Category Archives: Personality

Are Positive Illusions Really Good for You?

With 4,366 citations in WebOfScience, Taylor and Brown’s article “ILLUSIONS AND WELL-BEING: A SOCIAL PSYCHOLOGICAL PERSPECTIVE ON MENTAL-HEALTH” is one of the most cited articles in social psychology.

The key premises of the article is that human information processing is faulty and that mistakes are not random. Rather human information processing is systematically biased.

Taylor and Brown (1988) quote Fiske and Taylor’s (1984) book about social cognitions to support this assumption. “Instead of a naïve scientist entering the environment in search of the truth, we find the rather unflattering picture of a charlatan trying to make the data come out in a manner most advantageous to his or her already-held theories” (p. 88). 

30 years later, a different picture emerges. First, evidence has accumulated that human information processing is not as faulty as social psychologists assumed in the early 1980s. For example, personality psychologists have shown that self-ratings of personality have some validity (Funder, 1995). Second, it has also become apparent that social psychologists have acted like charlatans in their research articles, when they used questionable research practices to make unfounded claims about human behavior. For example, Bem (2011) used these methods to show that extrasensory perception is real. This turned out to be a false claim based on shoddy use of the scientific method.

Of course, a literature with thousands of citations also has produced a mountain of new evidence. This might suggest that Taylor and Brown’s claims have been subjected to rigorous tests. However, this is actually not the case. Most studies that examined the benefits of positive illusions relied on self-ratings of well-being, mental-health, or adjustment to demonstrate that positive illusions are beneficial. The problem is evident. When self-ratings are used to measure the predictor and the criterion, shared method variance alone is sufficient to produce a positive correlation. The vast majority of self-enhancement studies relied on this flawed method to examine the benefits of positive illusions (see meta-analysis by Dufner, Gebauer, & Sedikides, 2019).

However, there are a few attempts to demonstrate that positive illusions about the self predict well-being measures is measured by informant ratings to reduce the influence of shared method variance. The most prominent example is Taylor et al. (2003) article ” Portrait of the self-enhancer: Well adjusted and well liked or maladjusted and friendless.”
[Sadly, this was published in the Personality section of JPSP]

The abstract gives the impression that the results clearly favored Taylor’s positive illusions model. However, a closer inspection of reality shows that the abstract is itself illusory and disconnected from reality.

First, the study had a small sample size (N = 92). Second, only about half of these participants . Informant ratings were obtained from a single friend, but only 55 participants identified a friend who provided informant ratings. Even in 2003, it was common to use larger samples and more informants to measure well-being (e.g., Schimmack, & Diener, 2003). Moreover, friends are not as good as family members to report on well-being (Schneider & Schimmack, 2009). It only attests to Taylor’s social power that such a crappy, underpowered study was published in JPSP.

The results showed no significant correlations between various measures of positive illusions (self-enhancement) and peer-ratings of mental health (last row).

Thus, the study provided no evidence for the claim in the abstract that positive illusions about the self predict well-being or mental health without the confound of shared method variance.


Dufner, Gebauer, Sedikides, and Denissen (2019) conducted a meta-analysis of the literature. The abstract gives the impression that there is a clear positive effect of positive illusions on well-being.

Not surprisingly, studies that used self-ratings of adjustment/well-being/mental health showed positive association. The more interesting question is how self-enhancement measures are related to non-self-report measures of well-being. Table 3 shows that the meta-analysis identified 22 studies with an informant-rating of well-being and that these studies showed a small positive relationship, r = .12.

I was surprised that the authors found 22 studies because my own literature research uncovered fewer studies. So, I took a closer look at the 22 studies included in the meta-analysis (see APPENDIX).

Many of the studies relied on measures of social desirable responding (Marlow-Crowne Social Desirability Scale, Balanced -Inventory-of-Desirable Responding) as a measure of positive illusions. The problem with these studies is that social desirability scales also contain a notable portion of real personality variance. Thus, these studies do not conclusively demonstrate that illusions are related to informant ratings of adjustment. Paulhus’s studies are problematic because adjustment ratings were based on first-impressions in a zero-acquaintance relationship, and the results changed over time. Self-enhancers were perceived as better adjusted in the beginning, but as less adjusted later on. The problem here is that well-being ratings in this context have low validity. Finally, most studies were underpowered given the estimated population effect size of r = .12. The only reasonably powered study by Church et al. with 900 participants produced a correlation of r = .17 with an unweighted measure and r = .08 with a weighted measure. Overall, this evidence does not provide clear evidence that positive illusions about the self have positive effects. They actually show that any beneficial effects would be small.

New Evidence

In a forthcoming JRP article, Hyunji Kim and I present the most comprehensive test of Taylor’s positive illusion hypothesis (Schimmack & Kim, 2019). We collected data from 458 triads (students with both biological parents living together). We estimated separate models for students, mothers, and fathers as targets. In each model, targets self-ratings of the Big Five personality ratings were modelled with the halo-alpha-beta model, where the halo factor represents positive illusions about the self (Anusic et al., 2009). The halo factor was then allowed to predict the shared variance in well-being ratings by all three raters, and well-being ratings were based on three indicators (global life-satisfaction, average domain satisfaction, and hedonic balance, cf. Zou, Schimmack, & Gere, 2013).

The structural equation model is shown in Figure 1. The complete data, MPLUS syntax and output files and a preprint of the article are available on OSF (

The key findings are reported in Table 6. There were no significant relationships between self-rated halo bias and the shared variance among ratings of well-being across the three raters. Although this finding does not prove that positive illusions are not beneficial, the results suggest that it is rather difficult to demonstrate these benefits even in reasonably powered studies to detect moderate effect sizes.

The study did replicate much stronger relationships with self-ratings of well-being. However, this finding begs the question whether positive illusions are beneficial only in ways that are not visible to close others or whether these relationships simply reflect shared method variance.


Over 30 years ago, Taylor and Brown made the controversial proposal that humans benefit from distorted perceptions of reality. Only this year, a meta-analysis claimed that there is strong evidence to support this claim. I argue that the evidence in support of the illusion model is itself illusory because it rests on studies that relate self-ratings to self-ratings. Given the pervasive influence of rating biases on self-ratings, shared method variance alone is sufficient to explain positive correlations in these studies (Campbell & Fiske, 1959). Only a few studies have attempted to address this problem by using informant ratings of well-being as an outcome measure. These studies tend to find weak relationships that are often not significant. Thus, there is currently no scientific evidence to support Taylor and Brown’s social psychological perspective on mental health. Rather, the literature on positive illusions provides further evidence that social and personality psychologists have been unable to subject the positive illusions hypothesis to a rigorous test. To make progress in the study of well-being it is important to move beyond the use of self-ratings to reduce the influence of method variance that can produce spurious correlations among self-report measures.


Article#TitleStudy InformantsNSRIR
1Do Chinese Self-Enhance or Self-Efface?
It’s a Matter of Domain
1Table 4helpfulnessneuroticism1300.480.01
2How self-enhancers adapt well to loss: the mediational role of loneliness and social functioning1BIDR-SDSR symptoms (reversed) / IR mental health570.240.34
3Portrait of the self- enhancer:Well- adjusted and well- liked or maladjusted and friendless?1
4Social Desirability Scales: More Substance Than Style1Table 2MCSDdepression (reversed)2150.490.31
5Substance and bias in social desirability responding.12 FriendsTable 2SDEneuroticism (reversed)670.390.26
6Interpersonal and intrapsychic adaptiveness of trait self-enhancement: A mixed blessing1aZero-AquaintanceTable 2 Time 1Trait SEAdjustment124NA0.36
6Interpersonal and intrapsychic adaptiveness of trait self-enhancement: A mixed blessing1bZero-AquaintanceTable 2 Time 2Trait SEAdjustment124NA-0.11
6Interpersonal and intrapsychic adaptiveness of trait self-enhancement: A mixed blessing2Zero-AquaintanceTable 4 Time 1Trait SEAdjustment89NA0.35
6Interpersonal and intrapsychic adaptiveness of trait self-enhancement: A mixed blessing2Zero-AquaintanceTable 4 Time 1Trait SEAdjustment89NA-0.22
7A test of the construct validity of the Five-Factor Narcissism Inventory11 PeerTable 1FFNI VulnerabilityNeuroticism2870.50.33
8Moderators of the adaptiveness of self-enhancement: Operationalization, motivational domain, adjustment facet, and evaluator13 Peers/Family MembersSelf-ResidualsAdjustment1230.22-0.2
9Grandiose and Vulnerable Narcissism: A Nomological Network Analysis1NANA
10Socially desirable responding in personality assessment: Still more substance than style1a1 RoommateTable 1MCSDneuroticism (reversed)1280.410.06
10Socially desirable responding in personality assessment: Still more substance than style1bParentsTable 1MCSDneuroticism (reversed)1280.410.09
11Two faces of human happiness: Explicit and implicit life-satisfaction1a1 PeerTable 1BIDR-SDPANAS1590.450.17
11Two faces of human happiness: Explicit and implicit life-satisfaction1b1 PeerTable 1BIDR-SDLS1590.36-0.03
12Socially desirable responding in personality assessment: Not necessarily faking and not necessarily substance11 roommateTable 2BIDR-SDneuroticism (reversed)6020.260.02
13Depression and the chronic pain experience1noneMCSDNANA
14Trait self-enhancement as a buffer against potentially traumatic events: A prospective study1FriendsTable 5BIDR-SDmental health32NA-0.01
15Big Tales and Cool Heads: Academic Exaggeration Is Related to Cardiac Vagal Reactivity162NANA
16Are Actual and Perceived Intellectual Self-enhancers Evaluated Differently bySocial Perceivers?11 FriendTable 1 / above diagonalSE intelligenceneuroticism (reversed)3370.170.15
16Are Actual and Perceived Intellectual Self-enhancers Evaluated Differently bySocial Perceivers?3Zero-AquaintanceTable 1 / below diagonalSE intelligenceneuroticism (reversed)1830.190.38
17Response artifacts in the measurement of subjective well-being17 friends / familyTable 1MCSDLS1080.30.36
18A Four-Culture Study of Self-Enhancement and Adjustment Using the1a6 friends/ familyTable 6 SRM unweightedSRMLS9000.530.17
18A Four-Culture Study of Self-Enhancement and Adjustment Using the1b6 friends/ familyTable 6 SRM weightedSRMLS9000.490.08
19You Probably Think This Paper’s About You: Narcissists’ Perceptions of Their Personality and Reputation1NANA
20What Does the Narcissistic Personality Inventory Really Measure?4RoommatesNPI-GrandioseCollege Adjustment2000.480.27
21Self-enhancement as a buffer against extreme adversity: Civil war in Bosnia and traumatic loss in the United States1Mental Health ExpertsSelf-Peer Disadjustment difficulties (reversed)780.470.27
21Self-enhancement as a buffer against extreme adversity: Civil war in Bosnia and traumatic loss in the United States2Mental Health ExpertsTable 2  25 monthsBIDR-SDself distress / MHE PTSD740.30.35
22Self-enhancement among high-exposure survivors of the September 11th terrorist attack: Resilience or social maladjustment1Friend/FamilyBIDR-SDself depression 18 months / mental health450.290.33
23Decomposing a Sense of Superiority: The Differential Social Impact of Self-Regard and Regard for Others1Zero-AquaintanceSRMneuroticism (reversed)235NA0.02
24Personality, Emotionality, and Risk Prediction194NANA
24Personality, Emotionality, and Risk Prediction2119NANA
25Social desirability scales as moderator and suppressor variables1MCSD300NANA

Confirmation Bias is Everywhere: Serotonin and the Meta-Trait of Stability

Most psychologists have at least a vague understanding of the scientific method. Somewhere they probably heard about Popper and the idea that empirical data can be used to test theories. As all theories are false, these tests should at some point lead to an empirical outcome that is inconsistent with a theory. This outcome is not a failure. It is an expected outcome of good science. It also does not mean that the theory was bad. Rather it was a temporary theory that is now modified or replaced by a better theory. And so, science makes progress….

However, psychologists do not use the scientific method popperly. Null-hypothesis significance testing adds some confusion here. After all, psychologists publish over 90% successful rejections of the nil-hypothesis. Doesn’t that show they are good Popperians? The answer is no because the nil-hypothesis is not predicted by a theory. The nil-hypothesis is only useful to reject it to claim that there is a predicted relationship between two variables. Thus, psychology journals are filled with over 90% reports of findings that confirm theoretical predictions. While this may look like a major success, it actually shows a major problems. Psychologists never publish results that disconfirm a theoretical prediction. As a result, there is never a need to develop better theories. Thus, a root evil that prevents psychology from being a real science is verificationism.

The need to provide evidence for, rather than against, a theory led to the use of questionable research practices. Questionable research practices are used to report results that confirm theoretical predictions. For example, researchers may simply not report results of studies that did not reject the nil-hypothesis. Other practices can help to produce significant results by inflating the risk of a false positive result. The use of QRPs explains why psychology journals have been publishing over 90% results that confirm theoretical predictions for 60 years (Sterling, 1959). Only recently, it has become more acceptable to report studies that failed to support a theoretical prediction and question the validity of a theory. However, these studies are still a small minority. Thus, psychological science suffers from confirmation bias.

Structural Equation Modelling

Multivariate, correlational studies are different from univariate experiments. In a univariate experiment, a result is either significant or not. Thus, only tempering with the evidence can produce confirmation bias. In multivariate statistics, data are analyzed with complex statistical tools that provide researchers with flexibility in their data analysis. Thus, it is not necessary to alter the data to produce confirmatory results. Sometimes it is sufficient to analyze the data in a way that confirm a theoretical prediction without showing that alternative models fit the data equally well or better.

It is also easier to combat confirmation bias in multivariate research by fitting alternative models to the same data. Model comparison also avoids the problem of significance testing, where non-significant results are considered inconclusive, while significant results are used to confirm and cement a theory. In SEM, statistical inferences work the other way around. A model with good fit (non-significant chi-square or acceptable fit) is a possible model that can explain the data, while a model with significant deviation from the data is rejected. The reason is that the significance test (or model fit) is used to test an actual theoretical model rather than the nil-hypothesis. This forces researchers to specify an actual set of predictions and subject them to an empirical test. Thus, SEM is ideally suited to test theories popperly.

Confirmation Bias in SEM Research

Although SEM is ideally suited to test competing theories against each other, psychology journals are not used to model comparisons and tend to publish SEM research in the same flawed confirmatory way as other research is conducted and reported. For example, an article in Psychological Science this year published an investigation of the structure of personality and the hypothesis that several personality traits are linked to a bio-marker (Wright et al., 2019).

Their preferred model assumes that the Big Five traits neuroticism, agreeableness, and conscientiousness are not independent, but systematically linked by a higher-order triat called alpha or stability (Digman, 1997; DeYoung, 2007). In their model, the stability factor is linked to a marker of the serotonin (5-HT) prolactin response. This model implies that all three traits are related to the biomarker as there are indirect paths from all three traits to the biomarker that are “mediated” by the stability factor (for technical reasons the path goes from stabilty to the biomarker, but theoretically, we would expect the relationship to go the other way from a neurological mechanism to behaviour).

Thanks to the new world of open science, the authors shared actual MPLUS outputs of their models on OSF ( ). All the outputs also included the covariance matrix among the predictor variables, which made it possible to fit alternative models to the data.

Alternative Models

Another source of confirmation bias in psychology is that literature reviews fail to mention evidence that contradicts the theory that authors try to confirm. This is pervasive and by no means a specific criticism of the authors. Contrary to the claims in the article, the existence of a meta-trait of stability is actually controversial. Digman (1997) reported some SEM results that were false and could not be reproduced (cf. Anusic et al., 2009). Moreover, alpha could not be identified when the Big Five were modelled as latent factors (Anusic et al., 2009). This led me to propose that meta-traits may be an artifact of using impure Big Five scales as indicators of the Big Five. For example, if some agreeableness items have negative secondary loadings on neuroticism, the agreeableness scale is contaminated with valid variance in neuroticism. Thus, we would observe a negative correlation between neuroticism and agreeableness even across raters (e.g., self-ratings of neuroticism and informant ratings of agreeableness). Here I fitted a model with secondary loadings and independent Big Five factors to the data. I also examined the prediction that the biomarker is related to all three Big Five traits. The alternative model had acceptable fit, CFI = .976, RMSEA = .056.

The main finding of this model is that the biomarker shows only a significant relationship with conscientiousness, while the relationship with agreeableness trended in the right direction, but was not significant (p = .089) and the relationship for neuroticism was even weaker (p = .474). Aside from the question about significance, we also have to take effect sizes into account. Given the parameter estimates, the bimarker would produce very small correlations among the Big Five traits (e.g., r(A,C) = .19 * .10 = .019. Thus, even if these relationships were significant, they would not provide compelling evidence that a source of shared variance among the three traits has been identified.

The next model shows that the authors’s model ignored the stronger relationship between conscientiousness and the biomarker. When this relationship is added to the model, there is no significant relationship between the stability factor and the biomarker.

Thus, the main original finding of this study was that a serotonin related bio-marker was significantly related to conscientiousness, but not significantly related to neuroticism. This finding is inconsistent with theories that link neuroticism to serotonin, and evidence that serotonin reuptake inhibitors reduce neuroticism (at least in depressed patients). However, such results are difficult to publish because a single study with a non-significant results does not provide sufficient evidence to falsify a theory. However, fitting data to a theory only leads to confirmation bias.

The good news is that the authors were able to publish the results of an impressive study and that their data are openly available and can provide credible information for meta-analytic evaluations of structural models of personality, while the results of this study alone are inconclusive and compatible with many different theories of personality.

One way to take more advantage of these data would be to share the covariance matrix of items to model personality structure with a proper measurement model of the Big Five traits and to avoid the problem of contaminated scale scores, which is the best practice for the use of structural equation models. These models provide no evidence for Digman’s meta-traits (Schimmack, 2019a, Schimmack, 2019b).

In conclusion, the main point of this post is that (a) SEM can be used to test and falsify models, (b) SEM can be used to realize that data are consistent with multiple models and that better data are needed to find the better model, (c) studies of Big Five factors require a measurement model with Big Five factors and cannot rely on messy scale scores as indicators of the Big Five, and (d) personality psychologists need better training in the use of SEM.

32 Personality Types

Personality psychology is dominated by dimensional models of personality (Funder, 2019). There is a good reason for this. Most personality characteristics vary along a continuum like height rather than being categorical like eye color. Thus, a system of personality types requires some arbitrary decisions about a cutoff point. For example, a taxonomy of body types could do a median split on height and weight to assign people to the tall-heavy or the tall-light type.

However, a couple of moderately influential articles have suggested that there are three personality types (Asendorpf et al., 2001; Robins et al., 1996).

The notion that there are only three personality types is puzzling. The dominant framework in personality psychology is the Big Five model that conceptualizes personality traits as five independent continuous dimensions. If we were to create personality types by splitting each dimension at the median, it would create 32 personality types, where individuals are either above or below the median on neuroticism, extraversion, openness, agreeableness, and conscientiousness. if these five dimensions were perfectly independent of each other, we would see that individuals are equally likely to be assigned to one of the 32 types. There is no obvious way to reduce these 32 types to just 3.

Figure 1. small caps = below median, capitals = above mean

So, how did Robins et al. (1996) come to the conclusion that there are only three personality types? The data were Q-sorts. A Q-sort is similar to personality ratings on a series of attributes. The main difference is that the sorting task imposes a constraint on the scores that can be given to an individual. As a result, all individuals have the same overall mean across items. That is, nobody could be above average on all attributes. These kind of data are known as ipsative data. An alternative way to obtain ipsative data would be to subtract the overall mean of ratings from individual ratings. Although the distinction between ipsative and non-ipsative data is technically important, it has no implications for the broader understanding of Robins et al.’s work. The study could also have used ratings.

Robins et al. then performed a factor analysis. However, this factor analysis is different from a typical factor analysis that relies on correlations among items. Rather, the data matrix is transposed and the factor analysis is run on participants. With N = 300, there are three hundred variables and factor analysis is used to reduce this set of variables to a smaller set of factors, while minimizing the loss of information.

Everybody knows that the number of factors in a factor analysis is arbitrary and that a smaller number of factors implies a loss of information.

“Empirical research on personality typologies has been hampered by the lack of clear criteria for determining the number of types in a given sample. Thus, the costs and benefits of having a large number of types must be weighed against those of having relatively few types” (Robins et al., 1996).

The authors do not report Eigenvalues or other indicators of how much variance their three factor solution explained.

The three types are described in terms of the most and least descriptive items. Type 1 can be identified by high conscientiousness (“He is determined in what he does”), high extraversion (“He is energetic and full of life”), low neuroticism (reversed: “When he is under stress, he gives up and backs off”), high agreeableness (“He is open and straightforward”), and high openness (“He has a way with words”). In short, Type 1 is everybody’s dream child; a little Superman in the making.

Type 2 is characterized by high neuroticism (“He gets nervous in uncertain situations”), introversion (reversed: “He tries to be the center of attention”), low openness (reversed: he has a way with words,” but high agreeableness (“He is considerate and thoughtful of other people” ). Conscientiousness doesn’t define this type one way or the other.

Type 3 is characterized by low neuroticism (rerversed: “He is calm and relaxed; easy going”), high extraversion (“He tries to be the center of attention”), low conscientiousness (reversed: He plans things ahead; he thinks before he does something) and low agreeableness (He is stubborn”).

The main problem with this approach is that these personality profiles are not types. Take Profile 1 for example. While some participants’ profile correlated highly positively with Profile 1, some participants profile correlates highly negatively with Profile 1. What personality type are they? We might say that they are the opposite of Superman, but that would imply that we need another personality type for the Anti-Supermans. The problem doesn’t end here. As there are three profiles, each individual is identified by their correlations with all three profiles. Thus, we end up with eight different types depending on whether the correlation with the three profiles are positive or negative.

In short, profiles are not types. Thus, the claim that there are only three personality types is fundamentally flawed because the authors confused profiles with types. Even the claim that there are only 8 types would rest on the arbitrary choice of extracting only three factors. Four factors would have produced 16 types and five factors would have produced 32 types, just as the Big Five model predicted.

Asendorph et al. (2001) also found three profiles that they considered to be similar to those found by Robins et al. (1996). Moreover, they examined profiles in a sample of adult with a Big Five questionnaire (i.e., the NEO-FFI). Importantly, Asendorpf et al. (2001) use better terminology and refer to profiles as prototypes rather than types.

The notion of a prototype is that there are no clear defining features that determine class membership. For example, on average mammals are heavier than birds. So we can distinguish birds and mammals by their prototypical weight (how close their weight is to the average weight of a bird or mammal) rather than on the basis of a defining feature (lays eggs, has a uterus). Figure 2 shows the prototypical Big Five profile for the three groups of participants, when participants were assigned to three groups.

The problem is once more that the grouping into three groups is arbitrary. Clearly there are individuals with high scores on agreeableness and on openness, but this variation in personality was not used to create the three groups. Based on this figure, groupings are based on low N and high C, high N and low E, and low C. It is not clear what we should do with individuals who do not match any of these prototypical profiles. What type are individuals who are high in N and high in C?

In sum, a closer inspection of studies of personality types suggests that these studies failed to address the question. Searching for prototypical item-profiles is not the same thing as searching for personality types. In addition, the question may not be a good question. If personality attributes vary mostly quantitatively and if the number of personality traits is large, the number of personality types is infinite. Every individual is unique.

Are Some Personality Types More Common Than Others?

As noted above, the number of personality types that are theoretically possible is determined by the number of attributes and the levels of each attribute. If we describe personality with the Big Five and limit the levels to being above or below the median, we have 32 theoretical patterns. However, this does not mean that we actually observe all patterns. Maybe some types never occur or are at least rare. The absence of some personality types could provide some interesting insights into the structure of personality. For example, high conscientiousness might suppress neuroticism and we would see very few individuals who are high in C and low in N (Digman, 1997). However, when C is low, we could see equal numbers of individuals with high N and low N because conscientiousness only inhibits high N, while low conscientiousness does not lead to high N. It is impossible to examine such patterns with bivariate correlations (Feger, 1988).

A simple way to examine this question is to count the frequencies of personality traits (Anusic & Schimmack, unpublished manuscript that was killed in peer-review). Here, I present the results of this analysis based on Sam Gosling’s large internet survey with millions of visitors who completed the BFI (John, Naumann, & Soto, 2008).

Figure 3 simply shows the relative frequencies of the 32 personality types.

Figure 4 shows the results only for US residents. The results are very similar to those for the total sample.

The most notable finding is that the types nEOAC and Neoac are more frequent than all other types. These types are evaluatively positive or negative. However, it is important to realize that these types are not real personality types. Other research has demonstrated that the evaluative dimension in self-ratings of personality is mostly a rating or a perception bias (Anusic et al., 2009). Thus, individuals with a nEOAC profile do not have a better personality. Whether they simply rate themselves (other-deception) or actually see themselves (self-deception) as better than they are is currently unknown.

The next two types with above average frequency are nEoAC and NeOac. A simple explanation for this pattern is that openness is not highly evaluative and so some people will not inflate their openness scores, while they are still responding in a desirable way on the other four traits.

The third complementary pair are the neoAC and the NEOac types. This pattern can also be explained with rating biases because some people do not consider openness and extraversion desirable; so they will only show bias on neuroticism, agreeableness and conscientiousness. These people were called “Saints” by Paulhus and John (1998).

In short, one plausible explanation of the results is that all 32 personality types that can be created by combining high and low scores on the Big Five exists. Some types are more frequent than others, but at least some of this variation is explained by rating biases rather than by actual differences in personality.


The main contribution of this new look at personality types is to clarify some confusion about the notion of personality types. Previous researchers used the term types for prototypical personality profiles. This is unfortunate because it led to the misleading impression that there are only three personality types. You are either resilient, over-controlled, or under-controlled. In fact, even three profiles create more than three types. Moreover, the profiles are based on exploratory factor analyses of personality ratings and it is not clear why there are only three profiles. Big Five theory would predict five profiles where each profile is defined by items belonging to one of the Big Five factors. It is not clear why profile analyses yielded only three factors. One explanation could be that the item set did not capture some personality dimensions. For example, Robins et al.’s (1996) Q-sort did not seem to include many openness items.

Based on Big Five theory, one would expect 32 personality types that are about equally frequent. An analysis of a large data set showed that all 32 types exists, which is consistent with the idea that the Big Five are fairly independent dimensions that can occur in any combination. However, some types were more frequent than others. The most frequent combination was either desirable (nEOAC) or undesirable (Neoac). This finding is consistent with previous evidence that personality ratings are influenced by a general evaluative bias (Anusic et al., 2009). Additional types with higher frequencies can be attributed to variations in desirability. Openness and extraversion are not as desirable, on average, as low neuroticism and high agreeableness and conscientiousness. Thus, the patterns nEoAC and neoAC may also reflect desirability rather than actual personality structure. Multi-method studies or low evaluative items would be needed to examine this question.


Personality psychologists are frustrated that they have discovered the Big Five factors and created a scientific model of personality, but in applied settings the Myers-Briggs Type Indicator (MBTI) dominates personality assessment (Funder, 2019).

One possible reason is that the MBTI provides simple information about personality by classifying individuals into 16 types. These 16 types are defined by being high or low on four dimensions.

There is no reason, why personality psychologists could not provide simplified feedback about personality using a median split on the Big Five and assigning individuals to the 32 types that can be created by the Big Five factors. For example, I would be the NEOac type. Instead of using small caps and capitals, one could also use letters for both poles of the dimension, neurotic (N) vs. stable (S), extraverted (E) vs. introverted (I), variable (V) versus regular (R), agreeable (A) vs. dominant (D), and conscientious (C) vs. laid back (L). This would make me an NEVDL type. My son would be an SIRAC.

I see no reason why individuals would prefer Myer-Briggs types over Big Five types, given that the Big Five types are based on a well-established scientific theory. I believe the main problem in giving individuals feedback with Big Five scores is that many people do not think in terms of dimensions.

The main problem might be that we are assigning individuals to types even when their scores are close to the median and their classification is arbitrary. For example, I am not very high on E or low on C and it is not clear whether I am really an NEVDL or an NIVDC type. One possibility would be to use only scores that are one standard deviation above or below then mean or median. This would make me an N-VD- type.

To conclude, research on personality types has not made much progress for a good reason. The number of personality types depends on the number of attributes that are being considered and it is no longer an empirical question which types exists. With fairly independent dimensions all types exist and the number of types increases exponentially with the number of attributes. The Big Five are widely considered the optimal trade-off between accuracy and complexity. Thus, they provide an appealing basis for the creation of personality type and a viable alternative to the Myer-Briggs Type Indicator.

If you want to know what type you are, you can take the BFI online ( ). It provides feedback about your personality in terms of percentiles. To create your personality type, you only have to convert the percentiles into letters.

Negative Emotionality P < 50 = S P > 50 = N
Extraversion P < 50 = I P > 50 = E
Open-Mindedness P < 50 = R P > 50 = V
Agreeableness P < 50 = D P > 50 = A
Conscientiousness P < 50 = L P > 50 = C

However, keep in mind that your ratings and those of the comparison group are influenced by desirability.

If you are a NIRDL, you may have a bias to rate yourself as less desirable than you actually are

If you are an SEVAC, you may have a tendency to overrate your desirability.

A Psychometric Study of the NEO-PI-R

Galileo had the clever idea to turn a microscope into a telescope and to point it towards the night sky. His first discovery was that Jupiter had four massive moons that are now known as the Galilean moons (

Now imagine what would have happened if Galileo had an a priori theory that Jupiter has five moons and after looking through the telescope, Galileo decided that the telescope was faulty because he could see only four moons. Surely, there must be five moons and if the telescope doesn’t show them, it is a problem of the telescope. Astronomers made progress because they created credible methods and let empirical data drive their theories. Eventually even better telescopes discovered many more, smaller moons orbiting around Jupiter. This is scientific progress.

Alas, psychologists don’t follow the footsteps of natural sciences. They mainly use the scientific method to provide evidence that confirms their theories and dismiss or hide evidence that disconfirms their theories. They also show little appreciation for methodological improvements and often use methods that are outdated. As a result, psychology has made little progress in developing theories that rest of solid empirical foundations.

An example of this ill-fated approach to science is McCrae et al.’s (1996) attempt to confirm their five factor model with structural equation modeling (SEM). When they failed to find a fitting model, they decided that SEM is not an appropriate method to study personality traits because SEM didn’t confirm their theory. One might think that other personality psychologists realized this mistake. However, other personality psychologists were also motivated to find evidence for the Big Five. Personality psychologists had just recovered from an attack by social psychologists that personality traits does not even exist, and they were all too happy to rally around the Big Five as a unifying foundation for personality research. Early warnings were ignored (Block, 1995). As a result, the Big Five have become the dominant model of personality without subjecting the theory to rigorous tests and even dismissing evidence that theoretical models do not fit the data (McCrae et al., 1996). It is time to correct this and to subject Big Five theory to a proper empirical test by means of a method that can falsify bad models.

I have demonstrated that it is possible to recover five personality factors, and two method factors, from Big Five questionnaires (Schimmack, 2019a, 2019b, 2019c). These analyses were limited by the fact that the questionnaires were designed to measure the Big Five factors. A real test of Big Five theory requires to demonstrate that the Big Five factors explain the covariations among a large set of a personality traits. This is what McCrae et al. (1996) tried and failed to do. Here I replicate their attempt to fit a structural equation model to the 30 personality traits (facets) in Costa and McCrae’s NEO-PI-R.

In a previous analysis I was able to fit an SEM model to the 30 facet-scales of the NEO-PI-R (Schimmack, 2019d). The results only partially supported the Big Five model. However, these results are inconclusive because facet-scales are only imperfect indicators of the 30 personality traits that the facets are intended to measure. A more appropriate way to test Big Five theory is to fit a hierarchical model to the data. The first level of the hierarchy uses items as indicators of 30 facet factors. The second level in the hierarchy tries to explain the correlations among the 30 facets with the Big Five. Only structural equation modeling is able to test hierarchical measurement models. Thus, the present analyses provide the first rigorous test of the five-factor model that underlies the use of the NEO-PI-R for personality assessment.

The complete results and the MPLUS syntax can be found on OSF ( The NEO-PI-R data are from Lew Goldberg’s Eugene-Springfield community sample. Theyu are publicly available at the Harvard Dataverse



The NEO-PI-R has 240 items. There are two reasons why I analyzed only a subset of items. First, 240 variables produce 28,680 covariances, which is too much for a latent variable model, especially with a modest sample size of 800 participants. Second, a reflective measurement model requires that all items measure the same construct. However, it is often not possible to fit a reflective measurement model to the eight items of a NEO-facet. Thus, I selected three core-items that captured the content of a facet and that were moderately positively correlated with each other after reversing reverse-scored items. Thus, the results are based on 3 * 30 = 90 items. It has to be noted that the item-selection process was data-driven and needs to be cross-validated in a different dataset. I also provide information about the psychometric properties of the excluded items in an Appendix.

The first model did not impose a structural model on the correlations among the thirty facets. In this model, all facets were allowed to correlate freely with each other. A model with only primary factor loadings had poor fit to the data. This is not surprising because it is virtually impossible to create pure items that reflect only one trait. Thus, I added secondary loadings to the model until acceptable model fit was achieved and modification indices suggested no further secondary loadings greater than .10. This model had acceptable fit, considering the use of single-items as indicators, CFI = .924, RMSEA = .025, .035. Further improvement of fit could only be achieved by adding secondary loadings below .10, which have no practical significance. Model fit of this baseline model was used to evaluate the fit of a model with the Big Five factors as second-order factors.

To build the actual model, I started with a model with five content factors and two method factors. Item loadings on the evaluative bias factor were constrained to 1. Item loadings for on the acquiescence factor were constrained to 1 or -1 depending on the scoring of the item. This model had poor fit. I then added secondary loadings. Finally, I allowed for some correlations among residual variances of facet factors. Finally, I freed some loadings on the evaluative bias factor to allow for variation in desirability across items. This way, I was able to obtain a model with acceptable model fit, CFI = .926, RMSEA = .024, SRMR = .045. This model should not be interpreted as the best or final model of personality structure. Given the exploratory nature of the model, it merely serves as a baseline model for future studies of personality structure with SEM. That being said, it is also important to take effect sizes into account. Parameters with substantial loadings are likely to replicate well, especially in replication studies with similar populations.

Item Loadings

Table 1 shows the item-loadings for the six neuroticism facets. All primary loadings exceed .4, indicating that the three indicators of a facet measure a common construct. Loadings on the evaluative bias factors were surprisingly small and smaller than in other studies (Anusic et al., 2009; Schimmack, 2009a). It is not clear whether this is a property of the items or unique to this dataset. Consistent with other studies, the influence of acquiescence bias was weak (Rorer, 1965). Secondary loadings also tended to be small and showed no consistent pattern. These results show that the model identified the intended neuroticism facet-factors.

Table 2 shows the results for the six extraversion facets. All primary factor loadings exceed .40 and most are more substantial. Loadings on the evaluative bias factor tend to be below .20 for most items. Only a few items have secondary loadings greater than .2. Overall, this shows that the six extraversion facets are clearly identified in the measurement model.

Table 3 shows the results for Openness. Primary loadings are all above .4 and the six openness factors are clearly identified.

Table 4 shows the results for the agreeableness facets. In general, the results also show that the six factors represent the agreeableness facets. The exception is the Altruism facet, where only two items show a substantial loadings. Other items also had low loadings on this factor (see Appendix). This raises some concerns about the validity of this factor. However, the high-loading items suggest that the factor represents variation in selfishness versus selflessness.

Table 5 shows the results for the conscientiousness facets. With one exception, all items have primary loadings greater than .4. The problematic item is the item “produce and common sense” (#5) of the competence facet. However, none of the remaining five items were suitable (Appendix).

In conclusion, for most of the 30 facets it was possible to build a measurement model with three indicators. To achieve fit, the model included 76 out of 2,610 (3%) secondary loadings. Many of these secondary loadings were between .1 and .2, indicating that they have no substantial influence on the correlations of factors with each other.

Facet Loadings on Big Five Factors

Table 6 shows the loadings of the 30 facets on the Big Five factors. Broadly speaking the results provide support for the Big Five factors. 24 of the 30 facets (80%) have a loading greater than .4 on the predicted Big Five factor, and 22 of the 30 facets (73%) have the highest loading on the predicted Big Five factor. Many of the secondary loadings are small (< .3). Moreover, secondary loadings are not inconsistent with Big Five theory as facet factors can be related to more than one Big Five factor. For example, assertiveness has been related to extraversion and (low) agreeableness. However, some findings are inconsistent with McCrae et al.’s (1996) Five factor model. Some facets do not have the highest loading on the intended factor. Anger-hostility is more strongly related to low agreeableness than to neuroticism (-.50 vs. .42). Assertiveness is also more strongly related to low agreeableness than to extraversion (-.50 vs. .43). Activity is nearly equally related to extraversion and low agreeableness (-.43). Fantasy is more strongly related to low conscientiousness than to openness (-.58 vs. .40). Openness to feelings is more strongly related to neuroticism (.38) and extraversion (.54) than to openness (.23). Finally, trust is more strongly related to extraversion (.34) than to agreeableness (.28). Another problem is that some of the primary loadings are weak. The biggest problem is that excitement seeking is independent of extraversion (-.01). However, even the loadings for impulsivity (.30), vulnerability (.35), openness to feelings (.23), openness to actions (.31), and trust (.28) are low and imply that most of the variance in this facet-factors is not explained by the primary Big Five factor.

The present results have important implications for theories of the Big Five, which differ in the interpretation of the Big Five factors. For example, there is some debate about the nature of extraversion. To make progress in this research area it is necessary to have a clear and replicable pattern of factor loadings. Given the present results, extraversion seems to be strongly related to experiences of positive emotions (cheerfulness), while the relationship with goal-driven or reward-driven behavior (action, assertiveness, excitement seeking) is weaker. This would suggest that extraversion is tight to individual differences in positive affect or energetic arousal (Watson et al., 1988). As factor loadings can be biased by measurement error, much more research with proper measurement models is needed to advance personality theory. The main contribution of this work is to show that it is possible to use SEM for this purpose.

The last column in Table 6 shows the amount of residual (unexplained) variance in the 30 facets. The average residual variance is 58%. This finding shows that the Big Five are an abstract level of describing personality, but many important differences between individuals are not captured by the Big Five. For example, measurement of the Big Five captures very little of the personality differences in Excitement Seeking or Impulsivity. Personality psychologists should therefore reconsider how they measure personality with few items. Rather than measuring only five dimensions with high reliability, it may be more important to cover a broad range of personality traits at the expense of reliability. This approach is especially recommended for studies with large samples where reliability is less of an issue.

Residual Facet Correlations

Traditional factor analysis can produce misleading results because the model does not allow for correlated residuals. When such residual correlations are present, they will distort the pattern of factor loadings; that is, two facets with a residual correlation will show higher factor loadings. The factor loadings in Table 6 do not have this problem because the model allowed for residual correlations. However, allowing for residual correlations can also be a problem because freeing different parameters can also affect the factor loadings. It is therefore crucial to examine the nature of residual correlations and to explore the robustness of factor loadings across different models. The present results are based on a model that appeared to be the best model in my explorations. These results should not be treated as a final answer to a difficult problem. Rather, they should encourage further exploration with the same and other datasets.

Table 7 shows the residual correlation. First appear the correlations among facets assigned to the same Big Five factor. These correlations have the strongest influence on the factor loading pattern. For example, there is a strong correlation between the warmth and gregariousness facets. Removing this correlation would increase the loadings of these two facets on the extraversion factor. In the present model, this would also produce lower fit, but in other models this might not be the case. Thus, it is unclear how central these two facets are to extraversion. The same is also true for anxiety and self-consciousness. However, here removing the residual correlation would further increase the loading of anxiety, which is already the highest loading facet. This justifies the use of anxiety as the most commonly used indicator of neuroticism.

Table 7. Residual Factor Correlations

It is also interesting to explore the substantive implications of these residual correlations. For example, warmth and gregariousness are both negatively related to self-consciousness. This suggests another factor that influences behavior in social situations (shyness/social anxiety). Thus, social anxiety would be not just high neuroticism and low extraversion, but a distinct trait that cannot be reduced to the Big Five.

Other relationships are make sense. Modesty is negatively related to competence beliefs; excitement seeking is negatively related to compliance, and positive emotions is positively related to openness to feelings (on top of the relationship between extraversion and openness to feelings).

Future research needs to replicate these relationships, but this is only possible with latent variable models. In comparison, network models rely on item levels and confound measurement error with substantial correlations, whereas exploratory factor analysis does not allow for correlated residuals (Schimmack & Grere, 2010).


Personality psychology has a proud tradition of psychometric research. The invention and application of exploratory factor analysis led to the discovery of the Big Five. However, since the 1990s, research on the structure of personality has been stagnating. Several attempts to use SEM (confirmatory factor analysis) in the 1990s failed and led to the impression that SEM is not a suitable method for personality psychologists. Even worse, some researchers even concluded that the Big Five do not exist and that factor analysis of personality items is fundamentally flawed (Borsboom, 2006). As a result, personality psychologists receive no systematic training in the most suitable statistical tool for the analysis of personality and for the testing of measurement models. At present, personality psychologists are like astronomers who have telescopes, but don’t point them to the stars. Imagine what discoveries can be made by those who dare to point SEM at personality data. I hope this post encourages young researchers to try. They have the advantage of unbelievable computational power, free software (lavaan), and open data. As they say, better late than never.


Running the model with additional items is time consuming even on my powerful computer. I will add these results when they are ready.

What lurks beneath the Big Five?

Any mature science classifies the objects that it studies. Chemists classify atoms. Biologists classify organisms. It is therefore not surprising that personalty psychologists have spent a lot of their effort on classifying personality traits; that is psychological attributes that distinguish individuals from each other.

[It is more surprising that social psychologists have spent very little effort on classifying situations; a task that is now being carried out by personality psychologists (Rauthmann et al., 2014)]

After decades of analyzing correlations among self-ratings of personality items, personality psychologists came to a consensus that five broad factors can be reliably identified. Since the 1980s, the so-called Big Five have dominated theories and measurement of personality. However, most theories of personality also recognize that the Big Five are not a comprehensive description of personality. That is, unlike colors that can be produced by mixing three basic colors, specific personality traits are not just a mixture of the Big Five. Rather, the Big Five represent an abstract level in a hierarchy of personality traits. It is possible to compare the Big Five to the distinction of five classes of vertebrate animals: mammals, birds, reptiles, fish, and amphibians. Although there are important distinctions between these groups, there are also important distinctions among the animals within each class; cats are not dogs.

Although the Big Five are a major achievement in personality psychology, it also has some drawbacks. As early as 1995, personality psychologists warned that focusing on the Big Five would be a mistake because the Big Five are too broad to be good predictors of important life outcomes (Block, 1995). However, this criticism has been ignored and many researchers seem to assume that they measure personality when they administer a Big Five questionnaire. To warrant the reliance on the Big Five would require that the Big Five capture most of the meaningful variation in personality. In this blog post, I use open data to test this implicit assumption that is prevalent in contemporary personality science.

Confirmatory Factor Analysis

In 1996, McCrae et al. (1995) published an article that may have contributed to the stagnation in research on the structure of personality. In this article, the authors argued that structural equation modeling (SEM), specifically confirmatory factor analysis (CFA), is not suitable for personality researchers. However, CFA is the only method that can be used to test structural theories and to falsify structural theories that are wrong. Even worse, McCrae et al. (1995) demonstrated that a simple-structure model did not fit their data. However, rather than concluding that personality structure is not simple, they concluded that CFA is the wrong method to study personality traits. The problem with this line of reasoning is self-evident and was harshly criticized by Borsboom (2006). If we dismiss methods because they do not show a theoretically predicted pattern, we loose the ability to test theories empirically.

To understand McCrae et al.’s (1995) reaction to CFA, it is necessary to understand the development of CFA and how it was used in psychology. In theory, CFA is a very flexible method that can fit any dataset. The main empirical challenge is to find plausible models and to find data that can distinguish between competing plausible models. However, when CFA was introduced, certain restrictions were imposed on models that could be tested. The most restrictive model imposed that a measurement model should have only primary loadings and no correlated residuals. Imposing these restrictions led to the foregone conclusions that the data are inconsistent with the model. At this point, researchers were supposed to give up, create a new questionnaire with better items, retest it with CFA and find out that there were still secondary loadings that produced poor fit to the data. The idea that actual data could have a perfect structure must have been invented by an anal-retentive statistician who never analyzed real data. Thus, CFA was doomed to be useless because it could only show that data do not fit a model.

It took some time and courage to decide that the straight-jacket of simple structure has to go. Rather than giving up after a simple-structure model was rejected, the finding should encourage further exploration of the data to find models that actually fit the data. Maybe biologists initially classified whales as fish, but so what. Over time, further testing suggested that they are mammals. However, if we never get started in the first place, we will never be able to develop a structure of personality traits. So, here I present a reflective measurement model of personality traits. I don’t call it CFA, because I am not confirming anything. I also don’t call it EFA because this term is used for a different statistical technique that imposes other restrictions (e.g., no correlated residuals, local independence). We might call it exploratory modeling (EM) or because it relies on structural equation modeling, we could call it ESEM. However, ESEM is already been used for a blind computer-based version of CFA. Thus, the term EM seems appropriate.

The Big Five and the 30 Facets

Costa and McCrae developed a personality questionnaire that assesses personality at two levels. One level are the Big Five. The other level are 30 more specific personality traits.

Image result for costa mccrae facets

The 30 facets are often presented as if they are members of a domain, just like dogs, cats, pigs, horses, elephants, and tigers are mammals and have nothing to do with reptiles or bird. However, this is an oversimplification. Actual empirical data show that personality structure is more complex and that specific facets can be related to more than one Big Five factor. In fact, McCrae et al. (1996) published the correlations of the 30 facets with the Big Five factors and the table shows many, and a few substantial, secondary loadings; that is, correlations with a factor other than the main domain. For example, Impulsive is not just positively related to Neuroticism. It is also positively related to extraversion, and negatively related to conscientiousness.

Thus, McCrae et al.’s (1996) results show that Big Five data do not have a simple structure. It is therefore not clear what model a CONFIRMATORY factor analysis tries to confirm, when the CFA model imposes a simple structure. McCrae et al. (1995) agree: “If, however, small loadings are in fact meaningful, CFA with a simple structure model may not fit well” (p. 553). In other words, if an exploratory factor analysis shows a secondary loading of Anger/Hostility on Agreeableness (r = -.40), indicating that agreeable people are less likely to get angry, it makes no sense to confirm a model that sets this parameter to zero. McCrae et al. also point out that simple structure makes no theoretical sense for personality traits. “There is no theoretical reason why traits should not have meaningful loadings on three, four, or five factors:” (p. 553). The logical consequence of this insight is to fit models that allow for meaningful secondary loadings; not to dismiss modeling personality data with structural equations.

However, McCrae et al. (1996) were wrong about the correct way of modeling secondary loadings. “It is possible to make allowances for secondary loadings in CFA by fixing the loadings at a priori values other than zero” (p. 553). Of course, it is possible to fix loadings to a non-zero value, but even for primary loadings, the actual magnitude of a loading is estimated by the data. It is not clear why this approach could not be used for secondary loadings. It is only impossible to let all secondary loadings to be freely estimated, but there is no need to fix the loading of anger/hostilty on the agreeableness factor to a fixed value to model the structure of personality.

Personality psychologists in the 1990s also seemed to not fully understand how sensitive SEM is to deviations between model parameters and actual data. McCrae et al. (1996) critically discuss a model by Church and Burke (1994) because it “regarded loadings as small as ± .20 as salient secondaries” (p. 553). However, fixing a loading of .20 to a value of 0, introduces a large discrepancy that will hurt overall fit. One either has to free parameters or lower the criterion for acceptable fit. However, fixing loadings greater than .10 to zero and hoping to met standard criteria of acceptable fit is impossible. Effect sizes of r = .2 (d = .4) are not zero, and treating them as such will hurt model fit.

In short, exploratory studies of the relationship between the Big Five and facets show a complex pattern with many non-trivial (r > .1) secondary loadings. Any attempt to model these data with SEM needs to be able to account for this finding. As many of these secondary loadings are theoretically expected and replicable, allowing for these secondary loadings makes theoretical sense and cannot be dismissed as overfitting of data. Rather, imposing a simple structure that makes no theoretical sense should be considered underfiting of the data, which of course results in bad fit.

Correlated Residuals are not Correlated Errors

Another confusion in the use of structural equation modeling is the interpretation of residual variances. In the present context, residuals represent the variance in a facet scale that is not explained by the Big Five factors. Residuals are interesting for two reasons. First, they provide information about unique aspects of personality that are not explained by the Big Five. To use the analogy of animals, although cats and dogs are both animals, they also have distinct features. Residuals are analogous to these distinct features, and we would think that personality psychologists would be very interested in exploring this question. However, statistics textbooks tend to present residual variances as error variance in the context of measurement models where items are artifacts that were created to measure a specific construct. As the only purpose of the item is to measure a construct, any variance that does not reflect the intended construct is error variance. If we were only interested in measuring the Big Five, we would think about residual facet-variance as error variance. It does not matter how depressed people are. We only care about their neuroticism. However, the notion of a hierarchy implies that we do care about the valid variance in facets that is not explained by the Big Five. Thus, residual variance is not error variance.

The mistake of treating residual variance as error variance becomes especially problematic when residual variance in one facet is related to residual variance in another facet. For example, how angry people get (the residual variance in anger) could be related to how compliant people are (the residual variance in compliance). After all, anger could be elicit by a request to comply to some silly norms (e.g., no secondary loadings) that make no sense. There is no theoretical reason, why facets could only be linked by means of the Big Five. In fact, a group of researchers has attempted to explain all relations among personality facet without the Big Five because they don’t belief in broader factors (cf. Schimmack, 2019b). However, this approach has difficulties explaining the constistent primary loadings of facets on their predicted Big Five factor.

The confusion of residuals with errors accounts at least partially for McCrae et al.’s (1996) failure to fit a measurement model to the correlations among the 30 facets.

“It would be possible to specify a correlated error term between these two scales, but the interpretation of such a term is unclear. Correlated error usually refers to a nonsubstantive
source of variance. If Activity and Achievement Striving were, say, observer ratings, whereas all other variables were self-reports, it would make sense to control for this difference in method by introducing a correlated error term. But there are no obvious sources of correlated error among the NEO-PI-R facet scales in the present study” (p. 555).

The Big Five Are Independent Factors, but Evaluative Bias produces correlations among Big Five Scales

Another decision researchers have to make is whether they specify models with independent factors or whether they allow factors to be correlated. That is, are extraversion and openness independent factors or are extraversion and openness correlated. A model with correlated Big Five factors has 10 additional free parameters to fit the data. Thus, the model will is likely to fit better than a model with independent factors. However, the Big Five were discovered using a method that imposed independence (EFA and Varimax rotation). Thus, allowing for correlations among the factors seems atheoretical, unless an explanation for these correlations can be found. On this front, personality researchers have made some progress by using multi-method data (self-ratings and ratings by informants). As it turns out, correlations among the Big Five are only found in ratings by a single rater, but not in correlations across raters (e.g., self-rated Extraversion and informant-rated Agreeableness). Additional research has further validated that most of this variance reflects response styles in ratings by a single rater. These biases can be modeled with two method factors. One factor is an acquiescence factor that leads to higher or lower ratings independent of item content. The other factor is an evaluative bias (halo) factor. It represent responses to the desirability of items. I have demonstrated in several datasets that it is possible to model the Big Five as independent factors and that correlations among Big Five Scales are mostly due to the contamination of scale scores with evaluative bias. As a result, neuroticism scales tend to be negatively related to the other scales because neuroticism is undesirable and the other traits are desirable (see Schimmack, 2019a). Although the presence of evaluative biases in personality ratings has been known for decades, previous attempts at modeling Big Five data with SEM often failed to specify method factors; not surprisingly they failed to find good fit (McCrae et al., 1996. In contrast, models with method factors can have good fit (Schimmack, 2019a).

Other Problems in McCrae et al.’s Attempt

There are other problems with McCrae et al.’s (1996) conclusion that CFA cannot be used to test personality structure. First, the sample size was small for a rigorous study of personality structure with 30 observed variables (N = 229). Second, the evaluation of model fit was still evolving and some of the fit indices that they reported would be considered acceptable fit today. Most importantly, an exploratory Maximum Likelihood model produced reasonable fit, chi2/df = 1.57, RMS = .04, TLI = .92, CFI = .92. Their best fitting CFA model, however, did not fit the data. This merely shows a lack of effort and not the inability of fitting a CFA model to the 30 facets. In fact, McCrae et al. (1996) note “a long list of problems with the technique [SEM], ranging from technical difficulties in estimation
of some models to the cost in time and effort involved.” However, no science has made progress by choosing cheap and quick methods over costly and time-consuming methods simply because researchers lack the patients to learn a more complex method. I have been working on developing measurement models of personality for over a decade (Anusic et al., 2009). I am happy to demonstrate that it is possible to fit an SEM model to the Big Five data, to separate content variance from method variance, and to examine how big the Big Five factors really are.

The Data

One new development in psychology is that data are becoming more accessible and are openly shared. Low Goldberg has collected an amazing dataset of personality data with a sample from Oregon (the Eugene-Springfield community sample). The data are now publicly available at the Harvard Dataverse. With N = 857 participants the dataset is nearly four times larger than the dataset used by McCrae et al. (1996), and the ratio 857 observations and 30 variables (28:1) is considered good for structural equation modeling.

It is often advised to use different samples for exploration and then for cross-validation. However, I used the full sample for a mix of confirmation and exploration. The reason is that there is little doubt about the robustness of the data structure (the covariance/correlation matrix). The bigger issue is that a well-fitting model does not mean that it is the right model. Alternative models could also account for the same pattern of correlations. Cross-validation does not help with this bigger problem. The only way to address this is a systematic program of research that develops and tests different models. I see the present model as the beginning of such a line of research. Other researchers can use the same data to fit alternative models and they can use new data to test model assumptions. The goal is merely to boot a new era of research on the structure of personality with structural equation modeling, which could have started 20 years ago, if McCrae et al. (1996) had been more positive about the benefits of testing models and being able to falsify them (a.k.a. doing science).


I started with a simple model that had five independent personality factors (the Big Five) and an evaluative bias factor. I did not include an acquiescence factor because facets are measured with scales that include reverse scored items. As a result, acquiescence bias is negligible (Schimmack, 2019a).

In the initial model facet loadings on the evaluative bias factor were fixed at 1 or -1 depending on the direction or desirability of a facet. This model had poor fit. I then modified the model by adding secondary loadings and by freeing loadings on the evaluative bias factor to allow for variation in desirability of facets. For example, although agreeableness is desirable, the loading for the modesty facet actually turned out to be negative. I finally added some correlated residuals to the model. The model was modified until it reached criteria of acceptable fit, CFI = .951, RMSEA = .044, SRMR = .034. The syntax and the complete results can be found on OSF (

Table 3 shows the standardized loadings of the 30 facets on the Big Five and the two method factors.

There are several notable findings that challenge prevalent conceptions of personality.

The Big Five are not so big

First, the loadings of facets on the Big Five factors are notably weaker than in McCrae et al.’s Table 4 reproduced above (Table 2). There are two reasons for this discrepancy. First, often evaluative bias is shared between facets that belong to the same factor. For example, anxiety and depression have strong negative loadings on the evaluative bias factor. This shared bias will push up the correlation between the two facets and inflate factor loadings in a model without an evaluative bias factor. Another reason can be correlated residuals. If this extra shared variance is not modeled it pushes up loadings of these facets on the shared factor. The new and more accurate estimates in Table 3 suggest that the Big Five are not as big as the name implies. The loading of anxiety on neuroticism (r = .49) implies that only 25% of the variance in anxiety is captured by the neuroticism factor. Loadings greater than .71 are needed for a Big Five factor to explain more than 50% of the variance in a facet. There are only two facets where the majority of the variance in a facet is explained by a Big Five factor (order, self-discipline).

Secondary loadings can explain additional variance in some facets. For example, for anger/hostility neuroticism explains .48^2 = 23% of the variance and agreeableness explains another -.43^2 = 18% of the variance for a total of 23+18 = 41% explained variance. However, even with secondary loadings many facets have substantial residual variance. This is of course predicted by a hierarchical model of personality traits with more specific factors underneath the global Big Five traits. However, it also implies that Big Five measures fail to capture substantial personality variance. It is therefore not surprising that facet measures often predict additional variance in outcomes that it is not predicted by the Big Five (e.g., Schimmack, Oishi, Furr, & Funder, 2004). Personality researchers need to use facet level or other more specific measures of personality in addition to Big Five measures to capture all of the personality variance in outcomes.

What are the Big Five?

Factor loadings are often used to explore the constructs underlying factors. The terms neuroticism, extraversion, or openness are mere labels for the shared variance among facets with primary loadings on a factor. There has been some discussion about the Big Five factors and there meaning is still far from clear. For example, there has been a debate about the extraversion factor. Lucas, Diener, Grob, Suh, and Shao (2000) argued that extraversion is the disposition to respond strongly to rewards. Ashton, Lee, and Paunonen disagreed and argued that social attention underlies extraversion. Empirically it would be easy to answer these questions if one facet would show a very high loading on a Big Five factor. The more loadings approach one, the more a factor corresponds to a facet or is highly related to a facet. However, the loading pattern does not suggest that a single facet captures the meaning of a Big Five factor. The strongest relationship is found for self-discipline and conscientiousness. Thus, good self-regulation may be the core aspect of conscientiousness that also influences achievement striving or orderliness. However, more generally the results suggest that the nature of the Big Five factors is not obvious. It requires more work to uncover the glue that ties facets belonging to a single factor together. Theories range from linguistic structures to shared neurotransmitters.

Evaluative Bias

The results for evaluative bias are novel because previous studies failed to model evaluative bias in responses to the NEO-PI-R. It would be interesting to validate the variation in loadings on the evaluative bias factor with ratings of item- or facet-desirability. However, intuitively the variation makes sense. It is more desirable to be competent (C1, r = .66) and not depressed (N3, r = -69) than to be an excitement seeker (E5: r = .03) or compliant (A4: r = .09). The negative loading for modesty also makes sense and validates self-ratings of modesty (A5,r = -.33). Modest individuals are not supposed to exaggerate their desirable attributes and apparently they refrain from doing so also when they complete the NEO-PI-R.

Recently, McCrae (2018) acknowledged the presence of evaluative biases in NEO scores, but presented calculations that suggested the influence is relatively small. He suggested that facet-facet correlations might be inflated by .10 due to evaluative bias. However, this average is uninformative. It could imply that all facets have a loading of .33 or -.33 on the evaluative bias factor, which introduces a bias of .33*.33 = .10 or .33*-.33 = -.10 in facet-facet correlations. In fact, the average absolute loading on the evaluative bias factor is .30. However, this masks the fact that some facets have no evaluative bias and others have much more evaluative bias. For example, the measure of competence beliefs (self-effacy) C1 has a loading of .66 on the evaluative bias factor, which is higher than the loading on conscientiousness (.52). It should be noted that the NEO-PI-R is a commercial instrument and that it is in the interest of McCrae to claim that the NEO-PI-R is a valid measure for personalty assessment. In contrast, I have no commercial interest in finding more or less evaluative bias in the NEO-PI-R. This may explain the different conclusions about the practical significance of evaluative bias in NEO-PI-R scores.

In short, the present analysis suggests that the amount of evaluative bias varies across facet scales. While the influence of evaluative bias tends to be modest for many scales, scales with highly desirable traits show rather strong influence of evaluative bias. In the future it would be interesting to use multi-method data to separate evaluative bias from content variance (Anusic et al., 2009).

Measurement of the Big Five

Structural equation modeling can be used to test substantive theories with a measurement model or to develop and evaluate measurement models. Unfortunately, personality psychologists have not taken advantage of structural equation modeling to improve personality questionnaires. The present study highlights two ways in which SEM analysis of personality ratings is beneficial. First, it is possible to model evaluative bias and to search for items with low evaluative bias. Minimizing the influence of evaluative bias increases the validity of personality scales. Second, the present results can be used to create better measures of the Big Five. Many short Big Five scales focus exclusively on a single facet. As a result, these measures do not actually capture the Big Five. To measure the Big Five efficiently, a measure requires several facets with high loadings on the Big Five factor. Three facets are sufficient to create a latent variable model that separates the facet-specific residual variance from the shared variance that reflects the Big Five. Based on the present results, the following facets seem good candidates for the measurement of the Big Five.

Neuroticism: Anxiety, Anger, and Depression. The shared variance reflects a general tendency to respond with negative emotions.

Extraversion: Warmth, Gregariousness, Positive Emotions: The shared variance reflects a mix of sciability and cheerfulness.

Openness: Aesthetics, Action, Ideas. The shared variance reflects an interest in a broad range of activities that includes arts, intellectual stimulation, as well as travel.

Agreeableness: Straightforwardness, Altruism, Complicance: The shared variance represents respecting others.

Conscientiousness: Order, Self-Discipline, Dutifulness. I do not include achievement striving because it may be less consistent across the life span. The shared variance represents following a fixed set of rules.

This is of course just a suggestion. More research is needed. What is novel is the use of reflective measurement models to examine this question. McCrae et al. (1996) and some others before them tried and failed. Here I show that it is possible and useful to fit facet corelations with a structural equation model. Thus, twenty years after McCrae et al. suggested we should not use SEM/CFA, it is time to reconsider this claim and to reject it. Most personality theories are reflective models. It is time to test these models with the proper statistical method.

When Personality Psychologists are High

Correction (8/31/2019): In an earlier version, I misspelled Colin DeYoung’s name. I wrote DeYoung with a small d. I thank Colin DeYoung for pointing out this mistake.


One area of personality psychology aims to classify personality traits. I compare this activity to research in biology where organisms are classified into a large taxonomy.

In a hiearchical taxnomy, the higher levels are more abstract, less descriptive, but also comprise a larger group of items. For example, there are more mammals (class) than dogs (species).

in the 1980s, personality psychologists agreed on the Big Five. The Big Five represent a rather abstract level of description that combines many distinct traits into traits that are predominantly related to one of the Big Five dimensions. For example, talkative falls into the extraversion group.

To illustrate the level of abstraction, we can compare the Big Five to the levels in biology. After distinguishing vertebrate and invertebrate animals, there are five classes of vertebrate animals: mammals, fish, reptiles, birds, and amphibians). This suggests that the Big Five are a fairly high level of abstraction that cover a broad range of distinct traits within each dimension.

The Big Five were found using factor or pincipal component analysis (PCA). PCA is a methematical method that reduces the covariances among personality ratings to a smaller number of factors. The goal of PCA is to capture as much of the variance as possible with the smallest number of components. Evidently there is a trade-off. However, often the first components account for most of the variance while additional components add very little additional information. Using various criteria, five components seemed to account for most of the variance in personality ratings and the first five components could be identified in different datasets. So, the Big Five were born.

One important feature of PCA is that the components are independent (orthogonal). This is helpful to maximize the information that is captured with five dimensions. If the five dimensions would correlated, they would present overlapping variances and this redundancy would reduce the amount of explained variance. Thus, the Big Five are conceptually independent because they were discovered with a method that enforced independence.

Scale Scores are not Factors

While principal component analysis is useful to classify personality traits, it is not useful to do basic research on the causes and consequences of personality. For this purpose, personality psychologists create scales. Scales are usually created by summing items that belong to a common factor. For example, responses to the items “talkative,” “sociable,” and “reserved” are added up to create an extraversion score. Ratings of the item “reserved” are reversed so that higher scores reflect extraversion. Importantly, sum scores are only proxies of the components or factors that were identified in a factor analysis or a PCA. Thus, we need to distinguish between extraversion-factors and extraversion-scales. They are not the same thing. Unfortunately, personality psychologists often treat scales as if they were identical with factors.

Big Five Scales are not Independent

Now something strange happened when personalty psychologists examined the correlations among Big Five SCALES. Unlike the factors that were independent by design, Big Five Scales were not independent. Moreover, the correlations among Big Five scales were not random. Digman (1997) was the first to examine these correlations. The article has garnered over 800 citations.

Digman examined these correlations conducted another principal component analysis of the correlations. He found two factors. One factor for extraversion and openesss and the other factor for agreeableness and conscientiousness (and maybe low neuroticism). He proposed that these two factors represent an even higher level in a hierarchy of personality traits. Maybe like moving from the level of classess (mammals, fish, reptiles) to the level Phylum; a level that is so abstract that few people who are not biologists are familiar with.

Digman’s article stimulated further research on higher-order factors of personality, where higher means even higher than the Big Five, which are already at a fairly high level of abstraction. Nobody stopped to wonder how there could be higher-order factors if the Big Five are actually independent factors, and why Big Five scales show systematic correlations that were not present in factor analyses.

Instead personality psychologists speculated about the biological underpinning of the higher order factors. For example, Jordan B. Peterson (yes, them) and colleagues proposed that serotonin is related to higher stability (high agreeableness, high conscientiousness, and low neuroticism) (DeYoung, Peterson, and Higgins, 2002).

Rather than interpreting this finding as evidence that response tendencies contribute to correlations among Big Five scales, they interpreted this finding as a substantive finding about personality, society in the context of psychodynamic theories.

Only a few years later, separated from the influence of his advisor, DeYoung (2006) published a more reasonable article that used a multi-method approach to separate personality variance from method variance. This article provided strong evidence that a general evaluative bias (social desirable responding) contributes to correlations among Big Five Scales, which was formalized in Anusic et al.’s (200) model with an explicit evaluative bias (halo) factor.

However, the idea of higher-order factors was sustained by finding cross-method correlations that were consistent with the higher-order model.

After battling Colin as a reviewer, when we submitted a manuscript on halo bias in personality ratings, we finally were able to publish a compromise model that also included the higher order factors (stability/alpha; plasticity/beta), although we had problems identifying the alpha factor in some datasets.

The Big Mistake

Meanwhile, another article built on the 2002 model that did not control for rating biases and proposed that the correlation between the two higher-order factors implies that there is an even higher level in the hierarchy. The Big Trait of Personality makes people actually have more desirable personalities; They are less neurotic, more sociable, open, agreeable, and conscientious. Who wouldn’t want one of them as a spouse or friend? However, the 2006 article by DeYoung showed that the Big One only exists in the imagination of individuals and is not shared with perceptions by others. This finding was replicated in several datasets by Anusic et al. (2009).

Although claims about the Big One were already invalidated when the article was published, it appealed to some personality psychologists. In particular, white supremacist Phillip Rushton found the idea of a generally good personality very attractive and spend the rest of his life promoting it (Rushton & Irving, 2011; Rushton Bons, & Hur, 2008). He never realized the distinction between a personality factor, which is a latent construct, and a personality scale, which is the manifest sum-score of some personality items, and ignored DeYoung’s (2006) and other (Anusic et al., 2009) evidence that the evaluative portion in personality ratings is a rating bias and not substantive covariance among the Big Five traits.

Peterson and Rushton are examples of pseudo-science that mixes some empirical findings with grand ideas about human nature that are only loosely related. Fortunately, interest in the general factor of personality seems to be decreasing.

Higher Order Factors or Secondary Loadings?

Ashton, Lee, Goldberg, and deVries (2009) put some cold water on the idea of higher-order factors. They pointed out that correlations between Big Five Scales may result from secondary loadings of items on Big Five Factors. For example, the item adventurous may load on extraversion and openness. If the item is used to create an extraversion scale, the openness and extraversion scale will be positively correlated.

As it turns out, it is always possible to model the Big Five as independent factors with secondary loadings to avoid correlations among factors. After all, this is how exploratory factor analysis or PCA are able to account for correlations among personality items with independent factors or components. In an EFA, all items have secondary loadings on all factors, although some of these correlations may be small.

There are only two ways to distinguish empirically between a higher-order model and a secondary-loading model. One solution is to obtain measures of the actual causes of personality (e.g., genetic markers, shared environment factors, etc.) If there are higher order factors, some of the causes should influence more than one Big Five dimension. The problem is that it has been difficult to identify causes of personality traits.

The second approach is to examine the number of secondary loadings. If all openness items load on extraversion in the same direction (e.g., adventurous, interest in arts, interest in complex issues), it suggests that there is a real common cause. However, if secondary loadings are unique to one item (adventurous), it suggests that the general factors are independent. This is by no means a definitive test of the structure of personality, but it is instructive to examine how many items from one trait have secondary loadings on another trait. Even more informative would be the use of facet-scales rather than individual items.

I have examined this question in two datasets. One dataset is an online sample with items from the IPIP-100 (Johnson). The other dataset is an online sample with the BFI (Gosling and colleagues). The factor loading matrices have been published in separate blog posts and the syntax and complete results have been posted on OSF (Schimmack, 2019b; 2019c).


Neuroticism items show 8 out of 16 secondary loadings on agreeableness, and 4 out of 16 secondary loadings on conscientiousnes.

easily disturbed30.44-0.25
not easily bothered10-0.58-0.12-0.110.25
relaxed most of the time17-0.610.19-0.170.27
change my mood a lot250.55-0.15-0.24
feel easily threatened370.50-0.25
get angry easily410.50-0.13
get caught up in my problems420.560.13
get irritated easily440.53-0.13
get overwhelmed by emotions450.620.30
stress out easily460.690.11
frequent mood swings560.59-0.10
often feel blue770.54-0.27-0.12
panic easily800.560.14
rarely get irritated82-0.52
seldom feel blue83-0.410.12
take offense easily910.53
worry about things1000.570.210.09

Agreeableness items show only one secondary loading on conscientiousness and one on neuroticism.

indifferent to feelings of others8-0.58-0.270.16
not interested in others’ problems12-0.58-0.260.15
feel little concern for others35-0.58-0.270.18
feel others’ emotions360.600.220.17
have a good word for everybody490.590.100.17
have a soft heart510.420.290.17
inquire about others’ well-being580.620.320.19
insult people590.190.12-0.32-0.18-0.250.15
know how to comforte others620.260.480.280.17
love to help others690.140.640.330.19
sympathize with others’ feelings890.740.300.18
take time out for others920.530.320.19
think of others first940.610.290.17

Finally, conscientiousness items show only one secondary loading on agreeableness.

always prepared20.620.280.17
exacting in my work4-0.090.380.290.17
continue until everything is perfect260.140.490.130.16
do things according to a plan280.65-0.450.17
do things in a half-way manner29-0.49-0.400.16
find it difficult to get down to work390.09-0.48-0.400.14
follow a schedule400.650.070.14
get chores done right away430.540.240.14
leave a mess in my room63-0.49-0.210.12
leave my belongings around64-0.50-0.080.13
like order650.64-0.070.16
like to tidy up660.190.520.120.14
love order and regularity680.150.68-0.190.15
make a mess of things720.21-0.50-0.260.15
make plans and stick to them750.520.280.17
neglect my duties76-0.55-0.450.16
forget to put things back 79-0.52-0.220.13
shirk my duties85-0.45-0.400.16
waste my time98-0.49-0.460.14

Of course, there could be additional relationships that are masked by fixing most secondary loadings to zero. However, it also matters how strong the secondary loadings are. Weak secondary loadings will produce weak correlations among Big Five scales. Even the secondary loadings in the model are weak. Thus, there is little evidence that neuroticism, agreeableness, and conscientiousness items are all systematically related as predicted by a higher-order model. At best, the data suggest that neuroticism has a negative influence on agreeable behaviors. That is, people differ in their altruism, but agreeable neurotic people are less agreeable when they are in a bad mood.

Results for extraversion and openness are similar. Only one extraversion item loads on openness.

hard to get to know7-0.45-0.230.13
quiet around strangers16-0.65-0.240.14
skilled handling social situations180.650.130.390.15
am life of the party190.640.160.14
don’t like drawing attention to self30-0.540.13-0.140.15
don’t mind being center of attention310.560.230.13
don’t talk a lot32-0.680.230.13
feel at ease with people 33-0.200.640.160.350.16
feel comfortable around others34-0.230.650.150.270.16
find it difficult to approach others38-0.60-0.400.16
have little to say57-0.14-0.52-0.250.14
keep in the background60-0.69-0.250.15
know how to captivate people610.490.290.280.16
make friends easily73-0.100.660.140.250.15
feel uncomfortable around others780.22-0.64-0.240.14
start conversations880.700.120.270.16
talk to different people at parties930.720.220.13

And only one extraversion item loads on openness and this loading is in the opposite direction from the prediction by the higher-order model. While open people tend to like reading challenging materials, extraverts do not.

full of ideas50.650.320.19
not interested in abstract ideas11-0.46-0.270.16
do not have good imagination27-0.45-0.190.16
have rich vocabulary500.520.110.18
have a vivid imagination520.41-
have difficulty imagining things53-0.48-0.310.18
difficulty understanding abstract ideas540.11-0.48-0.280.16
have excellent ideas550.53-0.090.370.22
love to read challenging materials70-0.180.400.230.14
love to think up new ways710.510.300.18

The next table shows the correlations among the Big Five SCALES.

Scale CorrelationsNEOAC
Neuroticism (N)
Extraversion (E)-0.21
Openness (O)-0.160.13
Agreeableness (A)-
Conscientiousness (C)-

The pattern mostly reflects the influence of the evaluative bias factor that produces negative correlations of neuroticism with the other scales and positive correlations among the other scales. There is no evidence that extraversion and openness are more strongly correlated in the IPIP-100. Overall, these results are rather disappointing for higher-order theorists.

The next table shows the correlations among the Big Five Scales.

Scale CorrelationsNEOAC
Neuroticism (N)
Extraversion (E)-0.21
Openness (O)-0.160.13
Agreeableness (A)-
Conscientiousness (C)-

The pattern of correlations reflects mostly the influence of the evaluative bias factor. As a result, the neuorticism scale is negatively correlated with the other scales and the other scales are positively correlated with each other. There is no evidence for a stronger correlation between extraversion and openness because there are no notable secondary loadings. There is also no evidence that agreeableness and conscientiousness are more strongly related to neuroticism. Thus, these results show that DeYoung’s (2006) higher-order model is not consistent across different Big Five questionnaires.

Big Five Inventory

DeYoung found the higher-order factors with the Big Five Inventory. Thus, it is particularly interesting to examine the secondary loadings in a measurement model with independent Big Five factors (Schimmack, 2019b).

Neuroticism items have only one secondary loading on agreeableness and one on conscientiousness and the magnitude of these loadings is small.

emotionally stable24-0.610.270.18

Four out of nine agreeableness items have secondary loadings on neuroticism, but the magnitude of these loadings is small. Four items also have loadings on conscientiousness, but one item (forgiving) has a loading opposite to the one predicted by the hgher-order model.

find faults w. others20.15-0.42-0.240.19
helpful / unselfish70.440.100.290.23
start quarrels 120.130.20-0.50-0.09-0.240.19
trusting 220.150.330.260.20
cold and aloof27-0.190.14-0.46-0.350.17
considerate and kind320.040.620.290.23
like to cooperate420.15-0.100.440.280.22

For conscientiousness, only two items have a secondary loading on neuroticism and two items have a secondary loading on agreeableness.

thorough job30.590.280.22
careless 8-0.17-0.51-0.230.18
reliable worker13-
persevere until finished280.560.260.20
follow plans380.10-0.060.460.260.20
easily distracted430.190.09-0.52-0.220.17

Overall, these results provide no support for the higher-order model that predicts correlations among all neuroticism, agreeableness, and conscientiousness items. These results are also consistent with Anusic et al.’s (2009) difficulty of identifying the alpha/stability factor in a study with the BFI-S, a shorter version of the BFI.

However, Anusic et al. (2009) did find a beta-factor with BFI-S scales. The present analysis of the BFI do not replicate this finding. Only two extraversion items have small loadings on the openness factor.

full of energy110.34-0.110.580.20
generate enthusiasm160.070.440.110.500.20
shy and inhibited310.180.64-0.220.17

And only one openness item has a small loading that is opposite to the predicted direction. Extraverts are less likely to like reflecting.

ingenious 150.570.090.21
active imagination200.130.53-
value art300.120.460.090.160.18
like routine work35-
like reflecting40-0.080.580.270.21
few artistic interests41-0.26-0.090.15
sophisticated in art440.070.44-

In short, there is no support for the presence of a higher-order factor that produces overlap between extraversion and openness.

The pattern of correlations among the BFI scales, however, might suggest that there is an alpha factor because neuroticism, agreeableness and conscientiousness tend to be more strongly correlated with each other than with other dimensions. This shows the problem of using scales to study higher-order factors. However, there is no evidence for a higher-order factor that combines extraversion and openness as the correlation between these traits is an unremarkable r = .18.

Scale CorrelationsNEOAC
Neuroticism (N)
Extraversion (E)-0.26
Openness (O)-0.110.18
Agreeableness (A)-
Conscientiousness (C)-

So, why did DeYoung (2006) find evidence for higher-order factors? One possible explanation is that BFI scale correlations are not consistent across different samples. The next table shows the self-report correlations from DeYoung (2006) below the diagonal and discrepancies above the diagonal. Three of the four theoretically important correlations tend to be stronger in DeYoung’s (2006) data. It is therefore possible that the secondary loading pattern differs across the two datasets. It would be interesting to fit an item-level model to DeYoung’s data to explore this issue further.

Scale CorrelationsNEOAC
Neuroticism (N)0.100.03-0.06-0.08
Extraversion (E)-
Openness (O)-0.080.25-0.020.02
Agreeableness (A)-0.360.150.06-0.01
Conscientiousness (C)-0.310.210.090.24

In conclusion, an analysis of the BFI also does not support the higher-order model. However, results seem to be inconsistent across different samples. While this suggests that more research is needed, it is clear that this research needs to model personality at the level of items and not with scale scores that are contaminated by evaluative bias and secondary loadings.


Hindsight is 20/20 and after 20 years of research on higher-order factors a lot of this research looks silly. How could there be higher order factors for the Big Five factors if the Big Five are independent factors (or components) by default. The search for higher-order factors with Big Five scales can be attributed to methodological limitations, although higher-order models with structural equation modeling have been around since the 1980. It is rather obvious that scale scores are impure measures and that correlations among scales are influenced by secondary loadings. However, even when this fact was pointed out by Ashton et al. (2009), it was ignored. The problem is mainly due to the lack of proper training in methods. Here the problem is the use of scales as indicators of factors, when scales introduce measurement error and higher-order factors are method artifacts.

The fact that it is possible to recover independent Big Five factors from questionnaires that were designed to measure five independent dimensions says nothing about the validity of the Big Five model. To examine the validity of the Big Five as a valid model of the highest level in a taxonomy of personality trait it is important to examine the relationship of the Big Five with the diverse population of personality traits. This is an important area of research that could also benefit from proper measurement models. This post merely focused on the search for higher order factors for the Big Five and showed that searching for higher-order factors of independent factors is a futile endeavor that only leads to wild speculations that are not based on empirical evidence (Peterson, Rushton).

Even DeYoung and Peterson seems to have realized that it is more important to examine the structure of personality below rather than above the Big Five (DeYoung, Quility, & Peterson, 2007) . Whether 10 aspects, 16 factors (Cattell) or 30 facets (Costa & McCrae) represent another meaningful level in a hierarchical model of personality traits remains to be examined. Removing method variance and taking secondary loadings into account will be important to separate valid variance from noise. Also, factor analysis is superior to principle component analysis unless the goal is simply to describe personality with atheoretical components that capture as much variance as possible.

Correct me if you can

This blog post is essentially a scientific article without peer-review. I prefer this mode of communication over submitting manuscript to traditional journals where a few reviewers have the power to prevent research from being published. This happened with a manuscript that Ivana Anusic and I submitted and that was killed by Colin DeYoung as a reviewer. I prefer open reviews and I invite Colin to write an open review of this “article.” I am happy to be corrected and any constructive comments would be a welcome contribution to advancing personality science. Simply squashing critical work so that nobody gets to see it is not advancing science. The new way of conducting open science with open submissions, open reviews is the way to go. Of course, others are also invited to engage in the debate. So, let’s start a debate with the thesis “Higher-order factors of the Big Five do not exist.”

Personality and Self-Esteem

In the 1980s, personality psychologists agreed on the Big Five as a broad framework to describe and measure personality; that is, variation in psychological attributes across individuals.

You can think about the Big Five as a five-dimensional map. Like the two-dimensional map (or a three-dimensional globe), the Big Five are independent dimensions that create a space with coordinates that can be used to describe the vast number of psychological attributes that distinguish one person from another. One area of research in personality psychology is to correlate measures of personality attributes with Big Five measures to pinpoint their coordinates.

One important and frequently studied personality attribute is self-esteem, and dozens of studies have correlated self-esteem measures with Big Five measures. Robins, Tracy, and Trzesniewski (2001) reviewed some of these studies.

The results are robust and there is no worry about the replicability of these results. The strongest predictor of self-esteem is neuroticism vs. emotional stability. Self-esteem is located at the high end of neuroticism. The second predictor is extraversion vs. introversion. Self-esteem is located at the higher end of extraversion. The third predictor is conscientiousness which shows a slight positive location on the conscientious vs. careless dimension. Openness vs. closeness also shows a slight tendency towards openness. Finally, the results for agreeableness are more variable and show at least one negative correlation, but most correlations tend to be positive.

Evaluative Bias

Psychologists have a naive view of the validity of their measures. Although they sometimes compute reliability and examine convergent validity in methodological articles that are published in obscure journals like “Psychological Assessment,” they treat measures as perfectly valid in substantive articles that are published in journals like “Journal of Personality” or “Journal of Research in Personality.” Unfortunately, measurement problems can distort effect sizes and occasionally they can change the sign of a correlation.

Anusic et al. (2009) developed a measurement model for the Big Five that separates valid variance in the Big Five dimensions from rating biases. Rating biases can be content free (acquiescence) or respond to the desirability of items (halo, evaluative bias). They showed that evaluative bias can obscure the location of self-esteem in the Big Five space. Here, I revisit this question with better data that measure the Big Five with a measurement model fitted to the 44-items of the Big Five Inventory (Schimmack, 2019a).

I used the same data, which is the Canadian subsample of Gosling and colleagues large internet study that collects data from visitors who receive feedback about their personality. I simply added the single-item self-esteem measure to the dataset. I then fitted three different models. One model regressed the self-esteem item only on the Big Five dimensions. This model essentially replicates analyses with scale scores. I then added the method factors to the set of predictors.

Self-Esteem M1-0.430.300.08-0.030.16
Self-Esteem M2-0.330.190.00-

Results for the first model reproduce previous findings (see Table 1). However, results changed when the method factors were added. Most important, self-esteem is now placed on the negative side of agreeableness towards being more assertive. This makes sense given the selfless and other-focused nature of agreeableness. Agreeable people are less like to think about themselves and may subordinate their own needs to the needs of others. In contrast, people with high self-esteem are more likely to focus on themselves. Even though this is not a strong relationship, it is noteworthy that the relationship is negative rather than positive.

The other noteworthy finding is that evaluative bias is the strongest predictor of self-esteem. There are two interpretations of this finding and it is not clear which explanation accounts for this finding.

One interpretation is that self-esteem is rooted in a trait to see everything related to the self in an overly positive way. This interpretation implies that responses to personality items are driven by the desirability of items and individuals with high self-esteem see themselves as possessing all kinds of desirable attributes that they do not have (or have to a lesser degree). They think that they are kinder, smarter, funnier, and prettier than others, when they are actually not. In this way, the evaluative bias in personality ratings is an indirect measure of self-esteem.

The other interpretation is that evaluative bias is a rating bias that influences self-ratings, which includes self-ratings. Thus, the loading of the self-esteem item on the evaluative bias factor shows simply that self-esteem ratings are influenced by evaluative bias because self-esteem is a desirable attribute.

Disentangling these two interpretations requires the use of a multi-method approach. If evaluative bias is merely a rating bias, it should not correlated with actual life-outcomes. However, if evaluative bias reflects actual self-evaluations, it should be correlated with outcomes of high self-esteem.


Hopefully, this blog-post will create some awareness that personality psychology needs to move beyond the use of self-ratings in mapping the location of personality attributes in the Big Five space.

The blog post also has important implications for theories of personality development that assign value to personality dimensions (Dweck, 2008). Accordingly, the goal of personality development is to become more agreeable and conscientious and less neurotic among other things. However, I question that personality traits have intrinsic value. That is, agreeableness is not intrinsically good and low conscientiousness is not intrinsically bad. The presence of evaluative bias in personality items shows only that personality psychologists assign value to some traits and do not include items like “I am a clean-freak” in their questionnaires. Without a clear evaluation, there is no direction to personality change. Becoming more conscientious is no longer a sign of personal growth and maturation, but rather a change that may have positive or negative consequences for individuals. Although these issues can be debated, it is problematic that current models of personality development do not even question the evaluation of personality traits and treat the positive nature of some traits as a fundamental assumption that cannot be questioned. I suggest it is worthwhile to think about personality like sexual orientation or attractiveness. Although society has created strong evaluations that are hard to change, the goal should be to change these evaluations, not to change individuals to conform to these norms.

The Black Box of Meta-Analysis: Personality Change

Psychologists treat meta-analyses as the gold standard to answer empirical questions. The idea is that meta-analyses combine all of the relevant information into a single number that reveals the answer to an empirical question. The problem with this naive interpretation of meta-analyses is that meta-analyses cannot provide more information than the original studies contained. If original studies have major limitations, a meta-analytic integration does not make these limitations disappear. Meta-analyses can only reduce random sampling error, but they cannot fix problems of original studies. However, once a meta-analysis is published, the problems are often ignored and the preliminary conclusion is treated as an ultimate truth.

In this regard meta-analyses are like collateralized debt obligations that were popular until problems with CDOs triggered the financial crisis in 2008. A collateralized debt obligation (CDO) pools together cash flow-generating assets and repackages this asset pool into discrete tranches that can be sold to investors. The problem is when a CDO is considered to be less risky than the actual debt in the CDO actually is and investors believe they get high returns with low risks, when the actual debt is much more risky than investors believe.

In psychology, the review process and publication in a top journal give the appeal that information is trustworthy and can be cited as solid evidence. However, a closer inspection of the original studies might reveal that the results of a meta-analysis rest on shaky foundations.

Roberts et al. (2006) published a highly-cited meta-analysis in the prestigious journal Psychological Bulletin. The key finding of this meta-analysis was that personality levels change with age in longitudinal studies of personality.

The strongest change was observed for conscientiousness. According to the figure, conscientiousness doesn’t change much during adolescence, when the prefrontal cortex is still developing, but increases from d ~ .4 to d ~ .9 from age 30 to age 70 by about half a standard deviation.

Like many other consumers, I bought the main finding and used the results in my Introduction to Personality lectures without carefully checking the meta-anlysis. However, when I analyzed new data from longitudinal studies with large national representative samples, I could not find the predicted pattern (Schimmack, 2019a, 2019b, 2019c). Thus, I decided to take a closer look at the meta-analysis.

Roberts and colleagues list all the studies that were used with information about sample sizes, personality dimensions, and the ages that were studied. Thus, it is easy to find the studies that examined conscientiousness with participants who were 30 years or older at the start of the study.

Study NWeightStart1Max.IntervalES
Costa et al. (2000)22740.4441990.00
Costa et al. (1980)4330.08366440.00
Costa & McCrae (1988)3980.0835646NA
Labouvie-Vief & Jain (2002)3000.0639639NA
Branje et al. (2004)2850.064224NA
Small et al. (2003)2230.046866NA
P. Martin (2002)1790.03655460.10
Costa & McCrae (1992)1750.0353770.06
Cramer (2003)1550.03331414NA
Haan, Millsap, & Hartka (1986)1180.02331010NA
Helson & Kwan (2000)1060.02334247NA
Helson & Wink (1992)1010.0243990.20
Grigoriadis & Fekken (1992)890.023033
Roberts et al. (2002)780.024399
Dudek & Hall (1991)700.01492525
Mclamed et al. (1974)620.013633
Cartwright & Wink (1994)400.01311515
Weinryb et al. (1992)370.013922
Wink & Helson (1993)210.00312525
Total N / Average51441.00411119

There are 19 studies with a total sample size of N = 5,144 participants. However, sample sizes vary dramatically across studies from a low of N = 21 to a high of N = 2,274. Table 1 shows the proportion of participants that would be used to weight effect sizes according to sample sizes. By far the largest study found no significant increase in conscientiousness. I tried to find information about effect sizes from the other studies, but the published articles didn’t contain means or the information was from an unpublished source. I did not bother to obtain information from samples with less than 100 participants, because they contribute only 8% to the total sample size. Even big effects would be washed out by the larger samples.

The main conclusion that can be drawn from this information is that there is no reliable information to make claims about personality change throughout adulthood. If we assume that conscientiousness changes by half a standard deviation over a 40 year period, the average effect size for a decade is d = .12. For studies with even shorter retest intervals, the predicted effect size is even weaker. It is therefore highly speculative to extrapolate from this patchwork of data and make claims about personality change during adulthood.

Fortunately, much better information is now available from longitudinal panels with over thousand participants who have been followed for 12 (SOEP) or 20 (MIDUS) years with three or four retests. Theories of personality stability and change need to be revisited in the light of this new evidence. Updating theories in the face of new data is at the basis of science. Citing an outdated meta-analysis as if it provided a timeless answer to a question is not.

How Valid are Short Big-Five Scales?

The first measures of the Big Five used a large number of items to measure personality. This made it difficult to include personality measures in studies as the assessment of personality would take up all of the survey time. Over time, shorter scales became available. One important short Big Five measure is the BFI-S (Lang et al., 2011).  This 15-item measure has been used in several national representative, longitudinal studies such as the German Socio-Economic Panel (Schimmack, 2019a). These results provide unique insights into the stability of personality (Schimmack, 2019b) and the relationship of personality with other constructs such as life-satisfaction (Schimmack, 2019c). Some of these results overturn textbook claims about personality. However, critics argue that these results cannot be trusted because the BFI-S is an invalid measure of personality.

Thus, it is is critical importance to evaluate the validity of the BFI-S. Here I use Gosling and colleagues data to examine the validity of the BFI-S. Previously, I fitted a measurement model to the full 44-item BFI (Schimmack, 2019d). It is straightforward to evaluate the validity of the BFI-S by examining the correlation of the 3-item BFI-S scale scores with the latent factors based on all 44 BFI items. For comparison purposes, I also show the correlations for the BFI scale scores. The complete results for individual items are shown in the previous blog post (Schimmack, 2019d).

The measurement model for the BFS has seven independent factors. Five factors represent the Big Five and two factors represent method factors. One factor represents acquiescence bias. The other factor represents evaluative bias that is present in all self-ratings of personality (Anusic et al., 2009). As all factors are independent, the squared coefficients can be interpreted as the amount of variance that a factor explains in a scale score.

The results show that the BFI-S scales are nearly as valid as the longer BFI scales (Table 1).


For example, the factor-scale correlations for neuroticism, extraversion, and agreeableness are nearly identical. The biggest difference was observed for openness with a correlation of r = .76 for the BFI-scale and r = .66 for the BFI-S scale. The only other notable systematic variance in scales is the evaluative bias influence which tends to be stronger for the longer scales with the exception of conscientiousness. In the future, measurement models with an evaluative bias factor can be used to select items with low loadings on the evaluative bias factor to reduce the influence of this bias on scale scores. Given these results, one would expect that the BFI and BFI-S produce similar results. The next analyses tested this prediction.

Gender Differences

I examined gender differences three ways. First, I examined standardized mean differences at the level of latent factors in a model with scalar invariance (Schimmack, 2019d). Second, I computed standardized mean differences with the BFI scales. Finally, I computed standardized mean differences with the BFI-S scales. Table 2 shows the results. Results for the BFI and BFI-S scales are very similar. The latent mean differences show somewhat larger differences for neuroticism and agreeablness because these mean differences are not attenuated by random measurement error. The latent means also show very small gender differences for the method factors. Thus, mean differences based on scale scores are not biased by method variance.

Table 2. Standardized Mean Differences between Men and Women


Note. Positive values indicate higher means for women than for men.

In short, there is no evidence that using 3-item scales invalidates the study of gender differences.

Age Differences

I demonstrated measurement invariance for different age groups (Schimmack, 2019d). Thus, I used simple correlations to examine the relationship between age and the Big Five. I restricted the age range from 17 to 70. Analyses of the full dataset suggest that older respondents have higher levels of conscientiousness and agreeableness (Soto, John, Gosling, & Potter, 2011).

Table 3 shows the results. The BFI and the BFI-S both show the predicted positive relationship with conscientiousness and the effect size is practically identical. The effect size for the latent variable model is stronger because the relationship is not attenuated by random measurement error. Other relationships are weaker and also consistent across measures except for Openness. The latent variable model reveals the reason for the discrepancies. Three items (#15 ingenious, #l35 like routine work, and #10 sophisticated in art) showed unique relationships with age. The art-related items showed a unique relationship with age. The latent factor does not include the unique content of these items and shows a positive relationship between openness and age. The scale scores include this content and show a weaker relationship. The positive relationship of openness with age for the latent factor is rather surprising as it is not found in nationally representative samples (Schimmack, 2019b). One possible explanation for this relationship is that older individuals who take an online personality test are more open.


In sum, the most important finding is that the 3-item BFI-S conscientiousness scale shows the same relationship with age as the BFI-scale and the latent factor. Thus, the failure to find aging effects in the longitudinal SOEP data with the BFI-S cannot be attributed to the use of an invalid short measure of conscientiousness. The real scientific question is why the cross-sectional study by Soto et al. (2011) and my analysis of the longitudinal SOEP data show divergent results.


Science has changed since researchers are able to communicate and discuss research findings on social media. I strongly believe that open science outside of peer-controlled journals is beneficial for the advancement of science. However, the downside of social media of open science is that it becomes more difficult to evaluate expertise of online commentators. True experts are able to back up their claims with scientific evidence. This is what I did here. I showed that Brenton Wiernik’s comment has as much scientific validity as a Donald Trump tweet. Whatever the reason for the lack of personality change in the SOEP data will be, it is not the use of the BFI-S to measure the Big Five.

Personality Measurement with the Big Five Inventory

In one of the worst psychometric articles every published (although the authors still have a chance to retract their in press article before it is actually published), Hussey and Hughes argue that personality psychologists intentionally fail to test the validity of personality measures. They call this practice validity-hacking. They also conduct some psychometric tests of popular personality measures and claim that they fail to demonstrate structural validity.

I have demonstrated that this claim is blatantly false and that the authors failed to conduct a proper test of structural validity (Schimmack, 2019a). That is, the authors fitted a model to the data that is known to be false. Not surprisingly, they found that their model didn’t meet standard criteria of model fit. This is exactly what should happen when a false model is subjected to a test of structural validity. Bad models should not fit the data. However, a real test of structural validity requires fitting a plausible model to the data. I already demonstrated with several Big Five measures that these measures have good structural validity and that scale scores can be used as reasonable measures of the latent constructs (Schimmack, 2019b). Here I examine the structural validity of the Big Five Inventory (Oliver John) that was used by Hussay and Hughes.

While I am still waiting to receive the actual data that were used by Hussay and Hughes, I obtained a much larger and better dataset from Sam Gosling that includes data from 1 million visitors to a website that provides personality feedback (

For the present analyses I focused on the subgroup of Canadian visitors with complete data (N = 340,000). Subsequent analyses can examine measurement invariance with the US sample and samples from other nations. To examine the structure of the BFI, I fitted a structural equation model. The model has seven factors. Five factors represent the Big Five personality traits. The other two factores represent rating biases. One bias is an evaluative bias and the other bias is acquiescence bias. Initially, loadings on the method factors were fixed. This basic model was then modified in three ways. First, item loadings on the evaluative bias factor were relaxed to allow for some items to show more or less evaluative bias. Second, secondary loadings were added to allow for some items to be influenced by more than one factor. Finally, items of the same construct were allowed to covary to allow for similar wording or shared meaning (e.g., three arts items from the openness factor were allowed to covary). The final model and the complete results can be found on OSF (

Model fit was acceptable, CFI = .953, RMSEA = .030, SRMR = .032. In contrast, fitting a simple structure without method factors produced unacceptable fit for all three fit indices, CFI = .734, RMSEA = .068, SRMR = .110. This shows that the model specification by Hussey and Hughes accounted for the bad fit. It has been known for over 20 years that a simple structure does not fit Big Five data (McCrae et al., 1996). Thus, Hussay and Hughes claim that the BFI lacks validity is based on an outdated and implausible measurement model.

Table 1 shows the factor loading pattern for the 44 BFI items on the Big Five factors and the two method factors. It also shows the contribution of the seven factors to the scale scores that are used to provide visitors with personality feedback and in many research articles that use scale scores as proxies for the latent constructs.

emotionally stable24-0.610.270.18
full of energy110.34-0.110.580.20
generate enthusiasm160.070.440.110.500.20
shy and inhibited310.180.64-0.220.17
ingenious 150.570.090.21
active imagination200.130.53-
value art300.120.460.090.160.18
like routine work35-
like reflecting40-0.080.580.270.21
few artistic interests41-0.26-0.090.15
sophisticated in art440.070.44-
find faults w. others20.15-0.42-0.240.19
helpful / unselfish70.440.100.290.23
start quarrels 120.130.20-0.50-0.09-0.240.19
trusting 220.150.330.260.20
cold and aloof27-0.190.14-0.46-0.350.17
considerate and kind320.040.620.290.23
like to cooperate420.15-0.100.440.280.22
thorough job30.590.280.22
careless 8-0.17-0.51-0.230.18
reliable worker13-
persevere until finished280.560.260.20
follow plans380.10-0.060.460.260.20
easily distracted430.190.09-0.52-0.220.17

Most of the secondary loadings are very small, although they are statistically highly significant in this large sample. Most items also have the highest loading on the primary factor. Exceptions are the items blue/depressed, full of engery, and generate enthusiasm that have higher loadings on the evaluative bias factor. Except for two openness items, all items also have loadings greater than .3 on the primary factor. Thus, the loadings are consistent with the intended factor structure.

The most important results are the loadings of the scale scores on the latent factors. As the factors are all independent, squaring these coefficients shows the amount of explained variance by each factor. By far the largest variance component is the intended construct with correlations ranging from .76 for openness to .83 for extraversion. Thus, the lion share of the reliable variance in scale scores reflects the intended construct. The next biggest contributor is evaluative bias with correlations ranging from .36 for openness to .44 for extraversion. Although this means only 15 to 20 percent of the total variance in scale scores reflects evaluative bias, this systematic variance can produce spurious correlations when scale scores are used to predict other self-report measures (e.g., life satisfaction, Schimmack, 2019c).

In sum, a careful psychometric evaluation of the BFI shows that the BFI has good structural validity. The key problem is the presence of evaluative bias in scale scores. Although this requires caution in the interpretation of results obtained with BFI scales, it doesn’t justify the conclusion that the BFI is invalid.

Measurement Invariance

Hussey and Hughes also examined measurement invariance across age-groups and the two largest gender groups. They claimed that the BFI lacks measurement invariance, but this claim was based on a cunning misrepresentation of the results (Schimmack, 2019a). The claim is based on the fact that the simple-structure model does not fit in any group. However, fit did not decrease when measurement invariance was imposed on different groups. Thus, all groups showed the same structure and fit did not increase when measurement invariance was imposed, but this fact was hidden in the supplementary results.

I replicated their analyses with the current dataset. First, I fitted the model for the whole sample separately to the male and female samples. Fit for the male sample was acceptable, CFI = .949, RMSEA = .029, SRMR = .033. So was fit for the female sample, CFI = .947, RMSEA = .030, SRMR = .037.

Table 2 shows the results side by side. There are no notable differences between the parameter estimates for males and females (m/f). This finding replicates results with other Big Five measures (Schimmack, 2019a).

depressed/blue4.33/ .30-.18/-.11.19/ .20-.45/-.50.07/.05
relaxed9-.71/-.72.24/ .23.19/.18
tense14.52/ .49-.17/-.14.11/ .13-.27/-.32.20/ .20
worry19.58/ .57-.10/-.08.05/ .07-.22/-.22.17/ .17
emotionally stable24-.58/-.58.10/ .06.25/ .30.19/ .17
moody29.41/ .38-.26/-.25-.30/-.38.18/ .18
calm34-.55/-.59-.02/-.03.14/ .13.12/ .13-.27/-.24.21/ .19
nervous39.51/ .49-.21/.26-.10/-.10.08/ .08-.11/-.11-.27/-.25.18/ .17
SUM.78/ .77-.09/-.08-.01/-.01-.07/-.05-.02/-.02-.42-.46.05/ .04
talkative1.09/ .11.69/ .70-.10/-.08.24/ .24.19/ .18
reserved6-.55/-.60.08/.10.21/ .22.19/ .18
full of energy11.33/ .32-.09/-.04.56/ .59.21/ .20
generate enthusiasm16.04/ .03.44/ .43.12/ .13.48/ .50.20/ .20
quiet21-.79/-.82.03/ .04-.22/-.21.17/ .16
assertive26-.08/-.10.39/ .40.12/ .14-.23/-.25.18/ .17.26/ .24.20/ .18
shy and inhibited31.19/ .15.61/ .66.23/ .22.18/ .17
outgoing36.71/ .71.10/ .07.35/ .38.18/ .18
SUM-.02/-.02.82/ .82.04/ .05-.04-.06.00/ .00.45/ .44.07/ .06
original5.50/ .54-.12/-.12.40/ .39.22/ .20
curious10.40/ .42-.05/-.08.32/ .30.25/ .23
ingenious 150.00/0.00.60/ .56.18/ .16.10/ .04.22/ .20
active imagination20.50/ .55-.07/-.06-.17/-.18.29/ .26.23/ .21
inventive25-.07/ -.08.51/ .55-.12/-.10.37/ .34.21/ .19
value art30.10/ .03.43/ .52.08/ .07.17/ .14.18/ .19
like routine work35-.27/-.27.10/ .10.09/ .15-.22/-.21.17/ .16
like reflecting40-.09/-.08.58/ .58.28/ .26.22/ .20
few artistic interests41-.25/-.29-.10/-.09.16/ .15
sophisticated in art44.03/ .00.42/ .49-.08/-.08.09/ .09.16/ .16
SUM.01/ -.01-.01/-.01.74/ .78-.05/-.05-.03/-.06.38/ .34.20/ .19
find faults w. others2.14/ .17-.42/-.42-.24/-.24.19/ .19
helpful / unselfish7.45/ .43.09/.11.29/ .29.23/ .23
start quarrels 12.12/ .16.23/ .18-.49/-.49-.07/-.08-.24/-.24.19/ .19
forgiving17.49/ .46-.14/-.13.25/ .24.20/ .19
trusting 22-.14/-.16.38/ .32.27/ .25.21/ .19
cold and aloof27-.20/-.18.14/ .12.44/ .46-.34/-.37.18/ .17
considerate and kind32.02/.01.62/.61.28/ .30.22/ .23
rude37.10/ .12.12/ .12-.62/-.62-.13/-.08-.23/-.23.19/ .18
like to cooperate42.18/ .11-.09/-.10.43/ .45.28/ .29.23/ .22
SUM-.07/-.08.00/ .00-.07/-.07.78/ .77.03/ .03.43/ .44.04/ .04
thorough job3.58/ .59.29/ .28.23/ .22
careless 8-0.16-.49/-.51.24/ .23.19/ .18
reliable worker13-.10/-.09.09/ .10.55/ .55.30/ .31.24/ .24
disorganized18.13/ .16-.58/-.59-.21/-.20.17/ .15
lazy23-.52/-.51-.45/-.45.18/ .17
persevere until finished28.54/ .58.27/ .25.21/ .19
efficient33-.11/-.07.52/ .58.30/ .29.24/ .23
follow plans38.00/ .00-.06/-.07.45/ .44.27/ .26.21/ .20
easily distracted43.17/ .19.07/ .06-.53/-.53-.22/-.22.18/ .17
SUM-.05/-.05-.01/-.01-.05/-.06.04/ .04.81/ .82.43/ .41.03/ .03

I then fitted a multi-group model with metric invariance. Despite the high similarity between the individual models, model fit decreased, CFI = .925, RMSEA = .033, SRMR = .062. Although RMSEA and SRMR were still good, the decrease in fit might be considered evidence that the invariance assumption is violated. Table 2 shows that it is insufficient to examine changes in global fit indices. What matters is whether the decrease in fit has any substantial meaning. Given the results in Table 2, this is not the case.

The next model imposed scalar invariance. Before presenting the results, it is helpful to know what scalar invariance implies. Take extraversion as an example. Assume that there are no notable gender differences in extraversion. However, extraversion has multiple facets that are represented by items in the BFI. One facet is assertiveness and the BFI includes an assertiveness item. Scalar invariance implies that there cannot be gender differences in assertiveness if there are no gender differences in extraversion. It is obvious that this is an odd assumption because gender differences can occur at any level in the hierarchy of personality traits. Thus, evidence that scalar invariance is violated does not imply that we can not examine gender differences in personality. Rather, it would require further examination of the pattern of mean differences at the level of the factors and the item residuals.

However, imposing scalar invariance did not produce a decrease in fit, CFI = .921, RMSEA = .034, SRMR = .063. Inspection of the modification indices showed the highest modification index for item O6 “valuing art” with an implied mean difference of 0.058. This implies that there are no notable gender differences at the item-level. The pattern of mean differences at the factor level is consistent with previous studies, showing higher levels of neuroticism (d = .64) and agreeableness (d = .31), although the difference in agreeableness is relatively small compared to some other studies.

In sum, the results show that the BFI can be used to examine gender differences in personality and that the pattern of gender differences observed with the BFI is not a measurement artifact.

Age Differences

Hussey and Hughes used a median split to examine invariance across age-groups. The problem with a median split is that online samples tend to be very young. Figure 1 shows the age distribution for the Canadian sample. The median age is 22.

To create two age-groups, I split the sample into a group of under 30 and 30+ participants. The unequal sample size is not a problem because both groups are large given the large overall sample size (young N = 221,801, old N = 88,713). A published article examined age differences in the full sample, but the article did not use SEM to test measurement invariance (Soto, John, Gosling, & Potter, 2011). Given the cross-sectional nature of the data, it is not clear whether age differences are cohort differences or aging effects. Longitudinal studies suggest that age differences may reflect generational changes rather than longitudinal changes over time (Schimmack, 2019d). In any case, the main point of the present analyses is to examine measurement invariance across different age groups.

Fit for the model with metric invariance was similar to the fit for the gender model, CFI = .927, RMSEA = .033, SRMR = .062. Fit for the model with scalar invariance was only slightly weaker for CFI and better for RMSEA. More important, inspection of the modification indices showed the largest difference for O10 “sophisticated in art” with a standardized mean difference of .068. Thus, there were no notable differences between the two age groups at the item level.

The results at the factor level reproduced the finding with scale scores by Soto et al. (2011). The older group had a higher level of conscientiousness (d = .61) than the younger group. Differences for the other personalty dimensions were statistically small. There were no notable differences in response styles.

In sum, the results show that the BFI shows reasonable measurement invariance across age groups. Contrary to the claims by Hussey and Hughes, this finding is consistent with the results reported in Hussay and Hughes’s supplementary materials. These results suggest that BFI scale scores provide useful information about personality and that published articles that used scale scores produced meaningful results.


Hussey and Hughes accused personality researchers of validity hacking. That is, they do not report results of psychometric tests because these tests would show that personality measures are invalid. This is a strong claim that requires strong evidence. However, closer inspection of this claim shows that the authors used an outdated measurement model and misrepresented the results of their invariance analyses. Here I showed that the BFI has good structural validity and shows reasonable invariance across gender and age groups. Thus Hussay and Hughes’s claims are blatantly false.

So far, i have only examined the BFI, but I have little confidence in the authors’ conclusions about other measures like Rosenberg’s self-esteem scale. I am still waiting for the authors to share all of their data to examine all of their claims. At present, there is no evidence of v-hacking. Of course, this does not mean that self-ratings of personality are perfectly valid. As I showed, self-ratings of the Big Five are contaminated with evaluative bias. I presented a measurement model that can test for the presence of these biases and that can be used to control for rating biases. Future validation studies might benefit from using this measurement model as a basis for developing better measures and better measurement models. Substantive articles might also benefit from using a measurement model rather than scale scores, especially when the BFI is used as a predictor of other self-report measures to control for shared rating biases.