Identity and Culture, psychology homework help

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There is a great deal of research on individualism versus collectivism to explain the variety of cultural differences that make up a person’s identity. These studies show a variation in communication, expression, perception, and conflict avoidance that has become a framework of cultural theory (Matsumoto & Juang, 2008). Consider your own identity. Would you be the same person if you were raised in a culture that valued the group over the individual (collectivistic culture) or valued independence and the development of the self (individualistic culture)? How does your culture impact your identity?

Consider the impact of collectivistic and individualistic cultures on identity development.

1. Write a brief explanation of how a person’s identity may develop differently in a collectivistic versus an individualistic culture.

2. Explain how your own identity has been impacted by your culture (collectivistic or individualistic).

3. Explain how your identity might differ if you were raised in the other type of culture.

Support your responses using Evidenced based current literature. Please also utilize the below journal articles as references

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The Validity and Structure of Culture-Level Personality Scores: Data From Ratings of Young Adolescents Robert R. McCrae,1 Antonio Terracciano,1 Filip De Fruyt,2 Marleen De Bolle,2 Michele J. Gelfand,3 Paul T. Costa, Jr.,1 and 42 Collaborators of the Adolescent Personality Profiles of Cultures Project 1 National Institute on Aging 2 3 Ghent University University of Maryland ABSTRACT We examined properties of culture-level personality traits in ratings of targets (N 5 5,109) ages 12 to 17 in 24 cultures. The Adolescent Personality Profiles of Cultures Project collaborators include Maria E. Aguilar-Vafaie, Tarbiat Modarres University, Islamic Republic of Iran; Chang-kyu Ahn, Pusan National University, South Korea; Hyun-nie Ahn, Ewha Womans University, South Korea; Lidia Alcalay, Pontificia Universidad Catolica De Chile, Chile; Jüri Allik, University of Tartu, Estonia; Tatyana V. Avdeyeva, University of St. Thomas, USA; Marek Blatný, Academy of Science of the Czech Republic, Czech Republic; Denis Bratko, University of Zagreb, Croatia; Marina Brunner-Sciarra, Universidad Peruana Cayetano Heredia, Peru; Thomas R. Cain, Rutgers University, USA; Niyada Chittcharat, Srinakharinwirot University, Thailand; Jarret T. Crawford, The College of New Jersey, USA; Margarida P. de Lima, University of Coimbra, Portugal; Ryan Fehr, University of Maryland, USA; Emı́lia Ficková, Slovak Academy of Sciences, Slovak Republic; Sami Gülgöz, Koç University, Turkey; Martina Hřebı́čková, Academy of Science of the Czech Republic, Czech Republic; Lee Jussim, Rutgers University, USA; Waldemar Klinkosz, The John Paul II Catholic University of Lublin, Poland; Goran Kne&ević, Belgrade University, Serbia; Nora Leibovich de Figueroa, University of Buenos Aires, Argentina; Corinna E. Löckenhoff, Cornell University, USA; Thomas A. Martin, Susquehanna University, USA; Iris Marušić, Institute for Social Research, Zagreb, Croatia; Khairul Anwar Mastor, Universiti Kebangsaan Malaysia, Malaysia; Katsuharu Nakazato, Iwate Prefectural University, Japan; Florence Nansubuga, Makerere University, Uganda; Jose Porrata, San Juan, Puerto Rico; Danka Purić, Belgrade University, Serbia; Anu Realo, University of Tartu, Estonia; Norma Reátegui, Universidad Peruana Cayetano Heredia, Peru; Journal of Personality 78:3, June 2010 This article is a US Government work and is in the public domain in the USA. DOI: 10.1111/j.1467-6494.2010.00634.x 816 McCrae, Terracciano, De Fruyt, et al. Aggregate scores were generalizable across gender, age, and relationship groups and showed convergence with culture-level scores from previous studies of self-reports and observer ratings of adults, but they were unrelated to national character stereotypes. Trait profiles also showed crossstudy agreement within most cultures, 8 of which had not previously been studied. Multidimensional scaling showed that Western and non-Western cultures clustered along a dimension related to Extraversion. A culturelevel factor analysis replicated earlier findings of a broad Extraversion factor but generally resembled the factor structure found in individuals. Continued analysis of aggregate personality scores is warranted. The idea that the citizens of different nations have distinctive personalities can be traced to antiquity, and it was a central tenet of early 20th century culture and personality studies (LeVine, 2001). For a number of reasons, including the declining influence of psychoanalysis and ethical concerns about ethnocentrism (see Church, 2001), the topic fell out of favor, and interest has only recently been revived, this time from the perspective of trait psychology (Lynn & Martin, 1995; McCrae, Terracciano, & 79 Members of the Personality Profiles of Cultures Project, 2005; Schmitt et al., 2007). In this new approach, personality profiles of cultures can be obtained by averaging traits assessed in a sample of culture members, yielding a Jean-Pierre Rolland, Université Paris Ouest Nanterre La Défense, France; Vanina Schmidt, University of Buenos Aires, Argentina; Andrzej Sekowski, The John Paul II Catholic University of Lublin, Poland; Jane Shakespeare-Finch, Queensland University of Technology, Australia; Yoshiko Shimonaka, Bunkyo Gakuin University, Japan; Franco Simonetti, Pontificia Universidad Catolica De Chile, Chile; Jerzy Siuta, Jagiellonian University, Poland; Barbara Szmigielska, Jagiellonian University, Poland; Vitanya Vanno, Srinakharinwirot University, Thailand; Lei Wang, Peking University, People’s Republic of China; Michelle Yik, The Hong Kong University of Science and Technology, Hong Kong. Robert R. McCrae and Paul T. Costa, Jr., receive royalties from the Revised NEO Personality Inventory. This research was supported in part by the Intramural Research Program of the NIH, National Institute on Aging. The Czech contribution was supported by grant 406/07/1561 from the Grant Agency of the Czech Republic and is related to research plan AV0Z0250504 of the Institute of Psychology, Academy of Sciences of the Czech Republic. The authors are indebted to the following persons for their help with the data collection: Ana Butković, Sylvie Kouřilová, Valery E. Oryol, Ivan G. Senin, Vera V. Onufrieva, A. Maglio, I. Injoque Ricle, G. Blum, A. Calero, L. Cuenya, V. Pedrón, M. J. Torres Costa, D. Vion, Hamira Alavi, Kristina Burgetova, Shuo Chen, Irene Lee, Cindy Lo, and Javier Paredes. Correspondence concerning this article should be addressed to Robert R. McCrae, 809 Evesham Avenue, Baltimore, MD 21212. E-mail: RRMcCrae@gmail.com. Culture-Level Traits 817 set of aggregate personality traits. This is an etic approach, in which the same set of traits (usually identified in one culture) are studied across a range of cultures. The validity of these culture-level scores must be established, and there are at least two reasons to be skeptical about their accuracy. The first is that the personality trait scales that are aggregated may not themselves be commensurable across cultures: They may assess different constructs in different cultural contexts, or they may lack scalar equivalence (Nye, Roberts, Saucier, & Zhou, 2008; van de Vijver & Leung, 1997) due to problems in translation, in the relevance of particular items, or to cultural differences in response styles. These are theoretical threats to the validity of all cross-cultural measures. The second reason to doubt the validity of aggregate personality scores is that research to date suggests that they do not correspond to national character stereotypes (Perugini & Richetin, 2007). It is widely believed, for example, that the English are reserved—yet their aggregate personality scores suggest that they are in fact quite extraverted (McCrae, Terracciano, & 79 Members, 2005). This finding is not a fluke; analyses of data from 49 cultures suggested that national stereotypes are almost completely unrelated to aggregate personality traits (Terracciano et al., 2005). Many stereotypes have at least a kernel of truth (Madon et al., 1998), so the failure to find any association of national character stereotypes with aggregate personality scores is a legitimate source of concern. National character data from the Personality Profiles of Cultures (PPOC) project used by Terracciano and colleagues (2005)—and reanalyzed in the present article—were obtained by asking raters in each culture to describe the typical member of their own culture. Such judgments are sometimes called autostereotypes, in contrast to the heterostereotypes held by members of one culture about members of another. Several studies, however, have shown general agreement between these two kinds of stereotypes (Boster & Maltseva, 2006; Peabody, 1985). People around the world think that Americans are assertive and arrogant, and so do Americans (Terracciano & McCrae, 2007). Thus, the apparent inaccuracy of national character stereotypes is unlikely to be the result of ethnocentric or ethnophobic biases or of the way national character stereotypes were assessed. It is logically possible that both stereotypes and aggregated scores are invalid, but if forced to choose between them, researchers must rely on patterns of supporting evidence. Heine, Buchtel, and 818 McCrae, Terracciano, De Fruyt, et al. Norenzayan (2008), for example, showed that per capita Gross Domestic Product (GDP) is better predicted by stereotypes of Conscientiousness than by aggregate Conscientiousness scores. But this evidence is ambiguous, because in stereotypic thinking, industriousness is generally (mis)attributed to the wealthy (Fiske, Cuddy, Glick, & Xu, 2002), by a kind of variant of the fundamental attribution error. The weight of evidence to date favors the view that aggregate scores are accurate and national stereotypes are not (McCrae, Terracciano, Realo, & Allik, 2007b), largely because national stereotypes do not make psychological sense as indicators of national trait levels. For example, climate is one of the strongest correlates of national stereotypes of interpersonal warmth (McCrae, Terracciano, Realo, & Allik, 2007a), though few personality psychologists today believe that ambient temperature is a powerful influence on personality development. Stereotypes also fail to obey simple mathematical laws: The stereotype of Italians is not the mean of the stereotype of Northern and Southern Italians, but is almost identical with the latter (et al., 2007a). A number of cross-cultural methodologists (see Nye et al., 2008) have argued that the scalar equivalence of test items across cultures must be established before mean level comparisons are made—a strategy McCrae, Terracciano, and 79 Members (2005) labeled bottom-up. In contrast, McCrae and colleagues advocated a top-down strategy in which the construct validity of aggregate scores is examined directly. There is some support for the convergent validity of aggregate personality scores (e.g., Oishi & Roth, 2009), but it is still limited. Rentfrow, Gosling, and Potter (2008) provided validity data on aggregate personality scores for U.S. states, although those data do not address the difficulties posed by translation and cultural variations in response styles. McCrae, Terracciano, and 79 Members (2005) correlated culture-level scores from studies of self-reported personality traits with scores from observer-rated traits across 28 cultures. They found significant agreement for three (Neuroticism, Extraversion, and Openness) of the five factors and 26 of 30 facets of the Revised NEO Personality Inventory (NEO-PI-R; Costa & McCrae, 1992). Analyzed as profile agreement across the 30 facets within each culture, significant agreement was found for 22 of the 28 cultures. Aggregate personality scores also showed evidence of construct validity in their prediction of Hofstede’s (2001) dimensions of culture (Hofstede & McCrae, 2004) and in their geographical Culture-Level Traits 819 patterns (Allik & McCrae, 2004; McCrae, Terracciano, & 79 Members, 2005), in which Western cultures tended to cluster together in contrast to non-Western cultures. Using a different measure of personality, Schmitt and colleagues (2007) reported significant convergent validity between NEO-PI-R factor scores and Big Five Inventory (BFI) scales (John, Donahue, & Kentle, 1991) for three of the factors (Neuroticism, Extraversion, and Conscientiousness) across 27 cultures. (Discriminant validity was more problematic.) Persuasive evidence of the validity of culture-level aggregate personality scores would have important consequences for cross-cultural psychology. First, it would provide researchers with relatively accurate accounts of the prevailing personality traits in a variety of cultures, scores that might be used to predict a variety of nation-level outcomes of interest (McCrae & Terracciano, 2008). Second, it would reinforce the conclusion that national character stereotypes are almost completely unfounded—an observation with consequences both for the psychology of stereotypes and for the practice of international relations. Third, it would imply that the many theoretical concerns—potential threats to scalar equivalence—that have been raised about cross-cultural comparisons may have limited applicability in real-world data, and thus these concerns may have had an unwarranted chilling effect on mean comparisons in crosscultural research. Certainly, every cross-cultural researcher must continue to be vigilant against artifactual explanations of apparent cultural differences, but the validity of aggregate personality traits would serve as an encouragement to study such differences. With so much at stake, further evidence on the validity of aggregate personality traits is surely needed. The present article reports new data from the Adolescent Personality Profiles of Cultures (APPOC) Project, in which aggregate personality traits are scored from observer ratings of adolescents aged 12 to 17 in a sample of 24 cultures. Although this is a relatively small number, it includes 8 cultures (Argentina, Australia, Chile, Islamic Republic of Iran, Puerto Rico, Slovakia, Thailand, and Uganda) not previously included in culture-level studies of the validity of personality profiles. In studies of personality at the individual level, factor replication is an aspect of construct validity: If scales retain their validity in translation (and if the structure of personality is universal), then the same factor structure should emerge within each culture—as, for the most part, it does in analyses of the NEO-PI-R (McCrae, 820 McCrae, Terracciano, De Fruyt, et al. Terracciano, & 78 Members, 2005) and in world regional analyses of the BFI (Schmitt et al., 2007). However, replication of the individual-level factor structure at the culture level is not necessarily required, because the structure of personality may vary across levels of analysis. Previous research on the culture-level structure of the NEO-PI-R (McCrae, 2002; McCrae, Terracciano, & 79 Members, 2005) has suggested that the individual-level Five-Factor Model (FFM) is approximately replicated, but that the Extraversion factor is expanded to include aspects of other factors, including Impulsiveness, Openness to Fantasy and Values, and Competence—characteristics that appear to be higher in wealthier and more extraverted cultures. The present study provides an opportunity to replicate this culture-level finding. As a general rule, the analysis of aggregate scores ought to reproduce the individual level structure, unless there are specific effects on structure due to culture (J. Allik, personal communication, August 10, 2004; McCrae & Terracciano, 2008). The present study uses data on college students’ perceptions of adolescents ages 12 to 17, and previous analyses of these data at the individual level (De Fruyt et al., 2009) suggest one deviation from the universal adult factor structure: Openness to Ideas shows a substantial loading on Conscientiousness, perhaps because both diligence and an interest in ideas are attributed to adolescents who are known to be good students. It might therefore be hypothesized that a culture-level factor analysis of these adolescent data will show that aggregate Openness to Ideas loads on the Conscientiousness factor as well as the Openness factor. METHOD Procedure As detailed elsewhere (De Fruyt et al., 2009), collaborators from 27 sites representing 18 different languages from 24 cultures provided data. Ratings from multiple sites were available for the United States (three collaborating sites) and Poland (two collaborating sites). Collaborators were asked to collect anonymous observer ratings from college students who were randomly assigned one of four targets: a boy or girl ages 12 to 14 or 15 to 17 years. College student ratings were used instead of self-reports from adolescents for several reasons (convenience, data quality, comparability to PPOC data), but American studies (Costa, McCrae, & Martin, 821 Culture-Level Traits 2008; McCrae, Costa, & Martin, 2005) suggest that self-reports from adolescents would likely yield similar data. Collaborators were asked to provide data on 50 targets in each category. Participants received the following general instructions (cf. McCrae, Terracciano, & 78 Members, 2005): ‘‘This is a study of personality across cultures. We are interested in how people view others and rate their personality traits, and we will be comparing your responses to those of college students in other countries. Please think of a boy [girl] aged 12–14 [15–17] whom you know well. He [She] should be someone who is a native-born citizen of your country. He [She] can be a relative or a friend or neighbor—someone you like or someone you don’t like.’’ Valid ratings were obtained for 5,109 targets. Measures The NEO-PI-R (Costa & McCrae, 1992) is among the most frequently used inventories to assess the FFM and its dimensions of Neuroticism, Extraversion, Openness to Experience, Agreeableness, and Conscientiousness. The inventory has 30 facets, organized under the five domains, and includes 240 items (8 items per facet), presented with a 5-point Likert response scale. (For a discussion of the adequacy of this selection of facets to represent the five factors, see McCrae & Costa, 2008.) For the present study, participants were administered a questionnaire consisting of the 240 items of the NEO-PI-R and 37 additional items developed for the NEO-PI-3, a more readable version of the instrument (McCrae, Costa, et al., 2005). Previous analyses (De Fruyt et al., 2009) demonstrated that the psychometric properties of the NEO-PI-3 are maintained in the translations used in this study, and that the instrument is essentially equivalent to the NEO-PI-R in both structure and mean levels. It is therefore appropriate to compare NEO-PI-3 scores in the present sample with NEO-PI-R scores obtained in previous studies. NEO-PI-3 facet scales were standardized as T scores within the full sample (i.e., using individual level data, N 5 5,109, as international adolescent Form R NEO-PI-3 norms); factor scores were computed using the factor scoring weights for observer ratings presented in the manual (Costa & McCrae, 1992, Table 2, bottom panel). Aggregate scores were the mean T scores in each sample or subgroup. An index of data quality was also computed for each sample, based on four indicators: Number of protocols with more than 40 missing items, percentage of missing responses in valid protocols, number of protocols with evidence of acquiescence or naysaying, and responses in the unscreened sample to a single-item validity check asking respondents if they had answered honestly and accurately. Internal consistency of this quality index was .67. 822 McCrae, Terracciano, De Fruyt, et al. Criteria Validity of aggregate APPOC scores was examined by comparing scores to those previously reported in other samples. These include aggregate self-report NEO-PI-R data from a collection of available data sets (McCrae 2002; McCrae & Terracciano, 2008), observer rating NEO-PI-R data from the adult PPOC (McCrae, Terracciano, & 79 Members, 2005), and self-report BFI data (Schmitt et al., 2007). In addition, APPOC scores are also compared to national character stereotype (NCS) data (McCrae et al., 2007a), in which the ‘‘typical’’ member of a culture was rated by culture members on 30 scales corresponding to the facets of the NEO-PI-R. For example, the N1: Anxiety facet was assessed by asking if the typical culture member was ‘‘anxious, nervous, worrying vs. at ease, calm, relaxed.’’ When factored across nations, the structure of these stereotype ratings roughly replicated the structure of the NEO-PI-R (Terracciano et al., 2005). If stereotypes are, in fact, groundless, then NCS data provide information on the discriminant validity of aggregate trait scores. RESULTS AND DISCUSSION Preliminary Analyses We compared personality profiles in the three sites in the United States and the two sites in Poland. Using the SPSS Reliability program, treating sites as items and NEO-PI-3 facets as cases, we calculated average measure intraclass correlations under the absolute agreement definition. These values were .77 for the United States and .82 for Poland (pso.001). Data from these cultures were therefore collapsed (as the unweighted means of the different sites) for further analyses. In previous research (McCrae, Terracciano, & 79 Members, 2005), the variance of facet scores was related to geography, with larger standard deviations across the full range of facet scores for modern, Western cultures. The same pattern was found in the present study, with the lowest mean SDs in Malaysia, Peru, and Uganda, and the highest mean SDs in France, Australia, and Estonia. The correlation of mean SD in the present study with mean SD in the PPOC sample was r 5 .73, N 5 24, po.001. These geographical variations might be due to real differences in the homogeneity of traits in different cultures, to different response styles (e.g., acquiescence), or to differences in data quality, which also tends to be lower in nonWestern countries (see McCrae, Terracciano, & 79 Members, 2005). Culture-Level Traits 823 Also in previous research (Costa, Terracciano, & McCrae, 2001; Schmitt, Realo, Voracek, & Allik, 2008), the magnitude of gender differences was geographically ordered, with the most marked differences found in modern cultures. As in PPOC (McCrae, Terracciano, & 78 Members, 2005), we calculated gender difference indexes for each of the five factors, based on the facets on which adult women scored higher than men in self-reports (Costa, Terracciano, & McCrae, 2001). For example, because women scored higher than men on Openness to Aesthetics, Feeling, and Actions and lower on Openness to Ideas, a Female Openness/Closedness index was defined as (O2: Aesthetics1O3: Feelings1O4 Actions O5: Ideas)/4. Girls were rated significantly higher than boys in 74 of the 120 comparisons on the five indexes in 24 cultures. As in previous studies, the five indexes were positively intercorrelated and were summed to represent a general gender differentiation score (a 5 .78). As expected, the smallest differentiation was seen in Puerto Rico, Peru, and Uganda and the largest in Hong Kong, Slovakia, and Estonia. However, there were also some anomalous findings: Gender differentiation was low in Australia but relatively high in Malaysia. The correlation of gender differentiation in the present study with gender differentiation in the PPOC sample was only marginally significant (r 5 .37, N 5 23, po.05, one-tailed). In adult samples, lack of gender differentiation in traditional cultures has been attributed to the tendency of traditional men and women to compare themselves only to others of their own sex, in effect norming away gender differences in observed scores (Guimond et al., 2007). If so, then true gender differences are likely to be similar in all cultures. In any culture-level analysis it is necessary to recall that variation within cultures is usually far larger that variation across cultures. A components-of-variance analysis conducted on PPOC data (McCrae & Terracciano, 2008) showed that culture accounted for about 4% of the total variance, age (college vs. adult) for 3%, and sex for about 1%. Table 1 provides parallel information for APPOC. Here the effect of age is far smaller because the age groups differ very little. The effects of culture and sex, however, are similar to those seen in adult targets, although in adolescent targets, the effects of culture are most pronounced for Extraversion and least for Agreeableness. The top panel of Table 2 presents evidence on the generalizability of aggregate personality scores across gender and age groups. For these analyses, culture means for factor scores were derived for boys 824 McCrae, Terracciano, De Fruyt, et al. Table 1 Percentage of Variance in Observer-Rated NEO-PI-3 Factor Scores Attributable to Culture, Sex, and Age Factor Source Culture Sex Age Culture  Sex Culture  Age Sex  Age N E O A C Mean 3.6n 2.8n 0.2n 0.8n 0.8n 0.1n 5.0n 0.1n 0.0 0.6 0.7 0.0 2.9n 1.2n 0.0 0.9n 1.0n 0.0 1.5n 0.8n 0.2n 0.5 0.5 0.0 4.3n 2.2n 0.2n 0.5 0.7 0.0 3.46 1.42 0.12 0.66 0.74 0.02 Note. N 5 5,109. Age groups: 12 to 14 versus 15 to 17 years. Values are partial Z2 from a multivariate ANOVA. Three-way interactions were not significant. n po.05. and girls (or younger and older targets) separately and correlated across the 24 cultures. All correlations are significant, suggesting that similar estimates of culture-level means would be obtained regardless of the age or gender of the targets. We asked about the relationship of raters to targets and found that it varied somewhat across cultures. For example, 30% of the targets in Thailand were relatives of the raters, whereas 87% were relatives in Iran. De Fruyt and colleagues (2009) created a familiarity index based on questions about how well the raters knew the target, how often they saw them, and in how many different contexts. On a 0 to 4 scale, familiarity values ranged from 1.88 in Japan to 3.35 in Australia. Raters reported that they had known targets for from 0 to 17 years, with a mean of 9.2 years, but none of the raters had known their targets for over 10 years in Croatia or Portugal. Because of these differences across samples, we conducted analyses of variance on the five factors with culture and each of the dichotomized relationship categories as classifying variables. Most of the effects, even when significant in this large sample, were trivial in magnitude, and none of the main effects for relationship category or interaction effects accounted for more than 1% of the variance. The largest main effect showed that, unsurprisingly, well-known targets were rated higher in Extraversion (M 5 50.7) than less well-known targets (M 5 48.5). We also examined the generalizability of aggregate 825 Culture-Level Traits Table 2 Generalizability and Convergent Correlations of Culture-Level Factor Scores APPOC Factor N Generalizability Across gender .68nnn Across age .61nnn Across relationships Type .84nnn a Length .79nnn Familiarity .82nnn Convergent correlation Form R .50nn Form S .44n BFI .44n E O A C .82nnn .79nnn .56nn .50nn .54nn .49nn .83nnn .72nnn .80nnn .78nnn .65nnn .59nn .56nn .65nnn .49n .33 .43n .73nnn .63nn .76nnn .55nn .74nnn .45n .37n .14 .27 .02 .35 .05 .09 .36 .17 Note. Type 5 friend or acquaintance (N 5 2,456) versus relative (N 5 2,588). Length 5 known for less than (N 5 2,528) versus more than (N 5 2,300) 10 years. Familiarity 5 lower (N 5 2,327) versus higher (N 5 2,629). Form R 5 observer rating NEO-PI-R data, N 5 24, from McCrae, Terracciano, & 79 Members of the Personality Profiles of Cultures Project (2005); Form S 5 self-report NEO-PI-R data, N 5 16, from McCrae (2002) and McCrae and Terracciano (2008); BFI 5 self-report Big Five Inventory data, N 5 18, from Schmitt et al. (2007). a Across 22 cultures. n po.05, nnpo.01, nnnpo.001, one-tailed. scores across relationship categories. The top panel of Table 2 shows that, in general, there is strong replicability. Within this pool of generally well-acquainted raters, the details of the relationship do not seem to have major effects, so sample differences in these details are unlikely to affect results. Convergent and Discriminant Validity of Aggregate Scores Validity of Scales Across Cultures The bottom panel of Table 2 shows correlations with aggregate observer ratings (Form R) and self-reports (Form S) on the NEO-PI-R from previous studies. It also presents correlations with aggregated BFI self-reports. There is strong evidence of convergent validity for 826 McCrae, Terracciano, De Fruyt, et al. the Neuroticism and Extraversion factors, only weak evidence for Openness, and no evidence in these data for the validity of aggregate Agreeableness and Conscientiousness scores. Nonsignificant correlations for the Agreeableness factor across studies were also reported by McCrae, Terracciano, and 79 Members (2005) and Schmitt and colleagues (2007). Table 3 provides convergent validity information at the level of the facet scales. The intraclass correlation (first data column; ICC(1, k) 5 [BMS–WMS]/BMS) reflects agreement among raters on targets from each of the 24 cultures and estimates the reliability of the aggregate scores. These values are very slightly smaller than those found in analyses of adult targets (Mdn ICC 5 .91; McCrae, Terracciano, & 79 Members, 2005). The second and third data columns in Table 3 show convergent correlations with observer rating and self-report data on the NEO-PI-R. For Form R, 23 (76.7%) of the facets show significant cross-study agreement; for Form S, 20 (66.7%) are significant. E2: Gregariousness, O4: Actions, O5: Ideas, C3: Dutifulness, and C5: Self-Discipline failed to reach significance in either comparison; Dutifulness and Self-discipline also failed to show cross-study agreement in the PPOC study (McCrae, Terracciano, & 79 Members, 2005). However, the present data relate aggregate traits in ratings of adolescents using the NEO-PI-3 to aggregate traits in ratings and self-reports of adults using the original NEOPI-R; from this perspective the overall degree of convergence is striking. A comparison of Tables 3 and 2 highlights a puzzling finding: Why are the traits that define the Agreeableness and Conscientiousness factors generally related across studies, whereas the factors themselves are not? In both PPOC (McCrae, Terracciano, & 79 Members, 2005) and APPOC (reported below in Table 5), culturelevel analyses clearly show Agreeableness and Conscientiousness factors because the facets covary as expected. But the cross-facet, cross-study correlations are not consistently positive. For example, the correlation between aggregate A4: Compliance in adolescents and aggregate A5: Modesty in adults is .53, po.01. Such anomalies may be due to the small sample size (N 5 24), but they may also imply that there is more agreement on facet-specific variance than on common variance at the culture level. The last column of Table 3 reports correlations between APPOC aggregate traits and NCS scores across 22 cultures. Five correlations are significant, but three of them are negative. The positive associ- 827 Culture-Level Traits Table 3 Intraclass Reliability and Cross-Instrument Correlations for NEO-PI-3 Facet Scales ra NEO-PI-3 Facet Scale ICC(1,k) Form R Form S NCS N1: N2: N3: N4: N5: N6: Anxiety Angry Hostility Depression Self-Consciousness Impulsiveness Vulnerability .90 .79 .86 .77 .87 .90 .65nnn .52nn .55nn .40n .51nn .61nnn .79nnn .03 .46n .43n .60nn .72nnn .05 .18 .17 .10 .05 .54nn E1: E2: E3: E4: E5: E6: Warmth Gregariousness Assertiveness Activity Excitement Seeking Positive Emotions .90 .84 .76 .89 .91 .81 .60nnn .18 .37n .39n .49nn .43n .33 .27 .67nn .51n .82nnn .35 .40(n) .27 .00 .26 .35 .41(n) O1: O2: O3: O4: O5: O6: Fantasy Aesthetics Feelings Actions Ideas Values .91 .90 .90 .88 .84 .92 .54nn .58nn .78nnn .34 .28 .61nnn .40 .12 .56n .04 .08 .75nnn .10 .21 .14 .29 .07 .04 A1: A2: A3: A4: A5: A6: Trust Straightforwardness Altruism Compliance Modesty Tender-Mindedness .90 .82 .90 .91 .80 .89 .48nn .24 .74nnn .60nnn .63nnn .32 .48n .65nn .72nnn .44n .70nn .47n .20 .26 .04 .36n .08 .02 C1: C2: C3: C4: C5: C6: Competence Order Dutifulness Achievement Striving Self-Discipline Deliberation .81 .88 .86 .90 .84 .92 .52nn .47n .10 .44n .24 .58nn .63nn .48n .42 .52n .18 .68nn .37(n) .12 .10 .33 .31 .16 Mdn .89 .50 .48 .01 a Correlations with aggregate NEO-PI-R facet scores and NCS scales: Form R (observer rating data, N 5 24) from McCrae, Terracciano, and 79 Members (2005); Form S (self-report data, N 5 16) from McCrae (2002) and McCrae and Terracciano (2008); NCS data (N 5 22) from McCrae et al. (2007a). n po.05, nnpo.01, nnnpo.001, one-tailed. (n)Significant as one-tailed test in the wrong direction. 828 McCrae, Terracciano, De Fruyt, et al. ations of assessed Vulnerability and Compliance with corresponding national stereotypes and the negative correlation of Warmth with its stereotype replicate findings in observer rating data on adults but not in self-report data (Terracciano et al., 2005). Otherwise, these data are consistent with the findings of Terracciano and colleagues, who reported no association of assessed personality with national stereotypes. Validity of Profiles Within Cultures Table 4 provides data on comparisons of the 30-facet profiles within each culture. As in previous research, means for each facet were first standardized across the set of cultures used in each analysis; intraclass correlations were then calculated across the 30 facets by the double-entry method (see Griffin & Gonzalez, 1995). Comparing APPOC data to adult Form R data (first data column), significant profile agreement was found for 18 cultures (75.0%), including 6 of 8 cultures not included in the earlier PPOC comparison (McCrae, Terracciano, & 79 Members, 2005). Comparing APPOC data to adult Form S data (third data column), agreement was found for 9 of 16 cultures (56.3%). The magnitude of cross-study agreement was not related to data quality or n of targets in APPOC. The fifth data column of Table 4 reports ICC values for profile agreement with national character stereotypes for 22 cultures. Significant positive correlations were found for Argentina and Turkey, whereas significant negative correlations—contradicting the hypothesis of veridical stereotypes—were found for Australia, the Czech Republic, France, Hong Kong, and Peru. None of these correlations replicated findings reported by Terracciano and colleagues (2005), and the median intraclass correlation was .01. These analyses confirm that national character stereotypes in general do not reflect mean personality trait levels. The second, fourth, and sixth data columns of Table 4 report a second measure of profile agreement, rc (Cohen, 1969). Intraclass correlations are sensitive to the shape and relative elevation of profiles, but they do not take into account the direction of scoring. A profile that included measures of Introversion would look quite different from one that included measures of its polar opposite, Extraversion, and would generally yield different ICC values, but it would contain the same information. Cohen’s rc is invariant over the 829 Culture-Level Traits Table 4 Agreement of Adolescents’ NEO Personality Inventory-3 Profiles With Adults’ Revised NEO Personality Inventory Profiles and National Character Survey Scales Adult NEO-PI-R Form R Culture ICC Argentinaa Australiaa Chilea Croatia Czech Republic Estonia France Hong Kong Islamic Republic of Irana Japan Malaysia People’s Republic of China Peru Poland Portugal Puerto Ricoa Russia South Korea Serbia Slovakiaa Thailanda Turkey Ugandaa United States .43nn .45nn .24 .59nnn .13 .58nnn .65nnn .47nn .04 Mdn Form S rc NCS ICC rc ICC rc .43nn .47nn .51nn .63nnn .13 .59nnn .65nnn .58nnn .05 — — — .25 .53nn .84nnn .54nn .65nnn — — — — .26 .57nnn .85nnn .56nnn .70nnn — .39n .34(n) .10 .04 .33(n) .18 .37(n) .40(n) — .40n .32(n) .11 .06 .30 .18 .39(n) .34(n) — .77nnn .72nnn .48nn .78nnn .72nnn .58nnn .47nn .65nnn .03 .48nn .66nnn .06 .24 .18 .23 .25 .19 .30 .15 .35n .20 .41n .34(n) .52nn .51nn .56nn .42n .58nnn .58nnn .67nnn .15 .38n .56nnn .43nn .27 .52nn .51nn .64nnn .46nn .64nnn .61nnn .69nnn .23 .33n .28 — .12 .51nn .27 — — .16 — .51nn .24 .37n .42nn — .20 .52nn .30 — — .21 — .56nnn .54(nn) .27 .14 .26 .06 .18 .12 .17 — .42n .07 .17 .52(nn) .29 .07 .21 .03 .24 .12 .13 — .43nn .11 .18 .48 .54 .40 .45 .01 .07 Note. N 5 30 facets. ICC 5 intraclass correlations (double-entry method). rc 5 Cohen’s r. Form R (observer rating) data from McCrae, Terracciano, & 79 Members (2005). Form S (self-report) data from McCrae (2002) and McCrae and Terracciano (2008). NCS 5 National Character Survey; NCS data from McCrae et al. (2007a). a Not included in previous studies of culture-level convergent validity. n po.05, nnpo.01, nnnpo.001, one-tailed. (n),(nn)Significant as one-tailed test in the wrong direction. 830 McCrae, Terracciano, De Fruyt, et al. direction of scale scoring because each scale’s reflection around the mean (in this case, T 5 50) is also included in the profile. It is sensitive to both the shape and the absolute elevation of the two profiles. Reanalysis of data on profile agreement across observers (McCrae, 2008) showed that rc is as effective as ICC in identifying matched versus mismatched data. Table 4 reports rc values and provides further support for the view that aggregate adult personality scores, but not national character stereotypes, are related to aggregate adolescent scores. Adolescent profiles for Chile and Portugal are significantly related to adult profiles when rc is used as the measure of profile agreement. Geographical Patterns Associations among aggregate personality profiles were examined using nonmetric Multidimensional Scaling (MDS) to see if profile similarity was associated with geographical patterns. Analysis followed the methods used in previous research (Allik & McCrae, 2004; McCrae, Terracciano, & 79 Members, 2005): Aggregate scores for the 24 cultures were standardized across cultures, a distance matrix was calculated based on (1–Pearson r) across the 30 NEO-PI-3 facets, coordinates for two MDS dimensions were derived (StatSoft, 1995), and these coordinates were correlated with factor scores and rotated to maximize the correlations of the vertical axis with Neuroticism (r 5 .75) and the horizontal axis with Extraversion (r 5 .83). The standardized stress value for the two-dimensional solution was .21, which suggests the need for additional dimensions (five dimensions showed a stress value of .06), but because our intent was to compare these results to previous MDS results, we report the two-dimensional solution. Figure 1 displays results. As in previous studies, Western cultures are found on the right (extraverted) side of the plot, non-Western cultures on the left. French, Czechs, Argentines, and Hong Kong Chinese are again found at the top of the figure and Estonians and Mainland Chinese at the bottom. There is one notable difference: Russian adolescents are located in the bottom right of the figure and thus appear to be more adjusted and extraverted than older Russians (McCrae, Terracciano, & 79 Members, 2005). Resemblance to the MDS analysis of PPOC data can be quantified by correlating the coordinates across the two studies. Agreement was strong for the horizontal axis, r 5 .71, N 5 24, po.001; for the vertical axis, 831 Culture-Level Traits 1.5 °HK Chinese 1 French °Japanese °Chileans 0.5 °Australians °Argentines °Czechs °Thais °Portuguese °Turks °Malays °Ugandans 0 °Puerto Ricans °Peruvians –0.5 °Iranians °S. Koreans ° –1 –1.5 –1.5 ° Croatians °Slovaks °Poles Chinese °Serbians ° °Americans °Estonians –1 –0.5 0 0.5 1 1.5 Figure 1 Multidimensional scaling plot of 24 cultures based on a distance matrix of (1–Pearson r) for the 30 NEO Personality Inventory-3 facet scores, standardized across cultures. The vertical axis is maximally aligned with Neuroticism and the horizontal axis with Extraversion. HK Chinese 5 Hong Kong Chinese. S. Koreans 5 South Koreans. however, it was r 5 .34, ns. Omitting the Russians, the correlation for the vertical axis increased to r 5 .51, N 5 23, po.05. Culture-Level Factor Structure As in previous studies, principal component analyses at the culture level were undertaken using mean values from subsamples in order to obtain a reasonably large number of cases. For the present study, 108 subsamples were used, representing older and younger adolescent boys and girls from each of the 27 sites. Results after Procrustes rotation are reported in Table 5. Even in this small sample, the normative, adult, individual-level structure is reasonably replicated for Neuroticism, Extraversion, Agreeableness, and Conscientiousness 832 McCrae, Terracciano, De Fruyt, et al. Table 5 Culture-Level Factor Structure of NEO-PI-3 Facet Scales Procrustes-Rotated Principal Component A C VCa .03 .03 .04 .10 .23 .29 .22 .19 .19 .06 .33 .10 .19 .17 .01 .07 .49 .28 .90b .91c .91c .95b .94b .89c .64 .70 .53 .44 .51 .53 .27 .43 .39 .60 .16 .45 .41 .20 .21 .35 .51 .26 .20 .00 .39 .16 .17 .03 .91c .84 .96b .76 .88c .87c .38 .31 .27 .24 .14 .19 .65 .03 .59 .03 .09 .51 .01 .46 .44 .73 .02 .13 .09 .38 .27 .11 .18 .22 .34 .58 .17 .17 .50 .18 .49 .76 .91c .87c .24 .04 Trust Straightforwardness Altruism Compliance Modesty Tender-Mindedness .05 .07 .00 .05 .03 .19 .46 .17 .71 .47 .10 .41 .15 .26 .18 .08 .14 .18 .40 .59 .29 .48 .47 .46 .03 .01 .16 .23 .18 .52 .82 .80 .90c .71 .89c .65 Competence Order Dutifulness Achievement Striving Self-Discipline Deliberation .26 .02 .02 .10 .13 .09 .39 .13 .02 .15 .04 .38 .11 .22 .15 .24 .12 .02 .01 .28 .40 .11 .22 .41 .75 .78 .83 .88 .87 .72 .91c .79 .94b .99b .92c .96b Factor congruenced .93b .88b .47 .86b .88b .83b NEO-PI-3 Facet Scale N E N1: N2: N3: N4: N5: N6: Anxiety Angry Hostility Depression Self-Consciousness Impulsiveness Vulnerability .83 .80 .81 .77 .48 .77 .11 .01 .14 .20 .29 .13 E1: E2: E3: E4: E5: E6: Warmth Gregariousness Assertiveness Activity Excitement Seeking Positive Emotions .12 .03 .24 .18 .18 .06 O1: O2: O3: O4: O5: O6: Fantasy Aesthetics Feelings Actions Ideas Values A1: A2: A3: A4: A5: A6: C1: C2: C3: C4: C5: C6: O Note. These are principal components from 108 subsamples targeted to the American normative factor structure. Loadings greater than .40 in absolute magnitude are given in boldface. aVariable congruence coefficient; total congruence coefficient in the last row. bCongruence higher than that of 99% of rotations from random data. c Congruence higher than that of 95% of rotations from random data. dCongruence with American normative factor structure. Culture-Level Traits 833 factors (congruence 4.85; Lorenzo-Seva & ten Berge, 2006), and 26 of the 30 facets show loadings above .40 on the intended factor. Comparisons to randomly permuted data from an earlier study of the NEO-PI-R (McCrae, Zonderman, Costa, Bond, & Paunonen, 1996) suggested that 4 factor congruences and 19 of the 30 variable congruence coefficients exceeded chance values. However, the Openness factor is clearly not replicated. Three of its intended facets are unrelated to the factor, and three of the definers of the observed factor are facets of Extraversion. There appear to be two reasons for these deviations from the usual structure. First, Openness to Ideas loads on the Conscientiousness factor. This finding at the culture level is expected, given that, in these data, Openness to Ideas loads strongly (.48 to .51) on the Conscientiousness factor at the individual level (De Fruyt et al., 2009). Although sometimes seen in self-reports (Hřebı́čková, 2008), this phenomenon appears chiefly in observer ratings of adolescents. Costa et al. (2008) reported a loading of .39 for Openness to Ideas on the Conscientiousness factor when middle-school-aged respondents rated another child of the same age, but only .24 when they rated themselves. In observer ratings of college students and adults (McCrae, Terracciano, & 78 Members, 2005), the loading of O5: Ideas on Conscientiousness is .31; in self-reports from adults (Costa & McCrae, 1992), it is .16. It thus appears that high loadings of O5: Ideas on Conscientiousness are a joint function of method and target age: When outside observers assess intellectual curiosity in school children, they are apt to confuse it with academic success, which is also associated with Conscientiousness. Teachers, for example, attribute academic self-esteem to students they rate as high in both Conscientiousness and Openness (Graziano & Ward, 1992). By contrast, when American adolescents rate themselves, they can distinguish between intrinsic intellectual interest and academic achievement orientation (Costa et al., 2008). The Openness factor is also poorly defined because O1: Fantasy and O6: Values have their major loadings on the Extraversion factor. This is not unique to analyses of adolescents or of observer ratings; instead, it appears to be a culture-level phenomenon. Modern Western nations tend to be high on Extraversion, and they also tend to embrace such self-expressive values as imagination and tolerance (Inglehart, 1997). Raters from such cultures are thus more likely to describe their compatriot targets as high both in Extraversion and in traits like Fantasy and Values. As data simulations show (McCrae & Terracciano, 2008), 834 McCrae, Terracciano, De Fruyt, et al. the effect is to broaden the culture-level Extraversion factor to represent something more like individualism. This is, however, only part of the story. In adult data from PPOC, Openness to Fantasy and Values had joint loadings on the culturelevel Extraversion and Openness factors (McCrae, Terracciano, & 79 Members, 2005), whereas Table 4 shows no loadings at all for these facets on the Openness factor. At least with regard to Openness to Values, this may be because young adolescents do not yet have a clearly defined ideology, leading to very low internal consistency for this facet (Costa et al., 2008; De Fruyt et al., 2009). Conclusion The present study, using college students’ ratings of adolescents aged 12 to 17 on a modified version of the NEO-PI-R in 24 cultures, provides further evidence for three conclusions. First, there is general agreement about characterizations of cultures based on personality assessments of individuals: Adult self-reports, observer ratings of adults, and now observer ratings of adolescents all show similar patterns, whether one considers each trait across all cultures or the profile of all traits within each culture or the clustering of culture profiles in multidimensional space. Second, there is no consistent agreement between these aggregate characterizations of cultures and the corresponding collective beliefs about traits of the ‘‘typical’’ culture member: National character stereotypes again appear to be largely unfounded. Finally, there is further evidence that the culturelevel factor structure differs from the individual-level structure with regard to the Extraversion factor. In ratings of young adolescents, as in observer ratings and self-reports of college students and adults, Openness to Fantasy and Values, Competence, and low Compliance are associated with the Extraversion factor, but only at the culture level. This robust finding requires a culture-level explanation. The repeated finding that national character stereotypes are unrelated to assessed aggregate personality has seemed counterintuitive to some psychologists (e.g., Perugini & Richetin, 2007), but it makes sense if national stereotypes are, in fact, determined chiefly by such nonpsychological features as a nation’s wealth or mean temperature (McCrae et al., 2007a). This finding is not of merely academic interest: Beliefs about national character can have an important influence on political and social views and affect both ethnic and international relations. 835 Culture-Level Traits Psychologists should educate the public on the dangers of stereotypic thinking, especially with regard to national stereotypes. At the same time, they need to conduct more research on the origins of these beliefs and how they might be changed (Terracciano & McCrae, 2007). Other findings from the present study pose more purely intellectual challenges. At the individual level, aggregating facets to define broad domains generally leads to more reliable and valid scores. For example, among adolescents ages 14–20, the median cross-observer correlation for the five NEO-PI-3 domains is .53, whereas the median for the 30 facets is only .43 (McCrae, Costa, et al., 2005). That pattern is reversed at the culture level: In the present study, the median Form R cross-study correlation is .37 for the five domains but .50 for the 30 facets. It is possible that this finding is a fluke, attributable to the small number of cultures examined. Until that can be established, however, it would appear wise to conduct cross-cultural comparisons of aggregate traits chiefly at the facet level: We can have more confidence in the claim that a given culture is high in Altruism or Deliberation than that it is high in Agreeableness or Conscientiousness. Studies on the cultural origins or effects of personality traits should target specific facets. 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Journal of Personality and Social Psychology 2010, Vol. 99, No. 5, 870 – 882 © 2010 American Psychological Association 0022-3514/10/$12.00 DOI: 10.1037/a0020963 How People See Others Is Different From How People See Themselves: A Replicable Pattern Across Cultures Jüri Allik, Anu Realo, and René Mõttus Peter Borkenau University of Tartu Martin-Luther-Universität Halle–Wittenberg Peter Kuppens Martina Hřebı́čková University of Melbourne and Katholieke Universiteit Leuven Academy of Sciences of the Czech Republic Consensus studies from 4 cultures—in Belgium, the Czech Republic, Estonia, and Germany—as well as secondary analyses of self- and observer-reported Revised NEO Personality Inventory (NEO PI-R) data from 29 cultures suggest that there is a cross-culturally replicable pattern of difference between internal and external perspectives for the Big Five personality traits. People see themselves as more neurotic and open to experience compared to how they are seen by other people. External observers generally hold a higher opinion of an individual’s conscientiousness than he or she does about him- or herself. As a rule, people think that they have more positive emotions and excitement seeking but much less assertiveness than it seems from the vantage point of an external observer. This cross-culturally replicable disparity between internal and external perspectives was not consistent with predictions based on the actor– observer hypothesis because the size of the disparity was unrelated to the visibility of personality traits. A relatively strong negative correlation (r ⫽ ⫺.53) between the average self-minus-observer profile and social desirability ratings suggests that people in most studied cultures view themselves less favorably than they are perceived by others. Keywords: personality ratings, internal and external perspective, cross-cultural comparison, self-enhancement, the actor– observer hypothesis In addition to the consistent pattern of covariation among personality traits, several other surprisingly universal features of personality have been discovered. For example, it was found that in almost every culture, women reported themselves to be higher in Neuroticism, Agreeableness, Warmth, and Openness to Feelings, whereas men were higher in Assertiveness and Openness to Ideas (Costa, Terracciano, & McCrae, 2001; Feingold, 1994; Schmitt, Realo, Voracek, & Allik, 2008). Quite unexpectedly, these sex differences in personality increased with higher levels of human development, including a long and healthy life, equal access to knowledge and education, and economic prosperity (Costa et al., 2001; Schmitt et al., 2008). What is particularly remarkable is that the cross-cultural convergence between different studies on sex differences in personality was demonstrably stronger and more replicable than the convergence between the mean levels of the traits themselves (Schmitt et al., 2007, 2008). Another feature that seems to easily transcend cultures is age difference. In every human society explored so far, individuals become less extraverted and open to new experiences and more agreeable and conscientious with age (Costa et al., 2000; McCrae, Costa, Hřebı́čková et al., 2004; McCrae et al., 1999; Srivastava, John, Gosling, & Potter, 2003). In spite of some specific features characterizing particular cultures, the general pattern of difference between younger and older individuals is stable and highly replicable across the 30 personality traits measured by the NEO PI-R questionnaire, even across dissimilar cultures (Allik et al., 2009). In light of the discovery of these universal features, it is surprising that there is no consensus about the systematic differences More than a decade ago, McCrae and Costa (1997) proposed the hypothesis that the pattern of covariation among basic personality traits is a universal feature of the human species. Several recent large-scale cross-cultural studies have supported this idea, showing that a common five-factor structure of personality traits can be found in all languages and cultures examined so far (McCrae, Terracciano, & 78 Members of the Personality Profiles of Cultures Project, 2005; Schmitt, Allik, McCrae, & Benet-Martinez, 2007). However, the five-factor model is not the only replicable personality structure because Eysenck’s three-factor (van Hemert, van de Vijver, Poortinga & Georgas, 2002) and psycholexical six-factor (Lee & Ashton, 2008) structures have also been replicated in many cultures. Jüri Allik, Anu Realo, and René Mõttus, Department of Psychology, University of Tartu, Tartu, Estonia; Peter Borkenau, Institut für Psychologie, Martin-Luther-Universität Halle–Wittenberg, Halle, Germany; Peter Kuppens, School of Behavioral Science, University of Melbourne, Melbourne, Australia, and Department of Psychology, Katholieke Universiteit Leuven, Leuven, Belgium; Martina Hřebı́čková, Institute of Psychology, Academy of Sciences of the Czech Republic, Prague, the Czech Republic. This project was supported by Estonian Ministry of Science and Education Grant SF0180029s08 and Estonian Science Foundation Grant ESF7020 to Jüri Allik. The Czech participation was supported by Grant Agency of the Czech Republic Grant P407/10/2394. We are very thankful to Robert R. McCrae for his helpful comments and suggestions. Correspondence concerning this article should be addressed to Jüri Allik, Department of Psychology, University of Tartu, Tiigi 78, Tartu 50410, Estonia. E-mail: juri.allik@ut.ee 870 SELF VERSUS OTHER RATINGS IN PERSONALITY JUDGMENT between how people see others and how they see themselves, despite some theories that have been proposed. Social psychologists, for example, have invested a considerable amount of energy in the promotion of the idea that there is a fundamental disparity between the way people perceive themselves and the way they are perceived by others (Jones & Nisbett, 1971; Nisbett, Caputo, Legant, & Marecek, 1973; Watson, 1982). This disparity is believed to originate from an inevitable asymmetry between internal and external viewpoints: People are immersed in their own sensations, emotions, and cognitions at the same time that their experience of others is dominated by what can be observed externally (Pronin, 2008). Our aim in the current article is to systematically analyze the differences between how people judge their personality and how their personality is judged by others across different cultures. Analyzing the existing literature, it is possible to distinguish at least two principal mechanisms by which personality descriptions made from the perspective of the first person might be systematically different from those made from the vantage point of the third person. Different Information As an external observer cannot see the target person in all situations, it is almost inevitable that the information possessed by the target person must be different from the information that is available to an external observer. Even when it comes to information that is equally available to the target person and the external observer, it is possible that the target person could ignore some information that was attended to by the external observer in making his or her judgment. One consequence of this, as argued by Jones and Nisbett (1971), is the actor– observer asymmetry in attribution, that is the pervasive tendency for actors to attribute their behavior to external (situational) causes and for observers to attribute the same behavior to internal causes (dispositional qualities) of the actor. As a further evidence of the divergent perspectives of the actor and observer, among other things observers were found to ascribe more personality traits to other people than to themselves (Nisbett et al., 1973). However, a recent meta-analysis involving more than 170 studies established that the actor– observer hypothesis— counter to what textbook descriptions and commonly held beliefs suggest—is neither firmly established nor robust and that evidence for it is surprisingly limited (Malle, 2006). For example, the established beliefs that people ascribe more personality traits to others than to themselves (Sande, Goethals, & Radloff, 1988) and that they perceive more complexity in their behavior than that of others (Locke, 2002) were not confirmed in these later studies. It is not self-evident how the principle of “more personality traits” can be operationalized for responses to a fixed list of items. Personality psychology questionnaires are constructed on the assumption that all personality traits are applicable to all individuals; respondents thus need to assess themselves or the others on each of these traits. Despite the fact that considerable self– other agreement can be obtained for all personality traits, the convergence of judgments either between self and observer or between different observers is considerably stronger on some traits than on the other traits (Colvin, 1993; Funder, 1999; John & Robins, 1993). The most likely cause for these differences is the type and amount of information available to the self and the external observer. In 871 particular, results suggest that traits pertaining to extraversion are revealed relatively directly in social behavior and, therefore, are easy to judge, whereas traits pertaining to neuroticism are less visible and, so, are judged less accurately (Funder & Colvin, 1988; Funder & Dobroth, 1987). It has also been noticed that traits reflecting affective states are more difficult to judge than are traits which manifest themselves in overt behaviors (Spain, Eaton, & Funder, 2000; Watson, Hubbard, & Wiese, 2000). Lack of judgeability or visibility of traits can lead, in principle at least, to a disparity between self- and observer-ratings (John & Robins, 1993). It is not difficult to imagine, for example, why people see themselves as more neurotic compared to how they are seen by other people. Neuroticism reflects, to a considerable extent, inner states of an individual that are not necessarily accessible to an external observer. Provided that these externally unobservable instances of neurotic tendencies influence people’s self-evaluation of neuroticism, the expected outcome is a disparity between selfand observer-ratings. Thus, one possible scenario is that the self– other disparity is less pronounced on more observable traits—as operationalized using rank-order correlations between self- and observer-ratings—whereas the disparity between rater’s perspectives is more manifest in less externally observable traits. Self-Enhancement One of the most pervasive explanations for the disparity between self- and observer-perceptions is that people are systematically engaged in self-enhancement: They view themselves more favorably than they view others (Kenny, 1994; Kwan, John, Kenny, Bond, & Robins, 2004). Although some cross-cultural differences seem to exist (Heine & Hamamura, 2007; Heine & Renshaw, 2002), a recent cross-cultural study demonstrated that in all 56 cultures studied, people’s mean value of self-esteem was above the scale neutral point (Schmitt & Allik, 2005), suggesting that most people are motivated to maintain a positive view of themselves. Many studies have demonstrated that college students rate their own personality traits in more socially desirable terms than they do when rating the “average college student” (Alicke, 1985; Alicke, Klotz, Breitenbecher, Yurak, & Vredenburg, 1995; J. Krueger, 1998). However, the effect of this unrealistically positive view of themselves disappears, or is considerably reduced, when a specific person, not an average college student, is assessed (Alicke et al., 1995). Nevertheless, it is possible that the mean difference between self- and observer-ratings reflects, to a certain degree, the social desirability of personality traits. Thus, one expected outcome of self-enhancement is that mean self-rating scores are higher than are the observer-ratings on those personality traits believed to have higher social desirability. However, all these predictions about systematic differences between how people see others’ and their own personality traits still need to take into account the fact that normative self-rated personality mean scores converge almost perfectly with normative observer-rated mean scores. For example, the correlation between the mean profiles of the adult S-Form (self-ratings) and R-Form (observer-ratings) presented in the NEO PI-R Professional Manual is .94 (Costa & McCrae, 1992, Table B1 and Table B2). The largest mean difference is 3.3 raw-score points on the C6: Deliberation subscale, which in T-scores is 7.6 units higher for observerthan self-ratings. The average absolute self– other rating difference ALLIK ET AL. 872 across all 30 NEO PI-R subscales is 2.9 T-score units. Thus, given the high correlation and small difference in mean levels, there is little room for disparity between the perspectives from which personality is described. Looking at how similar self-rating and observer-rating normative profiles are, one has to conclude that the effect caused by self– other asymmetry must be small and, if at all, reliably detectable only when sufficiently large samples are used. Examination of the global distribution of aggregated self-rated personality profiles across cultures has revealed a regular pattern, with a clear contrast between European and American cultures on the one hand and Asian and African cultures on the other (Allik & McCrae, 2004). The global distribution of observer-rated personality traits generally followed the same pattern (McCrae, Terracciano, & 79 Members of the Personality Profiles of Cultures Project, 2005). However, it is possible that when there is an asymmetry between self- and observer-perspectives, it also displays itself at the aggregate level. If there is a systematic disparity in the perspective from which personality traits are described at culture-level personality scores as well, it will constitute another cross-cultural personality universal in addition to the relatively invariant pattern of sex and age differences. Aims of the Study Since many theoretical constructions assume a fundamental asymmetry between self- and observer-perspectives, the main goal of this study is to examine whether there is a replicable pattern of differences between self-rated and observer-rated personality traits which transcends different languages and cultures. There are two principal study designs available: an individual design, in which the targets of self-reports and observer reports are the same and in which the question is whether the mean difference between these two is positive or negative; and a culture-level design, in which different samples from the same culture are compared. The former is more powerful because targets serve as their own controls with regard to trait level, but the latter is also informative since it helps to test the robustness of the phenomenon. We used both designs. In Study 1 we compared data from four European languages and cultures—in Belgium, Czech Republic, Estonia, and Germany— where personality traits of participants were judged by themselves and by one or more observers. To many readers it may come as a surprise that the number of studies with consensual validation between self- and observer-ratings is rather limited (McCrae, Costa, Martin et al., 2004). Even so, consensus studies are the most powerful source to test whether how people see others is different from how people see themselves. Following the idea of the actor– observer hypothesis, we tested whether self– observer discrepancy is larger in less visible traits. Additionally, separate groups of judges from all four of the countries also rated the social desirability of each of the NEO-PI-R items. On the basis of this, it is possible to compare the difference between self- and observerratings with the social desirability of personality traits. Self-rated scores are expected to be higher than observer-rated scores on those personality traits considered higher on social desirability. Study 2 is devoted to analyses of published cross-cultural data sets about self- and observer-ratings of personality with the NEO PI-R. The main goal in Study 2 is to establish how robust are the asymmetries between self- and observer-ratings that were found in the first study, that is how well they generalize across languages, cultures, and different levels of analyses. Study 1 Samples Belgian sample. Flemish data were collected from 345 target participants (270 women and 75 men) who were psychology students at the Katholieke Universiteit Leuven and who, as a course requirement, rated their own personality with the Dutch version of the NEO PI-R (Hoekstra, Ormel, & DeFruyt, 1996). They also recruited a well-acquainted person (n ⫽ 345; 190 women, 112 men, and 43 did not specify sex), either a relative or a friend, who rated their personality using the observer-report form of the same instrument. The mean age of targets was 18.4 (SD ⫽ 3.0) years. The mean age of external raters was 29.5 (SD ⫽ 13.7) years. Czech sample. The Czech sample included 811 targets (330 men, 481 women) who were recruited in a series of studies (McCrae, Costa, Martin et al., 2004). They ranged in age from 14 years to 83 years, with a mean age of 35.7 years (SD ⫽ 14.2 years). Peer-ratings were provided by 909 raters (377 men, 532 women) aged 14 – 83 years (M ⫽ 35.8 years; SD ⫽ 14.3 years) who participated in one of two research designs. In the self– other agreement studies (N ⫽ 615), each target provided a self-report and was rated by one informant. In the consensus study, 195 targets (85 men and 110 women aged 17–77 years; mean age 36.4, SD ⫽ 15.2) provided a self-report and were each rated by three informants. All participants used the Czech version of the NEO PI-R questionnaire (Hřebı́čková, 2002). Estonian sample. Estonian data came from two already published studies. The first subsample consisted of 218 Estonianspeaking participants (180 women and 38 men; mean age 22.3 years, SD ⫽ 5.2) who answered the NEO PI-R questionnaire, which was accompanied by a standard instruction to describe themselves honestly and accurately (Konstabel, Aavik, & Allik, 2006). They were also asked to provide two peer-reports (n ⫽ 436) from their acquaintances, relatives, or close friends. The Estonian version of the NEO PI-R (Kallasmaa, Allik, Realo, & McCrae, 2000) was completed voluntarily; some students studying psychology received an extra credit towards the fulfillment of their course requirements. The second Estonian subsample consisted of 154 participants (53 men and 101 women; mean age 43.9 years, SD ⫽ 17.6) who were described by one or two judges (Mõttus, Allik, & Pullman, 2007). The sample of judges (n ⫽ 308) included 203 women, 67 men, and 38 participants who did not report their gender. The mean age of the judges was 38.2 (SD ⫽ 15.9) years. Both targets and judges used the Estonian version of the EE.PIP-NEO (Mõttus, Pullmann, & Allik, 2006), which has a facet-structure identical to the NEO PI-R but was designed to be linguistically simpler, containing shorter and grammatically less-complex items. German sample. Participants were 304 students (169 women, 134 men, 1 not reporting sex) at a German university, of whom only 3 studied psychology (Borkenau & Zaltauskas, 2009). Their mean age was 23.38 (SD ⫽ 2.68) years, ranging from 18 years to 35 years. They received 45 Euro for their participation and were recruited in 76 groups, each comprising 4 persons who all knew SELF VERSUS OTHER RATINGS IN PERSONALITY JUDGMENT each other well. First, the participants described the 3 other group members on 30 bipolar adjective scales; these, however, are not relevant to the present study. Next, each 4-person group was split into two dyads, and all participants described themselves and the other dyad member on several personality inventories including the German version of the NEO PI-R (Ostendorf & Angleitner, 2004). It is important to notice that in all four samples observers knew their targets well, being either their close relatives or friends. Measure of Social Desirability In order to develop a social desirability index for the NEO PI-R, the questionnaire items were assessed by 100 Czech judges (43 men and 57 women, mean age 40.5 years, SD ⫽ 15.1), 88 Estonian judges (24 men and 64 women, mean age 37.6 years, SD ⫽ 12.7), 30 Flemish judges (12 men, 16 women, 2 did not report their sex, mean age 20.3 years, SD ⫽ 2.0), and 20 German judges (9 men and 11 women, mean age 23.8 years, SD ⫽ 3.0), who independently rated the social desirability of each of the 240 NEO PI-R items. The instruction stated, Descriptors of people often contain evaluative information. Some personality characteristics are considered more desirable, receiving approval from other people, whereas others are undesirable. If someone agrees strongly with this item, does this present that person in a favorable or unfavorable light, or is agreeing with this item neutral as regards to others’ approval? Ratings were made on a 7-point Likert scale, ranging from extremely undesirable (⫺3) to extremely desirable (⫹3), with zero as a neutral point. Estonian desirability ratings were reported in a previous study (Konstabel et al., 2006). As the pairwise correlations between social desirability profiles of cultures were suffi- ciently high (from r ⫽ .80 to r ⫽ .93), we used an unweighted average of these four groups of judges. Results and Discussion The Flemish, German, Estonian, and Czech mean difference profiles (self-minus-observer) converted to z -scores (the mean difference divided by the average standard deviation) are shown in Figure 1. Even a brief inspection reveals that all four difference profiles are very similar. Indeed, the pairwise correlations between these four profiles range from .77 (Belgium and Germany) to .88 (Estonia and the Czech Republic) with the mean r ⫽ .83 (all highly significant). All four profiles are also strongly correlated with the difference profile of the U.S. adult normative data (S-Form minus R-Form) given in the NEO PI-R Professional Manual (Costa & McCrae, 1992, Table B1 and Table B2). The correlations are .85, .79, .75, and .63 for Belgium, the Czech Republic, Estonia, and Germany, respectively. In all four cultures (in addition to the United States), people perceive themselves as more neurotic and more open than they are seen by others. They also perceive their level of conscientiousness as lower than how they are seen from the vantage point of their observers. Particularly, people see themselves as less competent (C1), self-disciplined (C5), and altruistic (A3) than they are perceived by others. At the same time they think that they are more than others open to fantasy (O1) and ready to reexamine social, political, and religious values (O6). Even this short list of disparities suggests that in general, people do not view themselves more favorably than how they are viewed by others. For a more formal test we analyzed the combined sample of 1,768 targets, pooled from the four separate samples. Before pooling, all scores were normalized within the country dataset. Belgium Estonians Germans Czechs 0.4 0.0 -0.4 -0.8 -1.2 N1:Anxiety N2:Angry Hostility N3:Depression N4:Self-Consciousness N5:Impulsiveness N6:Vulnerability E1:Warmth E2:Gregariousness E3:Assertiveness E4:Activity E5:Excitement Seeking E6:Positive Emotions O1:Fantasy O2:Aesthetics O3:Feelings O4:Actions O5:Ideas O6:Values A1:Trust A2:Straightforwardness A3:Altruism A4:Compliance A5:Modesty A6:Tender-Mindedness C1:Competence C2:Order C3:Dutifulness C4:Achievement Striving C5:Self-Discipline C6:Deliberation Self-minus-observer Ratings (z-scores) 1.2 0.8 873 Figure 1. The mean difference profiles (self-minus-observer) for Belgium, Germany, Estonia, and the Czech Republic, converted to z-scores (the mean difference divided by the average standard deviation). Letters–number combinations are the NEO PI-R facet scale numbers. ALLIK ET AL. 874 Contrary to the concept of self-enhancement, the mean self-minusobserver profile was negatively correlated (r ⫽ ⫺.53, p ⫽ .003), with the profile of social desirability ratings (individual correlations were from r ⫽ ⫺.40, p ⫽ .026 to r ⫽ ⫺.62, p ⫽ .001 for Estonia and Belgium, respectively) suggesting that observerratings rather than self-ratings might be biased towards social desirability. Similarly, the normative U.S. adult self-minusobserver profile was negatively correlated with the profile of social desirability ratings (r ⫽ ⫺.45, p ⫽ .012). Figure 2 presents the average self-minus-observer profile for the four cultures studied in comparison with their average social desirability ratings on the items of the 30 NEO PI-R facet scales. Although these two profiles are not exact mirror images, their dissimilarity is obvious. In personality psychology, a correlation of ⫺.53 between two profiles is sufficiently high enough to speak of a substantial reverse link between social desirability and the disparity between perspectives. To test the prediction following from the actor– observer asymmetry that disparity is more pronounced on less visible traits, we computed the rank-order correlation between self-ratings and observer-ratings. The highest self– other agreement was found on the subscales E3: Assertiveness (.56), O2: Aesthetics (.56), and E5: Excitement Seeking (.55). A relatively low self– other agreement was found in the ratings of C1: Competence (.32), A1: Trust (.36), and N5: Impulsiveness (.38). The average self– other agreement was reasonably high (median ⫽ .43, p ⫽ .018). These values are comparable to typical self– observer agreement values obtained in previous studies using the NEO family questionnaires (Connolly, Kavanagh, & Viswesvaran, 2007; McCrae, Costa, Martin et al., 2004). Having obtained these self– other convergence values, we can ask whether the self– observer asymmetry is more pronounced on those personality traits on which people and their judges agree less. Figure 3 shows the correlation plot between self– other agreement and self-minus-observer differ- ence scores. There seems to be no systematic relationship between self– other agreement and the asymmetry in perception (r ⫽ .03, p ⫽ .88). Thus, on personality traits that are more visible or judge-able, there is no smaller disparity between selfand observer-ratings. In addition to social desirability and visibility, there are other potential explanations for the asymmetry between self- and observer-ratings. One of them could be a systematic age difference between targets and judges. College-age targets may have recruited adults to judge their personality, and adults may tend to endorse certain items differently compared with college-aged people. Indeed, this tendency was true for the Flemish and Estonian samples, but not for the German and a part of Czech samples in which the mean age of target and judges was identical. Thus, if the rater-age related explanation is true, we should expect a systematic difference in self– observer asymmetry between German and Czech samples on one side and Estonian and Flemish samples on the other side. There appears to be no such difference (see Figure 1), which speaks against the rater-age related explanation for the asymmetry. In addition, to see whether the observer-rated personality profile tends to be more adultlike, we compared the average values of the four self-minus-observer profiles displayed in Figure 1 with the mean age differences between adults and collegeage targets for 30 NEO PI-R facets. We took the latter values from the best observer-ratings database to date, results obtained from the international sample consisting of 50 cultures participating in the Personality Profiles of Cultures Project (McCrae, Terracciano, & 78 Members of the Personality Profiles of Cultures Project, 2005; McCrae, Terracciano, & 79 Members of the Personality Profiles of Cultures Project, 2005). Contrary to our prediction that adults may follow their age-specific response pattern even when rating college-aged people, the correlation between the two difference profiles— self– observer asymmetry and age-related differences 0.8 SELF-MINUS-OBSERVER RATINGS SOCIAL DESIRABILITY RATINGS 0.6 Z-scores 0.4 0.2 0.0 -0.2 -0.6 N1:Anxiety N2:Angry Hostility N3:Depression N4:Self-Consciousness N5:Impulsiveness N6:Vulnerability E1:Warmth E2:Gregariousness E3:Assertiveness E4:Activity E5:Excitement Seeking E6:Positive Emotions O1:Fantasy O2:Aesthetics O3:Feelings O4:Actions O5:Ideas O6:Values A1:Trust A2:Straightforwardness A3:Altruism A4:Compliance A5:Modesty A6:Tender-Mindedness C1:Competence C2:Order C3:Dutifulness C4:Achievement Striving C5:Self-Discipline C6:Deliberation -0.4 Figure 2. The average self-minus-observer profile for the four cultures studied in comparison with their average social desirability ratings on the Revised NEO Personality Inventory items. Letters–number combinations are the NEO PI-R facet scale numbers. SELF VERSUS OTHER RATINGS IN PERSONALITY JUDGMENT 875 0.60 E3 Correlation between self and observer ratings 0.55 O2 E5 E2 C2 O5 0.50 E4 E6 E1 O4 A4 N3 C5 0.45 C3 A5 N2 O1 C6 C4 N1 O6 0.40 A3 A2 N6 O3 A6 N5 N4 A1 0.35 C1 0.30 -0.8 -0.6 -0.4 -0.2 0.0 0.2 0.4 0.6 Self-minus-observer ratings (z-scores) Figure 3. Correlation plot between self– other agreement and self-minus-observer-difference scores (r ⫽ .03, p ⫽ .88). Letters-number combinations are the NEO PI-R facet scale numbers. (adults minus college age)—was positive (r ⫽ .64, p ⬍ .001), suggesting that the observer’s perspective was more characteristic to a younger person, not an older person. Thus, an external observer may tend to emphasize the younglike personality trait levels of their targets, whereas generally younger targets may tend to show themselves as more maturelike. Study 2 In spite of obvious language and cultural differences, all four studied countries are members of the European Union and have relatively high levels of human development and economic prosperity. Therefore, to make any claims concerning the universality of our findings in Study 1, we would need to have data from other regions of the World, including Asian and African countries. Disappointingly, the number of consensual studies between selfand observer-ratings done outside Europe and North America is in short supply. Fortunately, the NEO PI-R was translated into more than 40 languages and many researchers around the world have collected self-report data. In 2002, McCrae (2002) assembled self-report data that had been collected by other researchers using a variety of designs in 36 different cultures. Some cultures (e.g., Hong Kong) had only college-age respondents; some (e.g., Spain) had only adult data, and many had both. The ratio of men to women varied widely across cultures. In order to create overall culture scores that would be comparable across these diverse studies, McCrae (2002) first standardized each subsample using age- and gender-specific U.S. norms and then defined the overall culture T -score as the unweighted mean of all available subsamples. This strategy assumes that trait levels for a culture are generalizable across age and gender groups and that age and gender differences around the world are similar to those found in the U.S. norms. Both these assumptions were generally supported by the data (McCrae, 2002), and the validity of the resulting overall culture trait means was supported by their correlates (Allik & McCrae, 2004; McCrae & Terracciano, 2008). Unlike self-report data, the collection of the observer’s ratings has been much more systematic. During the Personality Profiles of Culture project, college students from 51 cultures identified an adult or college-aged man or woman whom they knew well and rated more than 12,000 targets using the R-form of the NEO PI-R (McCrae, Terracciano, & 78 Members of the Personality Profiles of Cultures Project, 2005; McCrae, Terracciano, & 79 Members of the Personality Profiles of Cultures Project, 2005). As there is a considerable overlap between the samples of nations for which aggregate self- or observer-rated personality scores are reported, it makes it possible to compare the self- and observer-rated personality traits at the aggregate national level. Provided that the pattern of difference between internal and external perspectives for the Big Five personality traits is pervasive, we could expect to observe it even if the targets of the self and observer’s ratings are not identical, to say nothing about other differences (such as age, sex and occupation) between the study designs. 876 ALLIK ET AL. Method The self-reported mean T-scores of the NEO PI-R subscales for 36 countries were published by McCrae (2002). An additional set of self-report data for Burkina Faso, Switzerland (Frenchspeaking), and Poland was published by McCrae and Terracciano (2008, Appendix C). In another large-scale project, observers’ ratings were collected from more than 50 different cultures using the R-Form of the NEO PI-R (McCrae, Terracciano, & 78 Members of the Personality Profiles of Cultures Project, 2005; McCrae, Terracciano, & 79 Members of the Personality Profiles of Cultures Project, 2005), meaning that standardized mean observer ratings for 51 cultures are now publicly available (McCrae & Terracciano, 2008). For 29 cultures, both self- and observer-reports are available. Although this overlapping set also contained data from Belgium, the Czech Republic, Estonia, and Germany, it is important to note that the data were different from what we used in Study 1. Because the published means in both cases are reported in T-scores, and the original data are not available, we converted them back to raw scores using the formula ([T-Score – 50] 䡠 SD)/10 ⫹ M, where M and SD are the mean and standard deviation of the U.S. adult or college age normative data (Costa & McCrae, 1992), dependent on the respective sample, and the average international sample data (McCrae & Terracciano, 2008) for self- and observer-ratings, respectively. It is important to note, however, that this back transformation from T-scores to raw scores is approximate, given that on the basis of T-scores alone it is impossible to reconstruct the exact scores for different sex and age groups. The reconstructed mean profile represents a hypothetical average person, without sex and age specification. To compute the self– other asymmetry index, we subtracted the mean score of the observer-ratings from the mean score of the self-ratings. In order to study correlation with societal-level indicators, we found the mean absolute difference between observer-ratings and self-ratings for each culture. This score showing the magnitude of the self-minus-observer differences was correlated with several indicators characterizing economic and social conditions. Gross domestic product (GDP). GDP at purchasing power parity in U.S. dollars, divided by the midyear population in 2006, were obtained from the Human Development Indices (2008). Life expectancy. Life expectancy at birth indicates the number of years a newborn infant would live if prevailing patterns of age-specific mortality rates at the time of birth were to stay the same throughout the child’s life (Human Development Indices, 2008). Human Development Index (HDI). The Human Development Index measures the level of human development by combining normalized measures of life expectancy, literacy, educational attainment, and GDP per capita for countries worldwide; the reported indices are for the year 2006 (Human Development Indices, 2008). Index of Shipping Difficulties. The Index of Shipping Difficulties is an indicator of required efforts and complications (border delays, fees, red tape, etc) met during shipping goods (World Development Report, 2009, Table A4). Days to Start Business. The goal of the Doing Business project was to provide an objective basis for understanding and improving the regulatory environment for business. We used the days required for starting a business as an index of the bureaucratic and legal hurdles an entrepreneur must overcome to incorporate and register a new firm. Data were retrieved from http://doingbusiness.org/ExploreTopics/StartingBusiness/ on June 29, 2009. Corruption Perception Index (CPI). The Corruption Perception Index ranks 180 countries by their perceived levels of freedom from corruption, as determined by expert assessments and opinion surveys. The CPI is compiled annually, and it was retrieved from the Transparency International homepage http:// www.transparency.org/policy_research/surveys_indices/cpi on June 29, 2009. Results and Discussion The mean differential profiles (self-minus-observer) of the 30 NEO PI-R subscales for the 29 countries or cultural groups are shown in Table 1. The average between-country profile correlation was .43, suggesting that the profiles of the self-minus-observer mean differences are rather similar. The last column in Table 1 shows how much the mean self-minus-observer profile of each country is similar to the average self-minus-observer profile of all 29 countries. As nearly all correlations are positive and significant (median ⫽ .67), a strong first principal component is suggested on which all individual profiles are loading. There is only one country-level self-minus-observer profile of the 29 that clearly deviates from the common shape—the Danish profile. We have no good explanation for why the Danish data are conspicuously different from other countries. The Danish self- and observerrating profiles alone are not outstanding from other profiles. The deviance of the self-minus-observer profile might reflect a real difference in perspective or might be a consequence of some measurement error, to say nothing about artifacts that could have been created by the back transformation from T-scores. Despite the ideosyncracy of the Danish self-minus-observer profile, we can still conclude that there is a remarkable cross-cultural similarity in the asymmetry of aggregate self- and observer-ratings. The critical issue, however, is how well the culture-level findings of Study 2 agree with the individual-level findings of Study 1 that were obtained with a more controlled design. The crossvalidity of the findings of the two studies is remarkable: Averaged self– observer difference profiles found in the four cultures investigated in Study 1 and the averaged self– observer difference profile of the 29 cultures investigated in Study 2 were correlated as highly as r ⫽ .80, p ⬍ .0001. Figure 4 presents the average self-minus-observer profile across 29 cultures, together with the average of four consensus studies (Study 1) and the differential profiles of the U.S. normative data (Costa & McCrae, 1992) for adults. All these three profiles are strongly correlated (from .70 to .82, p ⬍ .0001), suggesting that the average shape of the selfminus-observer profile remains essentially the same. Further comparison of data from the Study 1 and Study 2 shows that the average self– observer differences profile of the 29 countries is not correlated to the visibility of traits, operationalized as the rank-order correlation between self- and observer-ratings in the pooled data of four European countries (r ⫽ ⫺.04, p ⫽ .83). The lack of correlation is consistent with the findings in Study 1, further demonstrating that the self– observer asymmetry is probably not caused by the different amount of information available to self-raters and external observers. We also calculated the correla- SELF VERSUS OTHER RATINGS IN PERSONALITY JUDGMENT tion between the average self– observer differences profile of the 29 cultures and the mean social desirability ratings reported in Study 1. Similarly to the Study 1, the correlation was negative, but this time it was nonsignificant (r ⫽ ⫺.17, p ⫽ .37). Thus, assuming that social desirability is relatively universal, the lack of significant positive correlation shows that in culture-level analyses, self– observer asymmetry cannot be explained by selfenhancement. We were also interested whether the magnitude of the self-minusobserver differences is related to geographic, economic, and social indicators. In general, data of the 29 countries did not show significant correlation with these country-level indicators (Table 2), except a significant ( p ⫽ .008) negative correlation with the days required for starting a business. Provided that it is not a statistical fluke, it remains to elucidate why, in countries with low bureaucratic and legal hurdles that an entrepreneur must overcome to incorporate and register a new firm, people generally see others more differently from how they see themselves. General Discussion Mainstream social psychology, focusing on human inabilities, has been engaged in expanding the list of errors in judgment (J. I. Krueger & Funder, 2004). Although this study is also about the disagreement between two perspectives from which personality can be judged, its message is, overwhelmingly, about the remarkable accuracy of personality judgments. First, the level of self– other agreement (the median rank-order...
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Identity and Culture - Outline
Thesis statement: The paper requires to conduct a research on individualism versus
collectivism to explain the variety of cultural differences that make up a person’s identity.in
as provided in the course.
I.

Brief explanation of how a person’s identity may develop differently in a collectivistic versus
an individualistic culture.

II.
III.

How my own identity has been impacted by my culture (collectivistic or individualistic).
How my identity might differ if I were raised in the other type of culture.


Running head: IDENTITY AND CULTURE

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Identity and Culture
Name
Institution

IDENTITY AND CULTURE

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Identity and Culture

A person identity may develop differently in an individualistic culture as compared to a
collectivistic culture where one identifies mainly with self, the personal desires getting fulfilled
before those of a particular group. Also, it enhances the development of how an individual looks
after and takes good c...


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