Psychologists who emphasize trait perspectives expect a degree of cross-role consistency in all cultures, and consistency has been described as important for adjustment in Western theories of personality. At the same time, cultural psychologists have predicted less consistency and weaker relationships between consistency and adjustment in collectivistic cultures than in individualistic cultures (English & Chen, 2007
; Markus & Kitayama, 1994
; Suh, 2000; Triandis, 1995
). Both theoretical perspectives would be supported if (a) cross-role consistency was evident in all cultures, but less so in collectivistic cultures, and (b) the relative ability of trait consistency versus personal coherence, relationship harmony, and social appraisal to predict adjustment differed in the hypothesized manner in individualistic versus collectivistic cultures. Our results were generally supportive of trait perspectives, and cultural similarities were more prominent than cultural differences. Support for cultural psychology perspectives was more mixed, and there was some indication that lower consistency is more characteristic of East Asian cultures such as Japan than collectivistic cultures more generally. Strengths of the study included (a) our sampling of multiple individualistic and collectivistic cultures, including some less frequently sampled cultures; (b) our efforts to control for individual differences in scale use (i.e., response styles); and (c) the inclusion of multiple predictors of consistency and diverse measures of adjustment.
Cultural Similarities and Differences in Consistency
Consistent with trait perspectives, cross-role consistency was evident in a range of cultures. Across the six cultures, the mean cross-role correlations (.52 to .68) and mean general-specific role correlations (.41 to .68) were comparable to those reported by previous researchers (English & Chen, 2007
; Roberts & Donahue, 1994
; Suh, 2002
; Wood & Roberts, 2006
), and we examined a much broader range of cultures. The comparison with Suh’s (2002)
cross-cultural study is particularly intriguing, given the similarity of our results in Japan and the United States with his results in Korea and the United States. Suh (2002)
reported mean PCA values of 52.8% in Korea and 64.0% in the United States, while we obtained mean PCA values of 53.1% in Japan and 65.7% in the United States. From a “realist” perspective on person perception and from role identity theory, we can perhaps infer a degree of cross-role consistency across cultures in actual behavior as well. The realist perspective proposes that traits are real and observable (Funder, 1995
) and role identity theory proposes that role identities reflect, in part, one’s actual behaviors or traits in different roles (Wood & Roberts, 2006
). Nonetheless, cross-cultural comparisons of consistency in actual behavior will be needed to directly test this inference.
We also identified some cultural differences in cross-role trait consistency (Hypothesis 3). However, our results—particularly in combination with Suh’s (2002)
comparison of American and Korean samples—suggest that cultural psychology hypotheses regarding consistency may apply in a more limited manner to comparisons of the United States with East Asian cultures such as Japan and Korea, rather than to comparisons of individualistic and collectivistic cultures more generally. Indeed, although cultural psychology hypotheses have often been formulated in terms of the individualism-collectivism distinction, there is growing evidence that this distinction is too broad to adequately capture how cultures differ in consistency, “traitedness,” or other personality and self processes (del Prado et al., 2007
; Church, Katigbak, del Prado, Valdez-Medina, et al., 2006
; Malloy, Albright, Díaz-Loving, Dong, & Lee, 2004
). For example, the Mexican and Chinese cultures are both generally viewed as collectivistic. However, Malloy et al. (2004)
found that Chinese, but not Mexicans, exhibit less interobserver agreement in trait ratings—an indirect indicator of cross-situational consistency—than Americans. Similarly, Scollon et al. (2005)
found that the frequency and intensity of positive and negative moods were positively correlated among Asian Americans, Japanese, and Indians, but not among European or Hispanic Americans. In these two studies, as well as the present one, various types of inconsistency were most evident in selected Asian cultures, not collectivistic cultures generally (see also Bagozzi et al., 1999
; Kitayama et al., 2000
; Shimmack et al., 2002). Contrary to this pattern, however, Oishi et al. (2004)
found that Asian Indians and Japanese, but also Hispanic Americans, exhibited less within-individual consistency in affect than European Americans.
If the individualism-collectivism distinction cannot account for cultural differences in consistency, what alternative explanatory variable might? A plausible theoretical alternative is the dialecticalism of East Asian or Confucian cultures (Peng & Nisbett, 1999
). Indeed, this was the position taken by English and Chen (2007)
, who argued that East Asians’ dialectical thinking—with its greater acceptance of contradiction and change—leads to greater variability in East Asians’ self-concepts across relationship contexts. We conducted two follow-up analyses in hopes of augmenting the plausibility of this interpretation in the current study.
In the first analysis, we compared the cross-role consistency of ethnic Malays (n = 103) and ethnic Chinese (n = 102) in our Malaysian sample. If we assume that the ethnic Chinese have had greater exposure than the ethnic Malays to Confucian influences or Asian dialecticalism, then the ethnic Chinese should exhibit less cross-role consistency. Indeed, this was the case. In univariate ANOVAs, statistically significant (or marginally so) mean differences were found between the ethnic Malays and Chinese, respectively, for the SD index in the original data (M = .61 vs. .66, p < .08, eta2 = .02); for the SD index in the ipsatized data (M = .53 vs. .62, p < .001; eta2 = .05); for the PCA index (M = 71.6 vs. 63.7, p < .001, eta2 = .06); and for the residualized SD index in the ipsatized data (M = −.02 vs. .04, p < .001, eta2 = .05). For each of these indices, the means revealed lower consistency for the ethnic Chinese than for the ethnic Malays. Only the residualized SD index in the original data failed to show a significant mean difference (M = −.02 vs. .01, p > .05, eta2 = .003).
In a second follow-up analysis, we sought to replicate previous studies that found smaller inverse correlations, or even positive correlations, between positive and negative affects in East Asian cultures, as compared to Western cultures (Bagozzi et al., 1999
; Kitayama et al., 2000
: Schimmack et al., 2002
; Scollon et al., 2005
). Indeed, we found the following pattern of correlations between PANAS Positive and Negative Affect scores in the six cultures, ordered by size from negative to positive: Australia, r
= −.31; United States, r
= −.29; Malaysia, r
= −.05; Mexico, r
= .00; Philippines, r
= .08; and Japan, r
= .32. Only in the East Asian culture, Japan, was the correlation statistically significant in the positive direction (p
< .01). Some researchers have interpreted a positive correlation between positive and negative affect as consistent with Asian dialecticalism, because it suggests the simultaneous (and arguably contradictory) experience of emotions of opposite valence (Bagozzi et al., 1999
; Schimmack et al., 2002
; however, see Scollon et al., 2005
Although plausible, we believe the interpretation of our results in terms of East Asian or Confucian dialectical thinking remains tentative at his point. The nature and uniqueness of Asian dialectical thinking is still debated (e.g., Chan, 2000
; Ho, 2000
; Peng & Nisbett, 2000
) and needs to be further investigated in additional cultures. In this regard, a limitation of the present study was our inclusion of only one East Asian culture. Interpretations in terms of Asian dialectical thinking will also be strengthened by attempts to directly measure and test the ability of participants’ dialectical thinking to mediate individual and cultural differences in consistency (e.g., see English & Chen, 2007
). In the meantime, the more definitive finding of our study is that the individualism-collectivism distinction does not adequately account for cultural differences in cross-role consistency.
If participants in Japan or other East Asian cultures exhibit less cross-role consistency—in either self-ratings or actual behavior—does this necessarily imply that East Asians experience a less coherent sense of self? English and Chen (2007)
, drawing on recent theory and research on individual differences in “if-then” profiles or behavioral signatures, have argued that this is not the case. “If-then” profiles refer to stable cross-situational profiles of behaviors (i.e., if
in situation A, then
behavior A, but if
in situation B, then
behavior B) (e.g., Mischel, Shoda, & Mendoza-Denton, 2002
; Shoda, Mischel, & Wright, 1994
), Indeed, English and Chen found that Asian Americans’ trait ratings were less consistent across relationship contexts than those of European Americans, but just as stable over time within
contexts, suggesting the existence of stable self-concepts defined in “if-then” terms. Similarly, it is possible that our Japanese participants maintain a stable self-concept within
contexts across time, which might account for our finding that individual differences in consistency across
role contexts were not much related to adjustment in the Japanese sample. We did not include a longitudinal component in the present study, so we could not compare within-context stability across the six cultures. However, future studies could supplement traditional trait approaches, which largely treat situational effects as “error” (Mischel et al., 2002
), by examining the stability of if-then profiles across cultures.
Consistency and Alternative Predictors of Adjustment
Cross-role consistency predicted aspects of adjustment in all six cultures, particularly in the ipsatized data. These results, which are consistent with Western theories of personality and mental health (Erikson, 1950
; Jahoda, 1958
; Jourard, 1965
; Maslow, 1954
) and previous studies in the United States (Block, 1961
; Campbell et al., 1996
; Donahue et al., 1993
; Sheldon et al., 1997
), indicate that the relationship between consistency and adjustment is not limited to individualistic cultures. Prediction was most reliable in the American sample, and weakest in the Japanese sample, where consistency was primarily related (inversely) to negative affect. The more tenuous relationship between consistency and adjustment in the Japanese sample, as compared to the other collectivistic cultures, again favors an interpretation in terms of East Asian dialectical thinking, or some other aspect of East Asian or Confucian cultures, rather than individualism-collectivism.
The alternative constructs proposed by cultural psychologists—personality coherence, social appraisal, and relationship harmony—predicted adjustment in all six cultures, but were not better predictors in the collectivistic cultures than in the individualistic cultures. Suh (2002)
found that social appraisal—self-perceptions of how satisfied significant others are with one’s life—was a better predictor of life satisfaction in Korea than in the United States, but in the present study this relationship was equally strong in individualistic and collectivistic cultures. Similarly, Kwan et al. (1997)
and Kang et al. (2003)
had proposed that relationship harmony may be more important as a predictor of life satisfaction in collectivistic cultures than in individualistic cultures, but this was not the case in the present study. Rather, our results were similar to those of Chen, Chan, Bond, and Stewart (2006)
, who failed to find cultural differences in the predictive relationship between relationship harmony and depression in the United States and Hong Kong. Finally, Kitayama and Markus (1999)
proposed that, in Eastern cultures, personality coherence —the balance or coherence of multiple, even contradictory, aspects of self or personality—is more important than trait consistency. Our new measure, which was based on the conceptualization of Kitayama and Markus, was
successful in predicting adjustment, but was not uniquely predictive in the collectivistic (or Eastern) cultures in the study. The number of cross-cultural studies of this type is still small, and previous studies have been important in suggesting that the determinants of adjustment may differ across cultures (e.g., Chen et al., 2006
; Kang et al., 2003
; Kwan et al., 1997
; Suh, 2002
; Suh, Diener, Oishi, & Triandis, 1998
). However, the results of the present study, which included a broader sampling of cultures, suggest that the determinants of adjustment may be more similar than different across cultures. It is possible, of course, that alternative constructs that were not included in this study, such as self-efficacy (Chen et al., 2006
), fulfillment of social norms (Suh et al., 1998
), and perceived social support (Uchida et al., 2004
) will exhibit greater differential prediction of adjustment across cultures.
Consideration of Rival Interpretations
Before concluding, we discuss several rival interpretations of our results, as well as some limitations of the study. We first consider possible limitations associated with the trait-role rating task itself. A number of researchers have successfully used this task to investigate consistency. The total number of trait-role ratings has ranged from 100 to 175 in studies by Suh (2002)
, Campbell et al. (2003)
, and Baird et al. (2006)
, 240 in Sheldon et al.’s (1997)
study, and 360 in Donahue et al.’s (1993)
investigation. Although only Suh (202) applied the task cross-culturally, all of these researchers obtained results with this task that seem meaningful. Nonetheless, the task is repetitive and rival interpretations of our results should be considered.
First, in any study that uses Likert-type scales, the possible impact of response styles should be considered. For example, cross-role consistency could result from a respondent using similar rating points for most of his or her traits, leading to small cross-role standard deviations (i.e., SD
indices). Several considerations argue against this possibility, however. First, like most previous researchers, we attempted to reduce the impact of response sets by presenting the specific roles in different orders and by ordering the traits differently for each role. Second, the consistency indices were computed within
each individual, so that individual differences in the overall level of the ratings (e.g., some respondents using the middle of the rating scale and others the extremes) should have limited impact (beyond the statistical possibility that SD
indices can be larger for respondents who use the middle of the scale; Baird et al., 2006
). In any case, the vast majority of respondents in each culture (from 83% in the Philippines to 98% in the United States) used the full range of the rating scales, probably due in part to our inclusion of traits associated with both the positive and negative poles of each Big Five dimension. Finally, we analyzed the trait-role data in both original and ipsatized data. In the original data, the SD
indices already control for individual differences in the grand means of respondents’ ratings (i.e., subtracting out the grand mean has no effect on the standard deviation of the ratings for a given trait). By ipsatizing the data, we went further by controlling for individual differences in both the overall mean and variability of participants’ ratings. The overall pattern of results with the SD
index in the ipsatized data suggests that this method was effective in controlling for response styles.
A second rival interpretation of our results is the following: Rather than carefully envisioning and evaluating their traits in specific roles, respondents may at times rely on their general, abstract trait representations to rate their traits in these roles. If so, this would contribute artificially to greater cross-role consistency. We offer two types of evidence to counter this interpretation. First, if cross-role consistency is largely an artifact of the general trait representations, then the cross-role consistency correlations should be insignificant when the general trait ratings are partialled out. To obtain a single estimate of the cross-role correlations for each culture, we computed pooled correlations between each pair of roles across all individuals and traits in the ipsatized data (by using the ipsatized data we eliminated artifactual covariation due to individual differences in scale use). We then computed pooled partial correlations between each role-pair, controlling for the general trait ratings. Across the six cultures, all but 1 of 60 cross-role correlations remained statistically significant (p < .01) after controlling for the general trait ratings. The size of the cross-role partial correlations ranged from .21— .47 in the United States, .12— .47 in Australia, .19— .36 in Mexico, .28— .45 in the Philippines, .31— .49 in Malaysia, and .02— .49 in Japan. These (partial) consistency correlations probably represent an overcorrection, but they do show that cross-role consistency was not due solely to respondents’ general trait representations.
As an additional demonstration that participants distinguished between the various roles in their trait ratings, we computed Big Five trait scores for each role by averaging the raw ratings for the eight traits associated with each dimension (reverse-keying where necessary). We then computed a repeated measures ANOVA with gender as a between-subjects variable and Big Five dimension and role as repeated (within-subjects) factors. The most relevant finding, in all six cultures, was that the two-way interactions between Big Five dimension and role were statistically significant (p
< .001) and fairly large in size (partial ƞ2
values ranging from .18 to .25; Cohen, 1988
). This indicates that participants’ trait ratings were responsive to the specific role contexts. In addition, more modest three-way interactions between gender, Big Five dimension, and role were statistically significant (p
< .001 to .05) in all six cultures (partial ƞ2
values ranging from .01 to .03). This revealed that men and women exhibited somewhat different patterns of Big Five traits across the various roles, as one would expect. In summary, these findings suggest that the results reported in the study were not substantially impacted by individual or cultural differences in response styles, and that respondents in each culture did attend to, and differentiate, the specific role contexts when making their trait ratings.
A final limitation of the study was our sampling of only college students, who are probably more individualistic, and more exposed to Western or “global culture,” than more representative cultural samples. As noted in the method section, the samples were reasonably representative of the college student populations in each culture, however, and there were no marked differences in the relative prestige or eliteness of the university samples across cultures. Participants in one cultural sample, the Philippines, were a bit younger, on average, than participants in the other cultural samples because of the structure of the educational system in the Philippines (i.e., the absence of a middle-school level). However, there is no a priori reason to anticipate more or less consistency in a slightly younger sample. More importantly, the pattern of results in the Philippines did not stand out in significant ways from those in the other collectivistic cultures, with the exception of Japan. Finally, although the samples may be more individualistic or Westernized than broader samples within each culture, it should be noted that previous studies that have supported cultural psychology hypotheses have also been conducted primarily in college student samples.