Positive-Negative Asymmetry in Social Discrimination

PERSONALITY AND SOCIAL PSYCHOLOGY BULLETIN Mummendey et al. / POSITIVE-NEGATIVE ASYMMETRY

Positive-Negative Asymmetry in Social Discrimination: Valence of Evaluation and Salience of Categorization Amélie Mummendey Sabine Otten Uwe Berger Thomas Kessler Friedrich-Schiller-Universität, Jena, Germany Several studies have consistently demonstrated a positive-negative asymmetry in intergroup discrimination. As a possible explanation for this effect, the authors investigated whether stimulus valence has an impact on the salience of social categorization, which, in turn, is assumed to determine the degree of intergroup differentiation. It was hypothesized that the confrontation with negative stimuli instigates a change in the level of inclusiveness of self-categorization, inhibiting the differentiation based on the initial social categorization. Two studies with factors valence (positive, negative) and salience (low, high) were conducted to test these assumptions. Results were encouraging with respect to a category-based explanation of the valence effects on social discrimination. Implications of these findings for classical theories on behavior in minimal intergroup situations are discussed.

In his now classic monograph The Nature of Prejudice,

Allport (1954) defined social discrimination as “any conduct based on a distinction made on grounds of natural or social categories, which have no relation either to individual capacities or merits or to the concrete behavior of the individual person” (p. 51). Searching for antecedent conditions that may lead people to favor their own group and to disadvantage members of other groups, Tajfel, Billig, Bundy, and Flament (1971) published their striking Minimal Group Paradigm (MGP) studies: Social categorization per se turned out to be sufficient to lead participants to discriminate in favor of members of their own group and against members of the outgroup. Although there have been some critical voices questioning the validity or the interpretation of the MGP findings (e.g., Diehl, 1990; Messick & Mackie, 1989), the

finding that an arbitrary categorization into “us” (the ingroup) and “them” (the outgroup) can be sufficient to elicit social discrimination between groups is broadly accepted as a robust phenomenon in intergroup research (Brown, 1995; Tajfel & Turner, 1986). Social Identity Theory (SIT) (Tajfel & Turner, 1979, 1986) provides an explanation of this mere categorization effect: According to this theory, individuals strive to achieve or maintain a positive social identity, which is based to a large extent on favorable comparisons that can be made between the ingroup and some relevant outgroups. Thus, both ingroup favoritism and outgroup discrimination are tools to achieve positive distinctiveness of one’s own group. Subsequently, Self-Categorization Theory (SCT) (Turner, Hogg, Oakes, Reicher, & Wetherell, 1987) proposed positive distinctiveness as equivalent to the “relative prototypicality of the ingroup on valued dimensions of intergroup comparison” (p. 62). In both SIT and SCT, preferential treatment of one’s Authors’ Note: The studies reported here were conducted within a research project supported by the Deutsche Forschungsgemeinschaft (DFG) (No. MU-551-11-3). We are grateful to Alastair Coul, Christine Alterhoff, Sabine Greis, Dirk Pisula, Nils Wandersleben, Iris Kohnen, Alexandra Rosol, Charlotte Hirsch, Monika Schmalbrock, Hildegard Möllmann, Jorma Schubert, Mireille Herbert, and André Lammers for their assistance during conceptualization of the experiments and data collection. We are also extremely grateful to Rupert Brown and two anonymous reviewers for their helpful comments and suggestions on an earlier draft. Correspondence concerning this article should be addressed to Prof. Dr. Amélie Mummendey, Friedrich-SchillerUniversität, Lehrstuhl für Sozialpsychologie, Humboldtstr. 26, D-07743 Jena, Germany; e-mail: [email protected]. PSPB, Vol. 26 No. 10, October 2000 1258-1270 © 2000 by the Society for Personality and Social Psychology, Inc.

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Mummendey et al. / POSITIVE-NEGATIVE ASYMMETRY own group and discrimination against the outgroup is conceptualized as the result of individuals’ striving for positive social distinctiveness, in which they engage when they identify as members of a group, in a context where the categorization into ingroup and outgroup is made salient to them. Surprisingly, the mere categorization effect on ingroup favoritism does not appear when positive distinctiveness of one’s own group has to be established by comparisons on negatively valued dimensions or by differential allocations of negative resources such as costs or burdens. A series of experiments have demonstrated this so-called positive-negative asymmetry in social discrimination (e.g., Blanz, Mummendey, & Otten, 1995; Mummendey, Otten, & Blanz, 1994; Mummendey et al., 1992; Otten, Mummendey, & Blanz, 1996; for a review, see Mummendey & Otten, 1998). Under MGP conditions, group members who evaluated ingroup and outgroup on evaluation dimensions with positive connotations (such as intelligence or creativity) or who allocated positive resources (such as money or value points) showed the usual ingroup favoritism and outgroup discrimination. However, when using negatively valued dimensions (such as confusion or boredom) or allocating aversive treatments (such as duration of unpleasant noise or number of boring tasks), discrimination disappears (for a meta-analysis of discrimination on positive and negative resources, see Buhl, 1999). Thus, the same intergroup context that provides sufficient conditions for discrimination in the positive area seems insufficient to do so in the negative domain. How can this effect be explained? Neither SIT nor SCT predicts any impact of valence on intergroup behavior. On the face of it, the asymmetry effect seems to create difficulties for the postulate of “a general tendency to seek positive distinctiveness for oneself at any salient level of self-categorization” (Turner et al., 1987, p. 62). However, the explanatory approach in terms of valence-specific differences in the level of categorization that is crucial in the scope of this article provides an integrative framework that might solve the apparent contradiction between positive-negative asymmetry in social discrimination and SIT or SCT: If the postulate of a general tendency to seek positive distinctiveness at any salient level of self-categorization is not to be abandoned, then the absence of discrimination under negative valence conditions must be a consequence of changed salience of self-categorization. If distributing negative resources changes the salience of the group-level categorization, then a differential treatment of ingroup and outgroup (ingroup favoritism) must appear particularly inappropriate (see Blanz, Mummendey, & Otten, 1997, on valence-specific differences in the inappropriateness of ingroup favorit-

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ism): Why should members of the two groups be treated differently if the rationale for differentiating between them has disappeared? Integrating the cognitive analysis of positive-negative asymmetry in social discrimination as put forward by Otten, Mummendey, and Buhl (1998), it can be assumed that a more systematic, elaborate examination of the category information in the negative domain leads to the perception that the minimal group distinction provides no convincing, legitimate basis for differentiation between the categories and, therefore, a representation of the situation on another level of self-categorization is preferable. In line with our previous findings, we hypothesize the typical positive-negative asymmetry in social discrimination in situations with low (or medium) salience and negative valence.1 In contrast, as already shown in earlier studies introducing status and size differentials between groups (Blanz et al., 1995; Mummendey et al., 1992; Otten et al., 1996), high(er) category salience is expected to elicit ingroup favoritism in both positive and negative intergroup comparisons. Self-categorization theory predicts that increasing category salience will support a self-categorization on the intergroup level that causes group formation and group behavior (see, e.g., Hogg & Turner, 1987, p. 338). In the positive conditions, a rather minimal intergroup setting has proved to be sufficient to elicit self-categorization on the intergroup level and, hence, ingroup favoritism. Therefore, in principle, a further increase of category salience need not result in a change of the level of self-categorization but could still be reflected in a certain increase in ingroup biases (see Mullen, Brown, & Smith, 1992). In summary, we assume that both an experimentally induced variation of category salience as well as stimulus valence will have an impact on self-categorization (i.e., perceived category salience) and, as a result, ingroup favoritism. In statistical terms, the closest expression of our expectations is a contrast effect (Abelson & Prentice, 1997; Rosnow & Rosenthal, 1995) with an assumed difference between the low salience/negative condition and the three other conditions (which themselves are assumed to be equal). More specifically, our hypotheses are the following:

Hypothesis 1 (valence hypothesis): In the low salience conditions, there will be significant ingroup favoritism in the positive condition exclusively (i.e., the typical asymmetry effect). In the high salience conditions, we expect significant ingroup bias in both valence conditions (possibly with a larger effect size within the positive conditions); analogously, only for the low salience condition do we expect perceived salience to be lower with negative than with positive valence.

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PERSONALITY AND SOCIAL PSYCHOLOGY BULLETIN

Hypothesis 2 (salience hypothesis): The manipulation of category salience is expected to manifest in a main effect on perceived salience. Correspondingly, there will be more discrimination in the high salience condition than in the low salience condition. However, whether the effect sizes are identical for positive and negative valence cannot be postulated in advance. Hypothesis 3 (mediation hypothesis): The assumed valence-specific variation in the salience of the intergroup distinction will affect not only measures of intergroup bias but also variables indicating category salience. These measures are assumed to mediate the influence of valence.

Salience

Low

High

Valence

Positive

Negative

Positive

Negative

Intergroup Differentiation

IG > OG

IG = OG

IG > OG

IG > OG

Figure 1

Design and hypotheses for intergroup evaluations/allocations. NOTE: IG = ingroup, OG = outgroup.

Method PARTICIPANTS

To sum up, the aim of the research presented in this article is to explain the positive-negative asymmetry in social discrimination in terms of valence-specific differences in the salience of the social categorization on which intergroup evaluations are based. Therefore, we attempt to demonstrate that an active manipulation of the social categorization in a typical intergroup experiment has similar effects on ingroup favoritism as the valence factor (see also Kruglanski & Mackie, 1990, for this line of reasoning). If, as we assume, the valence effect on intergroup discrimination is mediated by salience of categorization on the group level, then the manipulation of salience will affect the outcome of social discrimination in a way similar to the manipulation of the valence of intergroup comparison dimensions. In addition, the assumed valence-specific decrease in the salience of the intergroup distinction is hypothesized to manifest itself not only in measures of intergroup bias but also on variables indicating category salience. Two studies, both based on a 2 × 2 design with factors Salience (low, high) and Valence (positive, negative) were conducted to test these hypotheses. Figure 1 shows the design and illustrates the hypotheses concerning the occurrence of ingroup favoritism. STUDY 1

Turner et al. (1987) posit that “the salience of some ingroup-outgroup categorization in a specific situation is a function of an interaction between the ‘relative accessibility’ of that categorization for the perceiver and the ‘fit’ between the stimulus input and category specifications” (p. 54). Oakes (1987) specified the concept of fit by differentiating between structural fit (which relates to the principle of meta-contrast) and normative fit (which relates to the social meaning of ingroupoutgroup categorizations). In our first experiment, we manipulated the salience of the ingroup-outgroup categorization by varying structural fit, adopting an experimental procedure by Oakes, Turner, and Haslam (1991) that will be described in more detail below.

As partial fulfillment of course requirements, 62 1st-year psychology students from a German university participated in this study. They were randomly assigned to the four experimental conditions. PROCEDURE

Participants observed a video presentation of a six-person discussion group (women only). The participants of this discussion argued either for the allocation (positive valence condition) or against the withdrawal (negative valence condition) of money in favor of their own faculty. As mentioned above, the salience manipulation was adopted from an experiment by Oakes et al. (1991): In the collective/conflict condition (representing high category salience), the video showed two equally sized groups, three students allegedly from the psychology faculty (the ingroup) and three from the education faculties who argued unequivocally in favor of their own faculty. This way, structural fit in terms of relative accentuation of intercategory differences as compared to intracategory differences was maximized (see Turner et al., 1987, p. 47). In the solo/deviance condition (representing low category salience), a single psychology student was pitted against five students from the education faculties, each of whom engaged in ingroup favoring arguments. Oakes et al. (1991, p. 132) posit that the singularity of one solo group member provides an alternative to simple category-based explanations of the target’s behavior: In contrast to the collective/conflict condition, the target’s behavior could be attributed more to idiosyncratic personality features than to category membership. Thus, perceived intracategory similarity and, correspondingly, the meta-contrast ratio and structural fit of the category (or prototypicality of the target group member) might decrease. To provide comparability with our previous studies, in contrast to the study by Oakes and associates (1991), participants did not make allocations to a specific target person but rather to their own group and to the other group. Each experimental session lasted about 20 minutes. After the video presentation, participants were asked to allocate (positive valence) or withdraw (negative

Mummendey et al. / POSITIVE-NEGATIVE ASYMMETRY valence) money in reference to the two faculties. Seven matrices (each presenting 11 allocation alternatives, ranging from maximum ingroup favoritism over fairness to maximum outgroup favoritism, or vice versa) were presented in a booklet, with each matrix representing one specific resource of a faculty’s department (research, library, rooms, personal computers, guest experts, teachers). Subsequently, participants answered a questionnaire comprising four parts: causal attribution, ingroup identification, manipulation checks, and demographic data. Finally, students were debriefed and thanked for their cooperation. DEPENDENT MEASURES

Discrimination. The mean score of the decisions on the seven allocation matrices represented the overall ingroup-outgroup difference created by the participants. Positive values indicate ingroup favoritism, whereas negative values indicate allocations in favor of the outgroup. Ingroup identification. Three items referred to ingroup identification (“I identify with the group of psychology students,” “After the discussion, I had a good impression of psychology students,” and “I like being a psychology student”). Items were measured on 7-point scales ranging from 1 (not true) to 7 (completely true). Causal attributions. Causal attributions referred to the participants’ own allocation decisions. It was assessed whether attributions made reference to individual characteristics (i.e., personal identity) or to the faculty membership (i.e., social category attribution): “Insofar as something about the kind of person you are was responsible . . . which of your characteristics was more important—something to do with your individual personality (0) or something to do with the fact that you are a psychology student (9) (see Oakes et al., 1991, p. 136)? Manipulation checks. Five items were included as checks on the manipulations of the experimental factors (on a 3-point scale ranging from true, don’t know, to not true). To check on the manipulation of valence, participants were asked whether the discussion in the video referred to the allocation of additional money to their department (item 1) or whether it referred to the withdrawal of money from their department (item 2). To check on the manipulation of the category salience, three items asked for the relative numbers of psychology versus education students in the discussion group on the video: Was there an equal number of psychology students and education students (item 3)? Were there fewer psychology students than education students (item 4)? and Were there more psychology students than education students (item 5)? Unfortunately, 13 of our participants did not reflect their experimental condition on at

TABLE 1:

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Study 1: Means and Standard Deviations for Intergroup Allocations, Social Category Attributions, and Ingroup Identification Low Salience Positive Valence n = 15

Negative Valence n = 20

High Salience Positive Valence n = 12

Negative Valence n = 15

Intergroup allocations .69* (.53) –.03 (.46) .55* (.46) .88* (1.04) Category attributions 5.33 (2.69) 2.40 (2.99) 4.18 (2.88) 4.87 (3.13) Ingroup identification 4.62 (1.31) 4.65 (.83) 3.78 (1.16) 4.44 (1.11) Contrast 1 –3 1 1 NOTE: Positive ingroup-outgroup differences indicate ingroup favoritism. Standard deviations are in parentheses. * Means differ significantly from zero at p < .05.

least one of these items (i.e., they gave “don’t know” or wrong answers). Irrespective of this finding, we decided to include all participants in our main analyses (for a discussion on the usefulness of manipulation checks as criteria for sample selection, see Sigall & Mills, 1998). This decision was further supported by an additional analysis on the selected sample, revealing essentially the same results.2 Results DISCRIMINATION

A 2 × 2 ANOVA (Salience × Valence) revealed a significant interaction, F(1, 58) = 9.53, p < .01, and a significant main effect for salience, F(1, 58) = 5.11, p < .05. As a sensible test of the valence hypothesis, the 2 × 2 design (Salience × Valence) was changed to a unifactorial design with four factor levels (1 × 4) and a contrast analysis was computed (see Table 1).3 The contrast (1, –3, 1, 1) was significant, t(58) = 4.11, p < .01, with an effect size of dcon = 1.08, indicating a large effect. Further residual analysis shows that there is no further systematic variance in the data. As predicted, there was significant lower discrimination in the low salience/negative valence condition, whereas the other conditions do not differ from one another. These results support our valence hypothesis. Furthermore, the only cell in which the mean discrimination did not differ from zero (i.e., in which ingroup favoritism was not significant) was the low salience/negative valence condition. IDENTIFICATION

Reliability analysis indicated that the three items could be reliably combined in a single identification measure (α = .67). For identification, there is only a mar-

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Figure 2

Study 1: Path models with valence as predictor, meta-contrast ratio as mediator, and discrimination as criterion. Path models for low salience (left) and high salience (right). †p < .10. *p < .05. **p < .01.

ginal significant effect of salience, F(1, 57) = 3.86, p = .08. The overall correlation between identification and intergroup allocations was marginally significantly positive (r = .19, p < .10, one-tailed). Separate calculations for each cell indicated a pattern reflecting the contrast effect that was found for the discrimination measure: The positive correlation was reflected in three conditions—positive valence/low salience (r = .53, p < .05, one-tailed), positive valence/high salience (r = .42, p < .10, one-tailed), and n egativ e v alence/ high salience (r = . 21, n s , one-tailed)—but was virtually zero in the negative valence/low salience condition (r = .06, ns, one-tailed). ATTRIBUTIONS

The overall correlation between social category attributions and ingroup-outgroup allocations was positive and significant (r = .45, p < .001). This result supports the predicted relation between category salience and categorybased differentiation: The higher the category salience (i.e., the more meaningful the social category), the higher the difference between ingroup and outgroup allocations. Again, separate correlations for each cell showed a pattern corresponding to the contrast effect found for discrimination: There was a significant correlation for positive valence/low salience (r = .67, p < .01), a marginally significant effect for positive valence/high salience (r = .49, p < .05), and a positive correlation under the negative valence/high salience condition (r = .29, ns, one-tailed). As found for the identification measures, the effect in the negative valence/low salience condition was not significant (r = .15, ns). Table 1 shows that the mean score for the social category attributions was lowest in the negative valence/low salience condition. There were no significant main effects but an interaction for the two factors of Valence and Salience, F(1, 58) = 5.70, p < .05. However, as expected, there was a significant contrast effect for the low salience/negative valence condition compared to

the three other cells, t(58) = 2.99, p < .01, with an effect size of dcon = .79. MEDIATIONAL ANALYSIS

It is hypothesized that perceived salience of social categories mediates the influence of valence on discrimination. According to our contrast analysis, this mediation should hold mainly in the low salience condition. For the high salience condition, no valence effects are expected. Hence, mediation was tested separately for the low salience and the high salience conditions. Following Baron and Kenny (1986), testing mediation requires multiple regression analyses in which a series of regression equations are estimated. In the first equation, the dependent variable (discrimination) is regressed on the independent variable (valence) to demonstrate that the independent variable predicts the dependent variable. Second, the independent variable (valence) is regressed on the proposed mediator (i.e., causal attribution). Finally, both the dependent variable and the potential mediating variable are regressed on the independent variable. A mediation is demonstrated to the extent that the influence of the independent variable on the dependent variable (first equation) is reduced if the mediator is simultaneously taken into account (third equation)or , in other words, if the independent variable shows an indirect effect on the dependent variable. Following this logic, we found evidence for a mediation of the Positive-Negative Asymmetry × Perceived Category salience (see Figure 2): In the low salience condition, valence shows a significant relation to intergroup differentiation (β = .60), t(34) = 4.33, p < .001, and a significant effect on causal attribution (β = .46), t(34) = 2.99, p < .01. Considering valence and causal attribution together, the third equation reveals a significant effect of causal attribution (β = .33), t(34) = 2.25, p < .05, and of valence on intergroup differentiation (β = .44), t(34) =

Mummendey et al. / POSITIVE-NEGATIVE ASYMMETRY 3.02, p < .01. In the high salience condition, valence showed no effect on intergroup differentiation (β = –.20), t(26) = 1.02, p > .10, as well as on causal attribution (β = –.12), t(26) = .59, p > .10. The relation of causal attribution remained roughly equal as the low salience condition (β = .32), t(26) = 1.71, p < .10. Taken together, these results reveal a marginally significant indirect effect of valence on intergroup differentiation in the low salience condition (β = .16), t = 1.85, p < .10, whereas in the high salience condition, the indirect effect is negligible (β = –.04, t = 0.57, p > .10). However, the difference between the two salience conditions did not reach significance. These results, if interpreted with caution, indicate that the mediating effect of causal attribution may be moderated by salience. In other words, causal attribution (perceived salience) mediates the relation of valence on intergroup differentiation in the low salience condition (manipulated salience) exclusively. Discussion Although some of the predicted effects did not reach the conventional level of significance, in sum, results from Study 1 support our main assumptions. There was a positive-negative asymmetry in social discrimination only under low salience conditions. Increasing the social category salience eliminated this asymmetrical effect and resulted in ingroup favoritism within both valence conditions. In accordance with prior research by Oakes and coworkers (1991), increased ingroup favoritism was (at least marginally) related to increased category-based attribution. However, in contrast to these authors, in the positive domain, we found no increased discrimination— and, correspondingly, no increased category-based attribution—for the high salience condition as compared to the low salience condition. The causal attribution data correspond to the discrimination data: Both dependent variables analogously reflect our independent manipulation of valence and salience in the predicted manner. Furthermore, the mediation analysis revealed some initial evidence that the effect of valence on discrimination is mediated by perceived salience. Evidence from Study 1 clearly supported a categorization account of the positive-negative asymmetry in social discrimination. However, it should be remembered that this was a first test of the assumed interplay between valence, perceived salience, and the probability of ingroup favoritism. Besides, some of the hypothesized effects reached only a marginally significant level. Hence, it seemed necessary and worthwhile to test the validity and generalizability of the results in a further study. This study was a modified, conceptual replication based on the same design but placed in a different experimental paradigm.

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STUDY 2

Study 2 tested the same set of hypotheses as reported in the introduction to Study 1: the valence hypothesis, the salience hypothesis, and the mediation hypothesis. In contrast to Study 1, the meta-contrast of perceived intergroup differences and the meta-contrast of perceived intergroup similarities were used as alternative measures of category salience (Turner et al., 1987). Study 2 not only used different measures for category salience but also varied three further aspects: First, because the salience manipulation by Oakes at al. (1991) has received some criticism (see Biernat & Vescio, 1993), we chose an operationalization based not only on a variation of structural fit but also on a variation of normative fit and investigated perceived similarities and differences instead of causal attributions. Second, we measured separate evaluations of ingroup and outgroup instead of intergroup allocations on matrices. In this way, we could not only gather information about the generalizability of effects for different dependent variables (which may be weak; see Otten et al., 1996) but also were able to distinguish the impact of salience and valence on ingroup treatment, on one hand, and outgroup treatment, on the other hand. Finally, we chose a different experimental paradigm, which more actively involved participants as group members. Method PARTICIPANTS AND DESIGN

The study consisted of 136 students from the faculties of natural sciences and social sciences at a large German university. However, as will be outlined below, our specific operationalization of category salience necessitated a reduction of the final sample to a total of 106 (58 women, 48 men). The mean age was 24 years. Participants were recruited through advertisements in a local newspaper and were paid 20 German marks for their participation. The design of this study has already been depicted in Figure 1. Valence was manipulated by the quality of attribute dimensions on which ingroup and outgroup were evaluated. The salience manipulation was based on variations in both structural as well as normative fit. PROCEDURE

Six to eight participants participated in one experimental session. They were divided into two subgroups of 3 to 4 persons. Care was taken that an equivalent number of participants from the two faculties attended each session. In the event that people did not show up at the arranged time, confederates took part in the experiment. Each session lasted about 50 minutes. Participants were randomly assigned to the experimental conditions.

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Ostensibly, the major concern of the experiment was processes in small group discussions. The intergroup distinction referred to the two alternative positions in a decision dilemma. In small groups, participants put forward arguments in favor of one of two opposing decisions. The ingroup was defined as the subgroup taking the same position as the respective participant, whereas the outgroup consisted of those persons arguing in favor of the decision alternative. The decision dilemma was introduced as follows:

The discussion lasted about 15 minutes. Afterward, participants completed a questionnaire about their experiences during the discussion. This questionnaire mainly comprised ingroup and outgroup evaluations concerning several attributes, which were either positive or negative (the sequence of ingroup judgments and outgroup judgments was varied between participants). Finally, there were questions on group identification and on perceived intergroup similarities and differences.

Due to changes in the financial budget, the city of Muenster has an extra amount of 2 million marks to be invested in either a social or an environmental project. In a committee meeting, supporters of two alternative projects have the opportunity to put forward arguments in favor of their project before the final decision is reached.

For all dependent measures, we used 10-point scales indicating participants’ degree of agreement with a given statement. The endpoints of these scales were labeled fully agree and fully disagree, respectively.

The participants’ task was to imagine this scenario and to argue as convincingly as possible in favor of either the environmental investment or the investment in a social work program. To reduce variance in the intergroup and intragroup interaction, the discussion was highly structured: After being assigned to one of the two discussion groups, participants were given 10 minutes to prepare for the task. By tossing a coin, it was decided which group started the discussion. After a member of the first group put forward an argument, a member of the opposing group commented on this argument and then put forward an argument of his or her own in favor of the decision alternative. This argument was commented on by a member of the first group, and so forth, until every group member had argued and commented once. Salience was manipulated in the following way: In the low category salience condition, equal numbers of students from two different faculties (natural sciences and social sciences) were invited for each experimental session; however, the experimenter overtly took care to ensure that each group discussing in favor of one of the alternative positions in the decision dilemma was mixed with respect to faculty. In the high salience condition, the faculty distinction was parallel to the distinction between positions in the decision dilemma, that is, students from social sciences argued in favor of a social work project, whereas students from natural sciences argued for supporting an environmental project. Normative fit was underlined by claiming that faculty membership would provide participants with specific expertise in putting forward arguments in favor of the respective position. In addition, structural fit was increased by putting signs of different colors and referring to the opposing decision alternatives on each group’s table in front of each group member.

DEPENDENT MEASURES

Ingroup and outgroup evaluations. The evaluative dimensions referred to the discussion style of ingroup and outgroup as well as to more general characteristics of the two groups. More specifically, the items concerned the following attributes (positive/negative): (a) discussion style: flexible/rigid, friendly/unfriendly, cooperative/ noncooperative, effective/ineffective, and considerate/ inconsiderate and (b) more general characteristics: tolerant/intolerant, confident/unconfident, likeable/ dislikeable, stimulating/boring, and open/reserved. Group identification. Two items referred to ingroup identification (“I identified with my discussion group” and “I appreciated being a member of my discussion group”) and two items referred to outgroup identification (“I would have preferred to be a member of the other discussion group” and “I would have preferred to put forward the arguments of the other discussion group”). The item “I would have preferred to put forward the arguments of the other discussion group” was not an identification item in the strict sense but was used to decide about the final sample selection. People who showed a strong preference for the discussion alternative (i.e., the outgroup’s position) were not included in the final analysis (see below). Category salience. Category salience was operationalized by the meta-contrast ratio, defined as the “average differences perceived between members of the category and the other stimuli . . . over the average difference perceived between members within the category” (Turner et al., 1987, p. 47). We computed two meta-contrast ratios, one for perceived similarities (mean similarities within groups over the similarities between groups) and one for differences (differences between groups over the mean differences within groups). High meta-contrasts indicated a high salience of social categorization. Perceived similarities and differences between and within ingroup and outgroup were measured by the following six items: (a/b) “There are several similari-

Mummendey et al. / POSITIVE-NEGATIVE ASYMMETRY TABLE 2:

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Study 2: Means and Standard Deviations for the Differences Between Ingroup and Outgroup Allocations, Separate Ingroup and Outgroup Allocations, Ingroup Identification, Intergroup Similarities, and Intergroup Differences Low Salience Positive Valencea

Difference IG-OG Ingroup evaluations Outgroup evaluations Ingroup identification Intergroup similarities Intergroup differences

.81* a 6.91 a 6.10 a 7.88 a 1.39 a 1.47

ab

(1.60) (1.23) (1.57) (1.94) (1.12) (1.69)

High Salience

Negative Valence a

.19 b 8.16 b 7.96 a 7.55 b .79 a .97

(1.40) (1.75) (1.53) (2.27) (0.44) (0.37)

Positive Valence b

1.35* a 6.46 a 5.11 a 7.98 ab 1.09 a 1.24

(1.43) (1.65) (1.38) (1.76) (0.76) (0.76)

Negative Valence ab

.85* b 8.27 b 7.42 a 7.82 ab 1.07 a 1.08

(1.55) (1.31) (1.70) (2.23) (0.69) (0.71)

NOTE: Positive ingroup-outgroup (IG-OG) differences indicate ingroup favoritism. Cell counts vary from 25 to 30. Means with different superscripts differ significantly according to the adjusted Bonferroni test. Standard deviations are in parentheses. * Means differ significantly from zero at p < .05.

ties/differences between the members of my group and the other discussion group,” (c/d) “In several respects, the members of my discussion group are similar/different,” and (e/f) “In several respects, the members of the other discussion group are similar/different.” Results MANIPULATION CHECKS

First, it was checked whether a decision dilemma with two equally plausible decision alternatives was achieved. In line with this aim, analyses of variance revealed neither a significant impact of the positions in the discussion (environmental investment/social investment) on the evaluation of the two groups in general nor an interaction of this factor with the two other independent variables. Although we tried to standardize intergroup interaction by structuring communication, there was still the possibility of systematic differences between the single experimental group sessions. However, nested design analyses of variance with Group Sessions as a nested factor showed neither a significant main effect nor significant interactions with the experimental factors. Hence, it was possible and legitimate to use individual participants as the unit of analysis. Identification The scales of ingroup and outgroup identification were reliable (α = .77 and .80, respectively). Generally, the identification with the experimental ingroup was very high (M = 7.28 on a 10-point scale) and significantly higher than the identification with the outgroup (M = 4.32), t(134) = 7.66, p < .001. Analysis of variance with the factors of Valence and Salience and ingroup identification as the dependent measure resulted in neither significant main effects, an interaction, nor any significant simple contrasts (see Table 2). In addition, we analyzed the link between ingroup identification and ingroup bias. Overall, this

correlation was significant (r = .21, p < .05). Differentiation of the sample according to stimulus valence indicated a significant correlation (r = .40, p < .01) in the positive domain exclusively (rnegative = .05, ns). An additional differentiation according to the salience factor did not qualify this pattern. Final Sample Selection In this experimental paradigm, group categorization referred to the two alternatives to be argued for in the decision dilemma. Normative fit in the high salience condition was emphasized by assigning students from the natural sciences to the proenvironment discussion group and the social sciences students to the pro–social project discussion group (referring to specific expertise in the fields, respectively), and salience was diminished by having members of both faculties discussing the respective alternatives (criss-cross categorization). However, because both social and environmental issues are relatively relevant in student contexts, some students of the natural sciences might in fact have been engaged in social projects, whereas others from the social sciences might have been more supportive for environmental projects. A clear preference for the other group’s discussion standpoint would have run counter to our salience manipulation (participants might have identified with the outgroup, resulting in a complete mismatch between faculty membership and a preferred discussion alternative). This aspect was measured by the item “I would have preferred to argue for the other groups’ decision alternative” (1 = completely disagree, 10 = completely agree). Although half of the participants clearly rejected this statement (scores 1 to 3), 24% scored 8 and above and were excluded from the final sample. Liking or dislike for one’s own group’s discussion alternative was not influenced by the quality of these alternatives itself . A t test with discussion alternative as an independent variable and liking or dislike as a dependent variable was not significant, t(104) = .27;

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there was no overall preference for either environmental or social welfare projects. In addition, an ANOVA analyzing whether the two experimental factors had an impact on preference for the discussion alternatives did not reveal any significant effects. Hence, our sample selection was clearly not confounded with any of our crucial experimental manipulations. Evaluations of Ingroup and Outgroup Subjective ratings concerning discussion styles and general group characteristics were summarized in one evaluative score for the ingroup and one evaluative score for the outgroup (negative scores recoded). Two subscales were excluded that were not sufficiently correlated with the other items (tolerant/intolerant, effective/ineffective). The reliability of the conjoint scales was high (α = .85 for the ingroup scale and .84 for the outgroup scale). A difference score—mean ingroup evaluation minus mean outgroup evaluation—was calculated to measure intergroup differentiation; positive differences indicate that the ingroup was favored over the outgroup. Difference scores. Data for the ingroup-outgroup difference scores broadly fit our hypotheses: The negative valence/low salience condition was the only one in which the difference in favor of the ingroup was not significant. Accordingly, we find the typical data pattern of the positive-negative asymmetry of social discrimination only within the low salience condition (see Table 2). A 2 × 2 ANOVA (Salience × Valence) revealed a marginal significant main effect of valence, F(1, 102) = 3.69, p < .10, with more differentiation in favor of the ingroup on the positive attribute dimensions than on the negative attribute dimensions and a significant main effect of salience, F(1, 102) = 4.24, p < .05, with more positive ingroup distinctiveness in the high salience condition than in the low salience condition. The interaction effect was not significant. A test of the contrast effects postulated by our valence hypothesis actually showed that, primarily, the low salience/negative valence condition differs from the other, t(102) = 2.52, p < .05, dcon = .50. The remaining residual variance shows no further significant systematic variation. Separate evaluations of ingroup and outgroup. Possible effects of the salience factor on ingroup evaluations as well as on outgroup evaluations were analyzed separately for both valence conditions. Only for outgroup evaluations, the main effect of salience was (marginally) significant; positive: F(1, 50) = 5.62, p < .05, negative: F(1, 54) = 3.01, p < .10. Table 2 shows that ingroup evaluations on negative dimensions were not affected by the salience manipulation, whereas the outgroup evaluations were signifi-

cantly more negative for high than for low salience. In fact, in the crucial negative valence/low salience cell, outgroup evaluations were as positive as the ingroup evaluations. CATEGORY SALIENCE

There was no main effect of salience on the meta-contrasts based on either perceived similarities or differences. However, we found a significant main effect of valence on meta-contrast/similarities, F(1, 102) = 4.12, p < .05, as well as a marginal effect on meta-contrast/differences, F(1, 102) = 3.02, p < .10. Furthermore, there was a marginally significant interaction between category salience and valence on perceived similarities, F(1, 102) = 3.69, p < .10. Both main effects support the assumption that negative valence decreases category salience (similarities: Mpos = 1.24, SD = .96; Mneg = .92, SD = .58; differences: Mpos = 1.35, SD = 1.29; Mneg = 1.02, SD = .55). In addition, a contrast analysis showed a significant effect for the meta-contrast on similarities, t(87) = 3.00, p < .01, dcon = .64,4 and a marginally significant effect for the meta-contrast on differences, t(61) = 1.98, p < .10, dcon = .51. The contrast effects replicate the findings of Study 1: The negative valence/low salience condition differed from the three other conditions, which themselves did not differ. The effects of stimulus valence on perceived salience correspond to the marginally significant effect of valence on intergroup differentiation, F(1, 102) = 3.69, p < .10 (Mpos = 1.09, SD = 1.52; Mneg = 0.49, SD = 1.49). As Table 2 shows, in line with the decrease in intergroup differentiation, there was the lowest meta-contrast ratio of perceived inter- and intragroup similarity as well as of inter- and intragroup differences in the negative valence/low salience condition. More important, the results of contrast analysis revealed essentially the same pattern of means on intergroup differentiation and both types of meta-contrast ratios. Hence, the findings of Study 1 were conceptually replicated. Bivariate correlations between the ingroup-outgroup difference scores in evaluations and the meta-contrast ratios were significant (r = .30, p < .01, meta-contrast/ similarities; r = .35, p < .001, meta-contrast/differences). As postulated by SCT, the larger the meta-contrast ratio, the higher the probability of ingroup biases. MEDIATIONAL ANALYSIS

Similar to Study 1, applying Baron and Kenny’s (1986) rationale for testing a mediation (see also Fiske, Kenny, & Taylor, 1982), we conducted a series of multiple regression analyses. In a first regression equation, valence was significantly related to ingroup favoritism (β = .20), t(104) = 2.02, p < .05. The second equation revealed a significant path of valence on perceived similarities between groups (β = .20), t(104) = 2.08, p < .05. Con-

Mummendey et al. / POSITIVE-NEGATIVE ASYMMETRY

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Figure 3

Study 2: Path models with valence as predictor, meta-contrast ratio as mediator, and discrimination as criterion. Path models for low salience (left) and high salience (right). *p < .05. **p < .01.

sidering valence and perceived similarities together as independent variables, results show a significant effect of perceived similarities (β = .27), t(104) = 2.83, p < .01, whereas the effect of valence was no longer significant (β = .13), t(104) = 1.48, p > .10. The indirect effect of valence on ingroup favoritism was marginally significant (b =.06), t(104) = 1.70, p < .10, which indicates that perceived similarities partially mediate the influence of valence on ingroup favoritism (see Baron & Kenny, 1986, p. 1177). When using meta-contrast ratio of intergroup differences as an indicator for salience, we obtain largely the same results, but coefficients are somewhat lower. The indirect effect of valence on ingroup favoritism fails to reach significance (β =.05), t(104) = 1.59, p > .10. We hypothesized that the mediation effect should hold in the low salience condition rather than the high salience condition. Therefore, a path model was established with valence as predictor, meta-contrast ratio of perceived similarities as a mediating variable, and ingroup favoritism as criterion (see Figure 3). In a multiple group analysis for high and low salience conditions with invariant matrices for both groups, results indicate no strong deviation of data from the common model (χ2 = 5.03, df = 6, p = .54). According to our hypothesis, the main difference of high and low salience conditions was expected for the relation between valence and perceived similarities. Thus, valence should influence the meta-contrast ratio of perceived similarities in the low salience condition exclusively. In a second multiple group analysis with a free path coefficient for the path of valence on perceived similarities in both groups, the model fit increases marginally significantly (χ2 = 1.34, df = 5, p = .93; χ2 difference = 3.69, df = 1, p < .10). These results support the assumption that salience moderated the mediation effect. Only in the low salience condition is the relation of valence and perceived similarities significant (β = .38, t = 2.87, p < .01; high salience condition:

b = .01, t = .08). Furthermore, the indirect effect of valence on ingroup favoritism was .11 (t = 2.05) in the low salience condition but zero in the high salience condition. Discussion Results of Study 2 replicated and extended the findings of Study 1 and are in line with our hypotheses (see Figure 1). The negative valence/low salience condition was the only one without significant ingroup favoritism. This manifested in the contrast effects between the low salience/positive valence condition, on one hand, and all other conditions, on the other hand. The same significant contrast also was shown for the salience measures, namely, the meta-contrast ratios for perceived similarities and differences. Finally, there was again evidence that the impact of valence on intergroup differentiation is mediated by perceived category salience. Although the different measures of category salience in Study 2 showed roughly the same pattern of means in the contrast analysis, there are some noteworthy differences: The influence of valence on meta-contrast ratio was lower for differences than for similarities and the interaction between salience and valence was more clear for similarities than for differences. Furthermore, valence had an impact not only on the mean structure but also on the variances of the two types of meta-contrast ratios: Variances within the negative valence condition were significantly lower than variances within the positive valence condition, similarities: F(1, 104) = 7.34, p < .01; differences: F(1, 104) = 4.26, p < .05. These valence effects on variances may reflect differences in the (perceived) consensus in judgments on either positive or negative evaluation dimensions. Hence, future research should further investigate these differences and should continue to investigate valid and sensible measures of category salience (see Berger, 1998).

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GENERAL DISCUSSION

In sum, our data provide considerable evidence for the assumption that varying the valence of intergroup comparison dimensions actually operated like a manipulation of category salience. Although in the single studies some of the expected effects failed to reach conventional levels of significance, the strong consistency between the two studies—using a different operationalization of both dependent as well as independent variables—is compelling. Nevertheless, a more direct test of whether negative stimuli decreased the salience of the intergroup distinction presupposed corresponding effects on salience measures. In this respect, results were encouraging as well: The data pattern for causal attributions in Study 1 and for perceived intergroup similarities in Study 2 correspond closely to the data pattern for intergroup differentiation. In both studies, the mediational analyses showed that the effect of valence on discrimination was mediated by perceived salience (meta-contrast).5 The results further indicated a moderation of this mediation by manipulated salience: In the low salience condition, valence has a strong indirect effect on differentiation that vanishes completely in the high salience condition. This moderation was due to the high effect of valence on category salience in the low salience condition. These results underline the conclusion that valence operates like a manipulation of category salience. Because studies on intergroup behavior have typically focused on the domain of positive resources, it had been an open question whether our manipulation of category salience would have identical effects within the two valence conditions. In fact, both studies indicated that the salience factor was more crucial within negative valence. Irrespective of valence, whereas a given high category salience was sufficient to elicit ingroup bias, experimentally induced low salience elicited significant bias in the positive condition exclusively. This finding is interesting with regard to the relationship between category salience and intergroup differentiation: Whereas the meta-analysis by Mullen and associates (1992) implies a linear relationship between these variables, our data stress the importance of a mediating subjective interpretation, that is, perceived category salience (see also Blanz, 1999). Here, the above-mentioned results on variances in the perception of intergroup similarities and differences might be read as a first hint that there is more consensus and less opportunity for subjective interpretation within the domain of negative as compared with positive stimuli (see also Mummendey, Berger, Buhl, & Otten, 1998; Mummendey & Otten, 1998).

Positive-Negative Asymmetry and Common Ingroup Identity In contrast to the findings of many other studies on intergroup behavior (e.g., Brewer, 1979; Brown, 1986), the variation in ingroup bias was not due to a variation in the evaluation of the ingroup but instead to a variation in the evaluation of the outgroup. In addition, there was a valence-specific asymmetry in the link between ingroup identification and ingroup favoritism, with strong positive correlations in the positive domain exclusively. This pattern of results, which cannot easily be accounted for by SIT, is in line with the recategorization model as postulated by Gaertner and colleagues (Gaertner, Dovidio, Anastasio, Bachman, & Rust, 1993; Gaertner, Mann, Murrell, & Dovidio, 1989; Gaertner, Rust, Dovidio, Bachman, & Anastasio, 1994). The recategorization process is characterized by three aspects: (a) a decrease in intergroup bias, based on (b) changes in the positivity of outgroup evaluations, whereas (c) the identification with the ingroup might still be high. Our data are consistent with Gaertner et al.’s (1993, 1994) findings concerning the representation of dual identities: They suggest that the development of a common ingroup identity does not necessarily require each group to forsake its subgroup identity completely. . . . Rather, generalization of benefits to additional outgroup members may be more likely to occur when the revised superordinate representation and the earlier group identities are salient simultaneously. (Gaertner et al., 1994, p. 245)

The work by Gaertner and associates, on one hand, and our findings concerning processes that may underlie positive-negative asymmetry in social discrimination, on the other hand, seem to approach the same phenomenon from two different angles: Work on the common ingroup identity model, on one hand, tries to clarify how well-known features of intergroup contact achieve their discrimination-decreasing effects. We, on the other hand, are searching for an explanation of why confrontations with negative instead of positive conditions make discrimination between groups disappear. A first explanato r y attemp t lead s to a co mmo n foc u s on recategorization processes. A perhaps noteworthy result is that neither ingroup identification nor the salience measures were correlated with ingroup favoritism in the negative valence/low salience condition. Although participants’ intergroup behavior was based on a new common ingroup identity, they might still have been aware of intergroup differences (or dissimilarities) and alternative levels of self-categorization. Correspondingly, Dovidio, Gaertner, and Validzic (1998) argue that there are conditions where a

Mummendey et al. / POSITIVE-NEGATIVE ASYMMETRY simultaneous recognition of both group difference and group communality is possible and beneficial for the reduction of intergroup biases. Conclusions In sum, results support a category-based account for the positive-negative asymmetry in social discrimination. More specifically, there is evidence that in intergroup settings that are minimal (i.e., do not provide clear-cut, objective information about evaluative differences between groups), the necessity to compare groups on negative evaluation dimensions reduces category salience and elicits a recategorization process on a higher level of inclusiveness. If we accept this conclusion, the question arises: Why does negative valence instigate recategorization? Currently, we can only speculate: Participants in the negative condition possibly experience a kind of common fate and are aware of some shared uneasiness about the task to make negative judgments on ingroup and outgroup. Furthermore, in line with the cognitive account for the asymmetry effect (see Otten et al., 1998), one could argue that a more careful, accurate information processing in the negative domain might decrease participants’ willingness to differentiate between groups that, from an objective viewpoint, are supposed to be very similar on the respective comparison dimensions. In other words, negative valence might raise doubts about whether the intergroup level is actually the appropriate one for self-categorization. Finally, some comments shall be made on the relevance of positive-negative asymmetry in social discrimination and its possible explanations in the domain of intergroup theories and research. Research on the positive-negative asymmetry in social discrimination provides more than just an additional variable in a long list of factors determining the link between mere categorization and social discrimination (see, e.g., Mummendey, 1995). In our opinion, it has crucial implications for the theoretical explanation of intergroup behavior in minimal intergroup situations as put forward by SIT (e.g., Tajfel & Turner, 1986). If one claims that striving for a positive social identity—based on positive ingroup distinctiveness in social comparisons—is the central motive for ingroup biases between minimal groups, then this motive should not only be relevant in comparisons involving positive resources. One might even argue that preventing the ingroup from any kind of negative treatment (allocation of burdens or characterization through negative traits) might be even more important for sustaining or establishing a positive social identity than claiming its superiority on positive comparison dimensions. Therefore, although the present article addressed the question, “What is specific about negative

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stimuli in intergroup allocations and evaluations?” the time has come to ask, “What is so specific about positive stimuli in intergroup allocations?“ and to substitute or to supplement the traditional interpretation of mere categorization effects in terms of a striving for positive distinctiveness with other explanatory approaches. As argued by Forgas and Fiedler (1996), it seems worthwhile to take into account that ingroup favoritism based on minimal categorizations might be strongly linked to a weakly reflected decision-making process, which is provided by positive rather than by negative stimuli. NOTES 1. It is important to state here that the term “low salience” has to be understood in relative terms: Salience is described as low as compared with high salience but is supposed to be still sufficiently high to elicit positive ingroup differentiation in the positive domain, at least. 2. The results for the reduced sample (excluding participants failing on the manipulation checks and 2 participants with out-of-range values; n = 47) basically equal those of the full sample. With respect to discrimination, a 2 × 2 ANOVA (salience valence) shows a marginally significant main effect for salience, F(1, 43) = 3.08, p < .10, and a marginally significant interaction between salience and valence, F(1, 43) = 3.65, p < .10. With respect to identification and causal attributions, there were neither main effects nor an interaction between salience and valence. Consistent with the results for the full sample, contrast analyses revealed highly significant contrasts for discrimination, t(43) = 3.25, p < .01, dcon = .99, and for causal attribution, t(43) = 2.09, p < .05, dcon = .64. 3. Note that both possible results are clearly compatible because the weights of the single contrast (1, –3, 1, 1) are simply the sum over the three orthogonal contrasts for both main effects (–1, –1, 1, 1; 1, –1, –1, 1) and the interaction (1, –1, 1, –1). 4. The degrees of freedom are due to the unequal variances of groups. 5. Most recently, Gardham and Brown (in press) corroborated our results concerning the valence effect on perceived salience.

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