Social identity and perceived group homogeneity: Evidence for the ingroup homogeneity effect

European Journal of Social Psychology, Vol. 20,269-286 (1990)

Social identity and perceived group homogeneity: Evidence for the ingroup

homogeneity effect

BERND SIMON Westfalische Wilhelms-Universitat, Munster, Federal Republic of Germany

and

THOMAS F. PETTIGREW University of California, Santa Cruz, U.S.A., and University of Amsterdam, The Netherlands

Abstract

On the basis of social identity theory, we argue that the search for a positive social identity is characterized by the accentuation of perceived ingroup homogeneity relative to perceived outgroup homogeneity (the ingroup homogeneity eflect). To test our specific hypotheses, we conducted an experiment in which some subjects were provided with a well-defined group membership and others were allocated to ill-defined groups. We also manipulated the information about the relative sizes of the groups. Subsequently, several measures ofperceived dimensional andgeneral group homogeneity were administered. As predicted, members of well-defined groups revealed the ingroup homogeneity effect for an attribute associated with the definition of their group. On the other hand, members of ill-defined groups displayed social creativity and selected alternative attributes to accentuate the relative ingroup homogeneity. Moreover, when intergroup comparisons did not contribute positively to the self-concept, subjects accentuated their positive personal identity and evinced the outgroup homogeneity effect. Consistent with previous research, subjects identified more strongly with a well-defined group when it was a minority rather than a majority. Minority members also perceivedmore general homogeneity within the ingroup than within the outgroup, whereas the majority members showed the opposite eflect. Finally, the interrelationship between personal and social identity is briefly discussed.

This article is based on a paper presented at the First European Congress of Psychology, Amsterdam (Holland), July 1989. The research reported here was conducted while Bernd Simon was a visiting scholar at the University of California, Santa Cruz. During that time, the first author was financially supported by the Friedrich-Ebert-Stiftung, Bonn. We are grateful to Rex Zimmermann and Howard Nussbaum for their assistance during the experiment.

0046-2772/90/040269-18$09.00 0 1990 by John Wiley & Sons, Ltd.

270 B. Simon and T. F. Pettigrew

INTRODUCTION

Previous research on the perception of group homogeneity has been guided largely by the outgroup homogeneity hypothesis which predicts that the outgroup is generally seen as more homogeneous than the ingroup (the outgroup homogeneity effect; e.g. Linville and Jones, 1980; Quattrone and Jones, 1980). This prediction has held up in many studies and various explanations have been proffered for the outgroup homogeneity effect (Jones, Wood and Quattrone, 198 1; Linville, Salovey and Fischer, 1986; Linville, Fischer and Salovey, 1989; Messick and Mackie, 1989; Quattrone, 1986; Wilder, 1986). Some writers argue that the ingroup is seen as less homogeneous than the outgroup because people are motivated to assert their own individuality and free themselves from constraining expectations simultaneously (e.g. Quattrone and Jones, 1980; Wilder, 1986). A second line of reasoning maintains that the outgroup homogeneity effect occurs because it justifies intergroup discrimination (Wilder, 1984, 1986). Finally, people might interact with ingroup members more frequently and in a greater variety of contexts than with outgroup members. Hence, greater familiarity with the ingroup should lead to greater differentiation among ingroup members than among outgroup members (Linville et al., 1986, 1989). This argument might even apply when the amount of knowledge about ingroup members (except self) and outgroup members is held constant (Judd and Park, 1988; Wilder 1984), because people might still have greater knowledge about the ingroup because of their superior knowledge about particular ingroup exemplars - namely themselves (Linville et al., 1989).

However, there is now growing evidence which throws doubt on the outgroup homogeneity effect as the ‘general rule’ and the explanations just listed. Simon and Brown (1987) found that experimentally created minorities assumed more ingroup than outgroup homogeneity (ingroup homogeneity effect), while non-minorities showed the opposite effect. A field study by Brown and Smith (1989) and a meta- analysis by Mullen and Hu (1989) have confirmed the ingroup homogeneity effect for minority members.

Based on social identity theory (Tajfel and Turner, 1986; Turner, Hogg, Oakes, Reicher and Wetherell, 1987; Hogg and Abrams, 1988), Simon and Brown (1987) argued that the search for a positive social identity can lead not only to ingroup favouritism (Brewer, 1979; Brown, Tajfel and Turner, 1980; Turner, 1981), but also to the establishment of greater ‘groupness’ of the ingroup compared to the outgroup. For, given that the ingroup is ascribed more groupness - i.e. is construed more in terms of a group - than the outgroup, the ingroup is likely to be seen as superior to the outgroup regarding the extent of social support and solidarity it offers its members (Allen, 1985, pp. 230-231). Following Campbell (1958), perceiving more ingroup than outgroup homogeneity should enhance the perceived groupness or ‘entitativity’ of the ingroup compared to that of the outgroup. Consequently, the accentuation of perceived ingroup homogeneity relative to perceived outgroup homogeneity will also contribute to a positive social identity.

Thus, when social identity is salient, group members may emphasize relative ingroup homogeneity at the expense of their own and other ingroup members’ individuality or uniqueness, which is otherwise often pursued (Codol, 1984, 1987; Snyder and Fromkin, 1980; Quattrone and Jones, 1980). Prior research indicates that this might be the case especially for members of dominated groups or minorities

Group homogeneity 27 1

(Deschamps, 1982a; McGuire and McGuire, 1988; Simon and Brown, 1987). More- over, contrary to Wilder’s (1986) assumption, the study by Brown and Smith (1989) suggests that ingroup favouritism can be associated with the ingroup homogeneity effect (see also Simon and Brown, 1987, p. 708). Also, the familiaritydifferentiation argument by Linville et al. (1986, 1989) is contradicted. Minority members in the study by Simon and Brown (1987) perceived the ingroup as more homogeneous than the outgroup in spite of identical knowledge about outgroup members and (other) ingroup members and superior knowledge about themselves.

The main aim of the present study is to further investigate Simon and Brown’s (1987) thesis that the search for a positive social identity can also lead to the perception of more ingroup than outgroup homogeneity. Yet we extend their thesis in one respect and specify it in another. Simon and Brown (1987) predicted and found stronger social identification in the minority ingroup than in the non-minority ingroup and the ingroup homogeneity effect in the former and the outgroup homogeneity effect in the latter condition. Given that minority members are more likely to define themselves in terms of their group membership or social identity than members of non-minorities (see also McGuire and McGuire, 1988), the ingroup homogeneity effect might indeed be more pronounced when the ingroup is in the minority position. But we assume that the ingroup homogeneity effect as such is not necessarily restricted to minorities.

However, we do not expect the ingroup homogeneity effect to occur irrespective of the specific attribute (or dimension) in reference to which homogeneity judgements are made, neither for minority nor for non-minority groups. Research on ingroup favouritism has demonstrated that group members are highly selective and often creative as to the dimensions they choose for discriminatory intergroup comparisons. For instance, the ingroup is usually most strongly favoured on ingroup-specific dimensions, i.e. on dimensions relevant to ingroup definition and identity (Mummendey and Schreiber, 1983, 1984; Mummendey and Simon, 1989; van Knippenberg and Ellemers, 1989). And new ingroup-specific dimensions are often introduced by group members (Lemaine, 1974; Tajfel, 1978). Accordingly, we also expect the ingroup homogeneity effect mainly on ingroup-specific dimensions or attributes (see also Kelly, 1989). Also, given an ill-defined ingroup lacking in distinctive attributes, group members should be motivated to search for attributes or dimensions on which to accentuate relative ingroup homogeneity or groupness.

To test these assumptions, we use a variant of the minimal group paradigm (see Tajfel, Billig, Bundy and Flament, 1971) that differs from the original paradigm in several ways. First, the minimal group paradigm usually involves a symmetrical social categorization into two distinct groups, both of which are explicitly defined (e.g. Klee group versus Kandinsky group). In the present experiment, we employ an asymmetrical social categorization, where only one group is explicitly defined (well-defined group, i.e. the ‘Wheeler group’) and the existence of another group (ill-defined group) can only be inferred (i.e. all people not belonging to the ‘Wheeler group’). Given the emphasis on membership or non-membership in the well-defined group, the experimental situation is very likely to prime self-definition in terms of group membership. Thus, social identity is made salient for all subjects irrespective of whether they are allocated to the well-defined group or not. Consequently, we expect members of the well-defined group to emphasize the comparatively greater ‘groupness’ or homogeneity of the ingroup regarding ingroup-specific attributes,


i.e. attributes associated with the definition of their group. On the other hand, members of the ill-defined group should be motivated to accentuate the relative ingroup homogeneity regarding alternative attributes not associated with the well- defined (out)group. Second, we also manipulate the information subjects are given about the relative sizes of the groups. Third, our study includes control conditions in which outside observers not belonging to groups are asked to provide estimates of homogeneity of two stimulus groups. Again, only one group is well-defined, and we manipulate the information about the relative sizes of the groups. In this manner, we assess the effects of our main independent variables (group definition and information about the relative group size) independent of the perceiver’s involvement as a member of one of the groups.

Finally, though not directly bearing on the perception of group homogeneity, another modification of the minimal group paradigm is included in order to demon- strate the potential contribution of social identity to a positive self-concept. In the present experiment subjects who are categorized into groups not only rate the ingroup and outgroup on various dimensions, but they also rate themselves on the same dimensions. We expect that, when the intergroup situation allows for ingroup superiority (i.e. for positive social identity), subjects will rely less on the contribution to a positive self-concept of their personal identity (Turner, 1978, 1982). Thus, they will feel less need to evaluate themselves positively when the ingroup is superior than when it is inferior to the outgroup.

Since we use minimal groups within and between which no face-to-face interaction had taken place, subjects must be assumed to be either equally unfamiliar with both ingroup and outgroup or - due to their superior knowledge about themselves - more familiar with the ingroup. Thus, the occurrence of the ingroup homogeneity effect would be most problematic for the theoretical model proposed by Linville et al. (1986, 1989) which assumes that category differentiation is directly related to category familiarity. To facilitate comparability further, we choose as one of our dependent measures the Pd measure (probability of differentiation) defined by Linville et al. (1986, 1989). Pd reflects the probability that a perceiver will differentiate between two randomly chosen instances of the category (or group members) in terms of the dimensional attribute in question (for further details see Results section). In addition to this measure of dimensional homogeneity (heterogeneity), we also apply measures of general homogeneity (Quattrone, 1986).

To summarize, our specific hypotheses are as follows: (1) Members of well-defined groups will ascribe more homogeneity to the ingroup

than to the outgroup regarding attributes associated with the well-defined group. (2) Members of ill-defined groups will show the ingroup homogeneity effect regard-

ing alternative attributes. (3) Minority members will be more likely than majority members to show the

ingroup homogeneity effect. (4) Minority members will identify more strongly with their ingroup than majority

members. ( 5 ) Group members will rate themselves less favourably individually on a dimen-

sion when the ingroup is superior to the outgroup on that particular dimension than when the ingroup is inferior.

Group homogeneity 273

METHOD

Subjects

One hundred and ninety-one undergraduates at the University of California, Santa Cruz (129 female and 62 male students), mean age 20 years (standard deviation: 3.8), volunteered as subjects for course credit.

Design

The experimental conditions yielded a 2 x 3 x 2 factorial design with two between- subjects factors, group definition (well-defined ingroup versus ill-defined ingroup) and information about the relative ingroup size (minority versus majority versus no information), and one within-subjects factor, stimulus group (ingroup versus outgroup). In the control conditions subjects were not allocated to any group. Thus, these conditions yielded a 3 x 2 factorial design with one between-subjects factor, information about the relative size of the well-defined group (minority versus majority versus no information), and one within-subjects factor, stimulus group (well-defined group versus ill-defined group).

Procedure

Experimental conditions

The experiment was introduced as a study of the relationship between aesthetic taste and personality. Subjects first viewed 18 slides of modern paintings and indicated - on eleven-point rating scales (0 =not at all to 10 = very much) - how much they liked each painting. The painter of each painting remained anonymous. While the experimenter and two assistants were then ostensibly busy analysing subjects’ responses, subjects worked on a filler task (i.e. describing two other paintings).

After about 15 minutes, a cover story was given that some of the paintings had been painted by a modem painter named J. P. Wheeler.’ The cover story also indicated that research evidence strongly suggests that the degree of liking for the Wheeler paintings relates to other personality characteristics. It was announced that those participants had been identified whose liking for the Wheeler painting was above a certain criterion. Those people would be called the Wheeler group (well- defined group). The experimenter emphasized that ‘to belong to the Wheeler group one does not necessarily have to show an extreme degree of liking for the Wheeler paintings, but rather it is sufficient to be just above a certain standard’. People below the cut-off point were not referred to as members of any specific group (ill- defined group). Some subjects were then told that many studies had reliably shown that either about 10 per cent or about 90 per cent of all people belong to the Wheeler group (minority or majority conditions); other subjects were given no information regarding the Wheeler group percentage (no information condition).

Each subject was then given a questionnaire containing the dependent measures. The first page of the questionnaire also indicated whether the subject belonged or

’ This name was fictitious. In fact, all paintings used in this experiment were painted by Paul Klee. More details regarding these paintings are available from the first author.


did not belong to the Wheeler group. Subjects had to endorse this information on all following pages. The relative size of the Wheeler group was varied across sessions. Membership or non-membership in the Wheeler group was, of course, randomly assigned to the subjects within each session. The experimental conditions were run in nine sessions. The number of subjects participating in a session ranged from nine to 19.

Control conditions

Although they were told the same cover story about the Wheeler group, subjects in these conditions were not themselves categorized into groups at any stage of the procedure. They viewed the same slides as subjects in the experimental conditions, but they never had to indicate their reactions to the slides. The information regarding the relative size of the Wheeler group was manipulated just as in the experimental conditions. Control subjects were also given questionnaires requiring estimates of the homogeneity of two stimulus groups (i.e. of the well-defined or Wheeler group and the ill-defined group). The rationale provided for the questions was the experimenter’s interest in subjects’ predictions of the reactions of people ‘who actually took part in our original experiments’. The control conditions were run in four sessions. The number of participants in a session ranged from 8 to 23.

Dependent measures

Self-ratings

Subjects in the experimental conditions (but not those in the control conditions) were asked for self-ratings on four dimensions: They indicated how much they liked two additional paintings shown to them, first one painting allegedly painted by J. P. Wheeler (WHE dimension) and then another painting allegedly not painted by Wheeler (NWH dimension). The ‘actual’ painter of the second painting was not revealed. Subsequently, subjects indicated how much they liked art in general (ART dimension) and how sociable they saw themselves (SOC dimension; for all dimensions: 0 = not at all to 10 = very much).

Frequency distributions

For each of the four dimensions mentioned above, all subjects (including control subjects) formed two separate frequency distributions, one for people belonging and another for people not belonging to the Wheeler group. For example, subjects were asked to imagine 100 people belonging (and 100 people not belonging) to the Wheeler group, and to estimate how those would like the Wheeler painting (WHE dimension). For that purpose, a series of 11 boxes was provided numbered from 0 (not at all) to 10 (very much). Subjects then estimated how many of the 100 people belonging (not belonging) to the Wheeler group would fall into each of the 11 boxes. The same procedure was followed for the other three dimensions, i.e. liking for the non- Wheeler painting (NWH), liking for art in general (ART) and sociability (SOC). Subjects in experimental conditions always started with the self-ratings on a dimension and then formed the two distributions before continuing with the next dimension.


The first dimensional attribute (WHE) was by definition associated with the Wheeler group. Because preference for the Wheeler paintings might also signify an affinity for art more generally we expected subjects to associate the third attribute (ART) with the Wheeler group as well. No associations with the Wheeler group as such were expected for the other two attributes (NWH and SOC). Yet since minority status is often associated with deviance and majority status with normality (see Moscovici, Mugny and van Avermaet, 1985), we assumed that sociability would be associated with majority rather than minority status.

From each frequency distribution, we later calculated the mean and the probability of differentiation (Pd measure). While the former index allows for a check of the expected group-attribute associations, the latter is an index of perceived dimensional homogeneity or heterogeneity.

Predictability, intragroup similarities and diferences

All subjects rated how easy it would be to make predictions about the personality of someone who belongs (does not belong) to the Wheeler group. Additionally, each subject estimated how similar to each other and - on a separate unipolar scale - how different from each other people belonging (not belonging) to the Wheeler group are. The order of the two questions - one asking for assumed similarities, the other for assumed differences - was randomized across subjects. All estimates regarding the two stimulus groups (i.e. people belonging and not belonging to the Wheeler group) were made on separate scales. The respective order was randomized across subjects.

Importance of group membership Subjects in the experimental conditions were also asked an additional question: Sub- jects allocated to the Wheeler group indicated how important it was to them to be a member of that group. On the other hand, subjects not allocated to the Wheeler group indicated how important it was to them not to belong to that group. This modification was necessary because it was impossible to refer to the latter subjects’ ingroup without defining it explicitly and invalidating the asymmetrical social categorization. Ratings were made on 1 l-point scales (0 = not at all to 10 = very).

After the questionnaires had been returned, subjects in the experimental and control conditions who had not been given information about the percentage of people usually belonging to the Wheeler group (no information condition) estimated ‘what percentage of the American population belongs to the Wheeler group’. Finally, all subjects were given another sheet asking for demographic data (gender, age, etc.) and for a description of the perceived purpose of the study. After the experiment had been completed, participants were fully debriefed.

RESULTS

Subjects’ description of the perceived purpose of the study revealed that 13 subjects doubted the existence of the Wheeler group and one determined the real purpose of the study. Moreover, three subjects did not understand how to form the frequency distributions. These 17 subjects (less than 9 per cent) were excluded from the analyses, with exclusions ranging from one to three per cell. The final sample size was 114


in the experimental conditions (18 to 21 per cell) and 60 in the control conditions (19 to 21 per cell). Preliminary analyses of variance revealed that none of the significant ANOVA effects reported in this study was significantly qualified by gender. Hence, the reported analyses include both women and men.

Manipulation check Perceived groupattribute associations Since control subjects were not involved as members of one of the two stimulus groups, their ratings provide a manipulation check regarding the group-attribute associations. For each control subject, the mean was calculated from the frequency distribution on each dimension for each stimulus group. Three (relative size of well- defined group) x 2 (stimulus group) ANOVAs were performed on these data. Four significant ANOVA effects emerged. As expected the attributes, liking for the Wheeler painting (WHE) and for art in general (ART), were significantly associated with the well-defined or Wheeler group (WHE) M(wel1) = 6.9, M(il1) = 4.0, F(1,57) = 101.83, p c 0.0005; (ART) M(wel1) = 6.9, M(il1) = 5.6, F(1,57) = 36.38, p < 0.0005).

On the sociability dimension (SOC), the interaction effect was significant (F(2,57) = 6 . 8 7 , ~ < 0.005, MSe (within subjects) = 1.26). When there was an explicit minority-majority relationship (i.e. in the minority and majority conditions), the group in the majority position was rated as more sociable than the group in the minority position irrespective of whether it was well-or ill-defined (M(wel1, min) = 5.2, M(il1,maj) = 6.2, t(57) = -2.75, p < 0.005; M(wel1,maj) = 6.1, M(il1,min) = 5.3, 457) = 2.25, p < 0.05, all t-tests are one-tailed unless otherwise noted). In the no information condition, the ill-defined group was seen as more sociable than the well-defined group (M(wel1,no info) = 5.3, M(il1,no info) = 5.9, t(57) = - 1.73, p c 0.05). And in this condition, the well-defined group was apparently seen in the minority position; the mean percentage estimate for the size of this group was significantly smaller than 50 per cent (M = 25.1, t(20) = 6 . 7 2 , ~ < 0.001). In sum, as expected sociability was associated with majority, rather than minority, groups. Finally, though not intended by the experimenters, control subjects perceived an association between liking for the non-Wheeler painting (NWH) and the ill-defined group (M(wel1) = 4.7, M(il1) = 5.8, F( 137) = 17.50, p c 0.0005).

The mean ratings from the experimental conditions together with all significant ANOVA effects are presented in Table 1 . The association between liking for the Wheeler painting and the well-defined (or Wheeler) group was replicated. On the other hand, experimental subjects’ ratings on the ART and SOC dimensions did not reflect the group-attribute associations perceived by control subjects. Yet given that these dimensions are value loaded, it is consistent with the phenomenon of ingroup favouritism that members of the ill-defined group or the minority were reluc- tant to ascribe more liking for art in general or more sociability to the outgroup (well-defined group or majority) than to the ingroup.

Perceived group homogeneity

Dimensional homogeneity: Diferentiation within groups For each subject, the probability of differentiation (Pd) was calculated from the frequency distribution on each dimension for each stimulus group. The Pd measure

Tabl

e 1.

Mea

n ra

tings

(Me)

and

pro

babi

lity

of d

iffer

entia

tion

(Pd)

for i

ngro

up a

nd o

utgr

oup

in e

xper

imen

tal c

ondi

tions

Wel

l-def

ined

(Whe

eler

) ing

roup

R

elat

ive

ingr

oup

size

Ill-d

efin

ed (n

on-W

heel

er) i

ngro

up

' R

elat

ive

ingr

oup

size

M

inor

ity

Maj

ority

N

o in

form

atio

n M

inor

ity

Maj

ority

N

o in

form

atio

n (n

= 1

8)

(n =

18)

(n

= 2

1)

(n =

18)

(n

= 1

9)

(n =

20)

In

grou

p O

utgr

oup

Ingr

oup

Out

grou

p In

grou

p O

utgr

oup

Ingr

oup

Out

grou

p In

grou

p O

utgr

oup

Ingr

oup

Out

grou

p

WH

E M

e 5.

7 *

4.4

6.4

* 3.

7 5.

8 *

3.8

4.4

* 6.

8 3.

5 *

6.4

4.8

* 6.

4 Pd

0.

79

0.79

0.

79

0.78

0.

79

0.80

0.

78

0.78

0.

77

* 0.

82

0.78

0.

79

Me

5.1

5.2

4.9

5.3

5.0

4.9

5.5

4.5

4.7

4.8

5.3

5.3

Pd

0.82

0.

83

0.80

0.

81

0.79

0.

81

0.79

*

0.82

0.

83

0.82

0.

81

0.83

Me

6.0

5.3

6.2

5.7

7.0

* 5.

3 6.

5 6.

2 6.

4 *

5.7

6.3

6.7

Pd

0.81

0.

82

0.73

*

0.78

0.

72

* 0.

77

0.80

0.

80

0.82

0.

79

0.77

*

0.71

Me

5.6

5.6

6.0

5.9

6.0

* 5.

2 5.

8 5.

9 6.

1 *

4.4

6.2

6.0

Pd

0.84

0.

84

0.79

0.

78

0.78

0.

80

0.79

0.

78

0.79

*

0.82

0.

81

* 0.

77

NW

H

AR

T

SOC

WH

E =

liki

ng fo

r the

Whe

eler

pain

ting,

NW

H =

liki

ng fo

r the

non

-Whe

eler

pai

ntin

g, A

RT

= li

king

for a

rt in

gen

eral

, SO

C =

soci

abili

ty.

*Ing

roup

and

out

grou

p sc

ores

(pe

r m

easu

re,

dim

ensi

on a

nd c

ell)

diff

er a

t th

e 0.

05 le

vel

of s

igni

fican

ce o

r be

tter

(one

-taile

d t-t

ests

bas

ed o

n m

ean

squa

re e

rror

(w

ithin

subj

ects

)).

Sign

ifica

nt A

NO

VA

effe

cts:

Mea

n ra

tings

: St

imul

usgr

oup(

SG):

AR

TF(

1,10

8) =

11.

56,~

< O.O

Ol;S

OC

F(1,

108)

= 6

.17

,~ < 0.

05.

Gro

up de

finiti

on (G

D):

WH

E F(

1,10

8) =

6.0

3,~

< 0.

05.

GD

X S

G W

HEF

(1,1

08)

= 6

8.66

,~ < 0.

0005

; AR

T 0

1,10

8) =

4.8

6,~

< 0.

05.

Rel

ativ

e in

grou

p si

ze (R

S) x

GD

: SO

C F

(2,1

08) =

4.76

, p <

0.01.

R

S X

GD

X S

G A

RT

02,

108)

= 4

.07

,~ < 0.

05; S

OC

02,

108)

= 3

.30

,~ < 0.

05.

MSe

(bet

wee

n su

bjec

ts):

WH

E 1.

53, N

WH

1.4

4, A

RT

2.66

, SO

C I .

69.

MSe

(with

in su

bjec

ts):

WH

E 3.

85, N

WH

3.5

6, A

RT

1.6

4, S

OC

1.9

2.

Pd m

easu

re:

SG: N

WH

01,

108)

= 6

.68

,~ < 0.

05.

GD

X S

G: A

RT

F(1,

108)

= 1

3.65

,~ < 0.

0005

. R

S X

GD

X S

G: A

RT

02,1

08) =

3.2

1,~

< 0.

05; S

OC

02,

108)

= 6

.60

,~

< 0.

005.

M

Se (b

etw

een

subj

ects

): W

HE

0.02

1, N

WH

0.0

09, A

RT

0.03

1, S

OC

0.01

4.

MSe

(with

in su

bjec

ts):

WH

E 0.

007,

NW

H 0

.001

, AR

T 0.

004,

SO

C 0.

002.

A

ll ot

her e

ffect

s wer

e non

-sig

nific

ant.


n

I = 1 is defined by Pd = 1 - Pi2 where i denotes the level of the dimensional attribute

in question and Pi denotes the probability for the i-th level of the attribute. For example, when a subject used three boxes (levels) in which she entered the numbers 40, 30, and 30 respectively, Pd then equals 1 - (0.42 + 0.32 + 0.33) = 0.66. S’ ince we used 11 point scales, Pd could theoretically range from 0 to 0.909. A higher value indicates a higher probability of differentiation or less perceived homogeneity.*

Hypothesis I We expected members of well-defined (or Wheeler) groups to show the ingroup homogeneity effect regarding attributes associated with their groups’ definition (WHE and ART). This prediction was supported on the ART, but not on the WHE dimension (see Table 1 for means and ANOVA effects). For ART the two-way interaction between group definition and stimulus group was highly significant (F(1,108) = 13.65, p < 0.0005). Members of well-defined groups showed the expected ingroup homogeneity effect (M(in) = 0.75, M(out) = 0.79, t(108) = -2.79, p < 0.005). In addition, a three-way interaction indicates (F(2,108) = 3.21, p < 0.05) that the ingroup homogeneity effect was largely due to subjects in the majority (t(l08) = -2.18, p < 0.05) and no information (t(l08) = -2.31, p < 0.05) conditions. Conversely, an outgroup homogeneity effect was found in the ill-defined, no information condition (t(lO8) = 3 . 1 5 , ~ < 0.005).

Hypothesis 2 We also predicted that members of ill-defined groups would show the ingroup homogeneity effect for the alternative attributes not associated with the well-defined (or Wheeler) group. Since liking for the Wheeler painting (WHE) and for art in general (ART) were associated with the well-defined group, liking for the non-Wheeler painting (NWI-I) and sociability (SOC) were such alternatives. However, since sociability was obviously associated with majority status, only members of the ill-defined majority were expected to show the ingroup homogeneity effect on the SOC dimension.

Our predictions were supported for the ill-defined minority and majority (see Table 1). Although on the NWH dimension, the ingroup was generally seen as somewhat more homogeneous than the outgroup (M(in) = 0.81, M(out) = 0.82, F(1,108) = 6.68, p < 0.05), this difference was due largely to subjects in the ill-defined, minority condition (t(l08) = -2.34, p < 0.05). On the SOC dimension, the three-way interaction was significant (F(2,108) = 6.60 ,~ < 0.005). There was the expected ingroup homogeneity effect in the ill-defined, majority condition (t( 108) = - 1.74, p < 0.05), but also an outgroup homogeneity effect in the ill-defined, no information condition (t(l08) = 3 . 0 3 , ~ < 0.005).

* Other researchers (see Judd and Park, 1988) used the standard deviation (S.D.) of the frequency distribution formed by each subject as the measure of perceived homogeneityiheterogeneity. In a comparative analyses of S.D. and Pd, Linville et al. (1986) concluded that the latter appears to be the ‘purer measure of stereotypic thinking and perceived degree of heterogeneity’ (p. 189). These authors also point out that both measures will be positively correlated if the perceiver’s frequency distribution is ‘bell-shaped’. Since our subjects were psychology students who were familiar with the notion of a normal distribution, they could be expected to form bell-shaped frequency distributions. In fact, S.D. and Pd correlated positively for all four dimensions in the experimental conditions. The coefficients ranged from 0.63 to 0.79 (allps < 0.0005, one-tailed). In the control conditions the coefficients were somewhat lower, specs- cally on the ART dimension where only the correlation coefficient regarding the well-defined group reached an acceptable level of statistical significance (dwell-defined group) = 0.28, p < 0.05, o n e - a d , r(il1-defined group) = 0.07, ns). On the other three dimensions the coefficients were again significantly positive. They ranged from 0.41 to 0.60 (all ps < 0.001, one-tailed). As one could expect analyses using S.D. as the dependent measure yielded patterns of means very similar to those presented above.


Turning to the control conditions, 3 (relative size of well-defined group) x 2 (stimulus group) ANOVAs revealed that control subjects saw the well-defined group generally as less homogeneous than the ill-defined group on the NWH dimension (M(wel1) = 0.83, M(il1) = 0.81, F(1,57) = 7.04, p 5 0.01), but as more homogeneous on the ART dimension (M(wel1) = 0.76, M(il1) = 0.80, F(1,57) = 8.04, p < 0.01). No other ANOVA effect was significant ((WHE) M(wel1) = 0.76, M(il1) = 0.78; (SOC) M(wel1) = 0.82, M(il1) = 0.82).

Analyses of covariance controlling for rating extremity (with the absolute distance from the mean of the frequency distribution to the scale midpoint as the covariate) replicated all effects in the experimental conditions, but only the main effect on the NWH dimension in the control conditions.

General homogeneity: Predictability and variability

The two questions that directly asked for assumed intragroup similarities and differences yielded virtually identical results. Hence, for each subject the respective responses were averaged. Thus, in addition to the predictability measure we obtained a single measure for the variability ascribed to each stimulus group. (Of course, the similarity ratings had been re-coded accordingly).

Hypothesis 3 Unlike the measure of dimensional homogeneity (Pd), the two measures of general homogeneity provide clear evidence that minority members are more likely than majority members to show the ingroup homogeneity effect (see Figure 1).

In 2 x 3 x 2 ANOVAs the within-subjects factor, stimulus group, interacted significantly with relative ingroup size on both measures ((PRE) F(2,108) = 15.8 1, p < 0.0005; (VAR) F(2,108) = 9 . 5 0 , ~ < 0.0005). Consistent with Simon and Brown’s (1987), results, minority members saw their ingroup as more predictable and as less variable than the outgroup ((PRE) M(in) = 3.4, M(out) = 2.4, t(108) = 2.93, p < 0.005; (VAR) M(in) = 6.3, M(out) = 7.0, t(lO8) = -2.10, p < 0.05), whereas majority members showed the opposite effects ((PRE) M(in) = 2.2, M(out) = 3.8, t(108) = -4.76, p < 0.001; (VAR) M(in) = 7.4, M(out) = 6.0, t(lO8) = 4.18, p < 0.001). Subjects in the no information condition showed no effect at all ((PRE) M(in) = 4.1, M(out) = 3.9, t(108) < 1, ns; (VAR) M(in) = 6.2, M(out) = 6.0, t(108) < 1, ns).

The two-way interaction between stimulus group and group definition was also significant for both measures ((PRE) F(1,108) = 21.41, p < 0.0005; (VAR) F(1,108) = 1 1 . 0 4 , ~ < 0.001). The well-defined ingroup was seen as more predictable, but not as less variable than the ill-defined outgroup ((PRE) M(in) = 3.9, M(out) = 3.1, t(108) = 2.95, p < 0.005; (VAR) M(in) = 6.4, M(out) = 6.8, t(108) = - 1.48, ns). The ill-defined ingroup was seen as both less predictable and more variable than the well-defined outgroup ((PRE) M(in) = 2.7, M(out) = 3.7, t(108) = -3.69, p < 0.001; (VAR) M(in) = 6.7, M(out) = 5.8, t(108) = 3.33, p < 0.001). All other ANOVA effects were non-significant.

Control subjects also ascribed more predictability and less variability to the well- defined group than to the ill-defined group ((PRE) M(wel1) = 4.8, M(il1) = 3.2, F(1,57) = 23.40, p < 0.0005; (VAR) M(wel1) = 5.4, M(il1) = 7.1, F(1,57) = 29.57, p < 0.0005). Unlike experimental subjects, however, they did not generally ascribe less predictability and more variability to the majority than to the minority group:


4 Predictability

3.8

7 2.4

Variabi l i ty

7.0 7

3

2 :i 0 1

. .-

-ma- Mlnorlty

-1 lngroup

Outgroup 4.1

Majority No Information

Figure 1. Predictability and variability ascribed to ingroup and outgroup as a function of the information about the relative ingroup size

Although stimulus group and relative size of well-defined group interacted significantly on the second measure ((PRE) F(2,57) = 1.26, ns; WAR) (F(2,57) = 5.33, p < 0.01), there was no consistent tendency for the majority to be perceived as more variable than the minority (M(wel1,min) = 4.9, M(il1,maj) = 7.7, t(57) = - 5.10, p < 0.001; (M(wel1,maj) = 5.9, M(il1,min) = 6.2, t(57) = -0.56, ns; M(wel1,no info) = 5.4, M(il1,no info) = 7.2, t(57) = - 3 . 4 4 , ~ < 0.001).

Importance of group membership

The importance ratings given by members of well-defined (or Wheeler) groups (importance of belonging to the well-defined group) and those given by members of ill-defined groups (importance of not belonging to the well-defined group) were analysed in separate three-level one-way ANOVAs (factor: relative ingroup size).

Hypothesis 4 We expected minority members to identify more strongly with their ingroup and therefore to attach more importance to their group membership than

Group homogeneity 28 1

majority members. For well-defined groups, the effect of relative ingroup size was significant (F(2,54) = 3 . 8 5 , ~ < 0.05). As expected, minority members attached greater importance to their group membership than majority members, while the no information condition did not differ significantly from either the minority or majority conditions (M(min) = 2.1, M(maj) = 0.5, M(no info) = 1.2, Newman-Keuls procedure,p I 0.05, MSe = 3.04). For ill-defined groups the overall effect was not significant (F(2,54) = 1.62, ns, M(min) = 2.5, M(maj) = 4.3, M(no info) = 2.7, MSe = 11.12).

Self-rating

Hypothesis 5 According to the control data, the well-defined (or Wheeler) group was superior to the ill-defined group on the ART dimension and the majority was superior to the minority on the SOC dimension. Thus, given that these dimensions are value loaded, well-defined group members were expected to rate themselves less favourably individually on the ART dimension than ill-defined group members, because the former could already count on (a priori) ingroup superiority. By the same token, majority members should show lower self-ratings on the SOC dimension than minority members.

These expectations were clearly confirmed in 3 x 2 ANOVAs. On both dimensions, one significant ANOVA effect emerged. On the ART dimension, members of well- defined groups gave significantly lower self-ratings than members of the ill-defined groups (M(wel1) = 7.7, M(il1) = 8.5, F(1,108) = 6.72, p < 0.05). On the SOC dimension, the main effect for relative ingroup size was significant (F(2,108) = 3.94, p < 0.05). Self-ratings in the majority condition were significantly lower than self- ratings in each of the other two conditions which did not significantly differ from each other (M(maj) = 6.1, M(min) = 7.1, M(no info) = 7.3, MSe = 4.08, Newman- Keuls procedure, p I 0.05).

Though not directly relevant to hypothesis 5 due to their lack of explicit value connotations, the self-ratings on the other two dimensions (WHE and NWH) were also analysed. A significant main effect for relative ingroup size emerged on the WHE dimension (F(2,108) = 4 . 6 5 , ~ < 0.05). Although the interaction effect remained only suggestive (F(2,108) = 2 . 5 2 , ~ = 0.085; on all other dimensions: Fs(2,108) I 1.17, ns), the pattern of the means indicates that the main effect was due to the low liking for the Wheeler painting in the ill-defined, majority condition. Comparisons within each level of the group definition factor revealed that the majority condition differed significantly from each of the other two conditions when the ingroup was ill-defined (M(maj) = 3.8, M(min) = 6.1, M(no info) = 6.3), but there was no significant difference at all when the ingroup was well-defined (M(maj) = 5.9, M(min) = 5.7, M(no info) = 6.8, MSe = 5.79, Newman-Keuls procedure, p I 0.05). The minority and no information conditions did not differ significantly from each other within any level of the group definition factor. On the NWH dimension, no effects were significant (M(tota1) = 5.2).

DISCUSSION

In accordance with our first hypothesis, members of well-defined (or Wheeler) groups perceived the ingroup as more homogeneous than the outgroup regarding liking


for art in general - an attribute more strongly associated with the well-defined than with the ill-defined group. On the other hand, members of well-defined groups did not show an ingroup homogeneity effect regarding liking for the Wheeler painting, although that attribute was also associated with the ingroup. This pattern of results suggests that these subjects based their social identity largely on the more abstract and value loaded attribute (ART).

When the ingroup was ill-defined, minority and majority members showed the ingroup homogeneity effect regarding alternative attributes: the former regarding liking for the non-Wheeler painting, the latter regarding sociability. This selection of alternative attributes was not arbitrary. The attributes, liking for the Wheeler painting and for art in general, were obviously reserved for the well-defined (or Wheeler) group, while sociability was regarded as an attribute of majority members. Consequently, liking for the non-Wheeler painting was the most appropriate, if not the only available, alternative attribute on which members of the ill-defined minority could base their social identity. Consistent with this assumption, they also tended to ascribe more liking for the non-Wheeler painting to the ingroup than to the outgroup (M(in) = 5.5, M(out) = 4.5, t(108) = 1.59, p < 0.06). On the other hand, the shared belief that majority members are more sociable than minority members offers a plausible explanation why members of the ill-defined majority chose sociability to establish their positive social identity. The mean ratings confirmed that these subjects attributed more sociability to their ingroup than to the outgroup (see Table 1).

The members of the ill-defined majority also accentuated the relative ingroup homogeneity on the W E dimension. At first glance, this might seem inconsistent with our hypothesis that ill-defined group members will show that effect regarding alternative attributes. However, the mean ratings of the members of the ill-defined majority indicate that, compared to the outgroup, the ingroup was seen as being characterized by significantly less liking for the Wheeler painting. Thus, this effect did indeed occur for an alternative attribute, namely the attribute ‘absence of any liking (or even dislike) for the Wheeler painting’ (see Table 1). Moreover, self-ratings on the WHE dimension of the same subjects (M = 3.8) neatly match the low mean ratings for the ingroup, suggesting that they were highly motivated to establish a unified ingroup.

It should also be noted that the ingroup homogeneity effects observed in the experimental conditions cannot simply be due to differences in the perception of well- or ill-defined groups per se. First, in the control conditions the higher homogeneity ratings for the well-defined group (on WHE) were not replicated in the analysis controlling for rating extremity (ANCOVA). However, in the experimental conditions, where subjects were involved as members of one of the two groups, such an analysis did confirm greater perceived homogeneity for the well-defined ingroup. Second, while control subjects perceived the ill-defined group generally as more homogeneous than the well-defined group (on NWH), experimental subjects showed that effect only when the former group was a minority ingroup. Finally, on the SOC dimension homogeneity effects were found only in the experimental, but not in the control conditions. In conclusion, the perceiver’s own group membership is apparently a crucial factor which influences the perception of group homogeneity.

Contrary to our first hypothesis, subjects in the ill-defined, no information condition showed no ingroup homogeneity effect. Rather, outgroup homogeneity effects


occurred on both the ART and SOC dimensions. The absence of any ingroup homogeneity effect suggests that these subjects did not define themselves in terms of their social identity. Interestingly, the outgroup homogeneity effects occurred on dimensions on which intergroup comparisons could hardly contribute positively to those subjects’ social identity. As confirmed by the control conditions, the well-defined group was likely to be seen as u priori superior to the ill-defined group on the ART dimension. Hence, the outcome of the intergroup comparison on that dimension was likely to be unfavourable for members of the latter group. Further, they could not rely on an explicit majority status of the ingroup, which would have allowed for unequivocal ingroup superiority on the SOC dimension.

Thus, it stands to reason that in the ill-defined, no information condition self- enhancement qua individual - the accentuation of a positive personal identity (Turner, 1978,1982) -was a better source for a positive self-concept than intergroup comparisons. This reasoning is supported by subjects’ self-ratings on both the ART and SOC dimensions. Members of ill-defined groups rated themselves more favourably individually on the ART dimension than members of well-defined groups, while on the SOC dimension subjects both in the minority and no information conditions gave more favourable self-ratings than subjects in the majority condition. In short, group membership that provides a positive social identity contributes positively to a person’s self-concept and makes that person less dependent on positive self-evaluations at an individual level.

There is additional evidence in our data that the ingroup (outgroup) homogeneity effect becomes more (less) likely on a dimension when the intergroup comparison on that particular dimension contributes positively to the self-concept. The ingroup outgroup differences of the mean ratings on each of the two value loaded dimensions (ART and SOC) were negatively correlated with the respective ingroupoutgroup differences of the Pd measure ((ART) r = -0.41, (SOC) r = - 0 . 3 7 , ~ ~ < 0.0005, one- tailed). The more the ingroup was favoured relative to the outgroup, the more ingroup relative to outgroup homogeneity was perceived. Moreover, those correlations were substantial when the ingroup was ill-defined ((ART) r = -0.58, (SOC) r = -0.53, ps < 0.0005, one-tailed), but negligible when it was well-defined ((ART) r = -0.11, (SOC) r = -0.15, ns). The close relation between the ingroup homogeneity effect and ingroup favouritism, specifically for people who still have to establish their social identity (i.e. members of ill-defined groups), supports our thesis that both phenomena characterize the search for a positive social identity.

Though the measure of dimensional homogeneity (Pd) provides little evidence that minority members are more likely than majority members to show the ingroup homogeneity effect, this relationship is clearly supported at the level of general homogeneity: Minority members ascribed more predictability and less variability to the ingroup than to the outgroup, while majority members showed the opposite effects. According to our theorizing, the minority or majority position of the ingroup is the decisive factor that produces these effects. One could argue that larger groups per se are generally perceived as less homogeneous than smaller ones. But there are three problems with this alternative explanation. First, control subjects ascribed more predictability to the well-defined group than to the ill-defined group irrespective of relative group size. Second, they perceived the majority as more variable than the minority only when the majority was ill-defined, but not when it was well-defined. Finally, the percentage estimates for the relative size of the well-defined group in


the no information conditions did not correlate with the relative degree of predictability and variability ascribed to it, neither in the experimental nor in the control conditions. The coefficients ranged from -0.22 to 0.21 (all ns). Consequently, differences in group size per se cannot account for the effects observed at the level of general homogeneity.

We also found that, provided the ingroup was well-defined, minority members attached greater importance to their group membership than majority members. Notice that in the Simon and Brown (1987) study the ingroup was also well-defined. Thus, if we take predictability and variability as indices of general homogeneity (heterogeneity) and the importance rating as an index of identification with the ingroup, those results replicated the Simon and Brown (1987) findings. In line with our emphasis on social identification processes, the importance ratings correlated positively with ingroupoutgroup differences of ascribed predictability ( r = 0.41, p < 0.001, one-tailed) and negatively with the analogous differences of ascribed variability ( r = -0.40, p < 0.001, one-tailed). The stronger the identification with the well-defined ingroup, the more general homogeneity was ascribed to the ingroup relative to the outgroup.

Yet when the ingroup was ill-defined the importance ratings did not indicate that minority members identified more strongly with their ingroup than majority members. The importance measure administered to members of ill-defined groups may have been inappropriate for our purpose. Remember that it referred to the importance attached to non-membership in the well-defined outgroup, rather than directly to the importance attached to membership in the ill-defined ingroup. Thus, the indirect reference to the ingroup may have rendered this measure a less valid index of identification than the direct measure administered to well-defined groups. The much higher mean square error term for the former measure supports this possibility (1 1.12 versus 3.04). Not surprisingly, the importance ratings given by members of ill-defined groups did not correlate with any of the two indices of perceived general homogeneity (r(PRE) = 0.01, r(VAR) = 0.12, ns).

The present study demonstrates that both ingroup and outgroup homogeneity effects occur and that a thorough understanding of intergroup behaviour and perception requires the specification of the underlying processes. Although at a general level minority members seem to be more likely to perceive the ingroup as more homogeneous than the outgroup, there is clear evidence that both minority and majority members show the ingroup homogeneity effect regarding specific attributes. The close relationship between the ingroup homogeneity effect and other indices of social identification with the group (ingroup favouritism and importance attached to group membership) supports our thesis that the perception of relative ingroup homogeneity is motivated by the search for a positive social identity. Yet we also replicated the outgroup homogeneity effect. And our findings suggest that this effect occurs when people define themselves largely in terms of their personal identity and attach little importance to their group membership. Thus, social identity theory, drawing on the distinction between personal and social identity (Turner, 1982; Turner et al., 1987), offers a promising framework to understand both ingroup and outgroup homogeneity effects.

A final point raised tangentially by these findings concerns self-esteem. Our data point to the interplay of the contributions of personal and social identity. The demands of our experimental situation made social identity salient. And when it


proved sufficiently positive, subjects were less dependent on self-enhancement qua individual. But when the search for positive social identity failed, individual self- evaluations were higher, suggesting that subjects accentuated their positive personal identities. Yet this line of reasoning does not predicate that the contributions to self- esteem of personal and social identity are negatively related. Rather, people may try to tap both sources (see also Deschamps, 1982b), though one source might contribute more than the other in a given social situation. In fact, there were positive, though very weak, correlations between individual self-evaluations and ingroup favouritism (i.e. ingroup minus outgroup evaluation) (r(ART) = 0.19, r(S0C) = 0.17, p < 0.05, one-tailed) and substantial positive correlations between individual self-evaluations and absolute ingroup evaluations @(ART) = 0.47, r(UM) = 0.39, p < 0.001, one- tailed). Moreover, the latter correlations rule out mere contrast effects (i.e. the higher the rating for the reference group, the lower the individual self-rating) as an alternative explanation for the lower individual self-evaluations given ingroup superiority.

REFERENCES

Allen, V. L. (1985). ‘Infra-group, intra-group and inter-group: Construing levels of organisa- tion in social influence’. In: Moscovici, S., Mugny, G. and van Avermaet E. (Eds) Perspectives on Minority Influence, Cambridge University Press, Cambridge pp. 21 7-238.

Brewer, M. B. (1979). ‘In-group bias in the minimal intergroup situation: A cognitive-motiva- tional analysis’, Psychological Bulletin, 86: 307-324.

Brown, R. J. and Smith, A. (1989). ‘Perceptions of and by minority groups: the case of women in academia’, European Journal of Social Psychology, 19 61-75.

Brown, R. J., Tajfel, H. and Turner, J. C. (1980). ‘Minimal group situations and intergroup discrimination: Comments on the paper by Aschenbrenner and Schaefer’, European Journal of Social Psychology, 10: 399-414.

Campbell, D. T. (1958). ‘Common fate, similarity, and other indices of the status of aggregates of persons as social entities’, Behavioral Science, 3: 14-25.

Codol, J.-P. (1 984). ‘Social differentiation and non-differentiation’. In: Tajfel, H. (Ed.) The Social Dimension, Vol. 1, Cambridge University Press, Cambridge, pp. 3 14-337.

Codol, J.-P. (1987). ‘Comparability and incomparability between oneself and others: Means of differentiation and comparison reference points’, European Bulletin of Cognitive

Deschamps, J.-C. (1982a). ‘Social identity and relations of power between groups’. In: Tajfel, H. (Ed.) Social Identity and Intergroup Relations, Cambridge University Press, Cambridge,

Deschamps, J. C. (1982b). ‘Differenciations entre soi et autrui et entre groupes’. In: Codol, J.-P and Leyens, J.-P, (Eds) Cognitive Analysis of Social Behavior, Martinus Nijhoff Publishers, The Hague, pp. 247-266.

Hogg, M. A. and Abrams, E. (1988). Social Identifications (A social psychology of intergroup relations and group processes), Routledge, London.

Jones, E. E., Wood, G. C. and Quattrone, G. A. (1981). ‘Perceived variability of personal characteristics in ingroups and outgroups: The role of knowledge and evaluation’, Person- ality and Social Psychology Bulletin, 7: 523-528.

Judd, C. M. and Park, B. (1988). ‘Out-group homogeneity: Judgments of variability at the individual and group levels’, Journal of Personality and Social Psychology, 54: 778-788.

Kelly, C. (1989). ‘Political identity and perceived intragroup homogeneity’, British Journal of Social Psychology, 28: 239-250.

Lemaine, G. (1974). ‘Social differentiation and social originality’, European Journal of Social

Linville, P. W., Fischer, G. W. and Salovey, P. (1989). ‘Perceived distributions of thecharacter- istics of in-group and out-group members: Empirical evidence and a computer simulation’, Journal of Personality and Social Psychology, 57: 165-1 88.

Psychology, 7: 87-105.

pp. 85-98.

Psychology, 4: 17-52.


Linville, P. W. and Jones, E. E. (1980). ‘Polarized appraisals of out-group members’, Journal of Personality and Social Psychology, 38: 689-703.

Linville, P. W., Salovey, P. and Fischer, G. W. (1986). ‘Stereotyping and perceived distributions of social characteristics: An application to ingroupoutgroup perception’. In: Dovidio, J. F. and Gaertner, S. L. (Eds) Prejudice, Discrimination, and Racism, Academic Press, Inc., Orlando, pp. 165-208.

McGuire, W. J. and McGuire, C. V. (1988). ‘Content and process in the experience of self‘. In: Berkowitz, L. (Ed.) Advances in Experimental Social Psychology, Vol. 21, Academic Press, Inc., New York, pp. 97-144.

Messick, D. M. and Mackie, D. M. (1989). ‘Intergroup relations’, Annual Review of Psychology, 40: 45-81.

Moscovici, S., Mugny, G. and van Avermaet, E. (Eds) (1985). Perspectives on Minority Influence, Cambridge University Press, Cambridge, England.

Mullen, B. and Hu, L. (1989). ‘Perceptions of ingroup and outgroup variability: A meta- analytic integration’, Basic and Applied Social Psychology, 10: 233-252.

Mummendey, A. and Schreiber, H.-J. (1983). ‘Better or different? Positive social identity by discrimination against or by differentiation from outgroups’, European Journal of Social

Mummendey, A. and Schreiber, H.-J. (1984). ‘“Different” just means “better”: Some obvious and some hidden pathways to in-group favouritism’, British Journal of Social Psychology,

Mummendey, A. and Simon, B. (1989). ‘Better or different? 111: The impact of importance of comparison dimension and relative ingroup size upon intergroup discrimination’, British Journal of Social Psychology, 28: 1-16.

Quattrone, G. A. (1986). ‘On the perception of a group’s variability’. In: Worchel, S. and Austin, W. G. (Eds) Psychology of Intergroup Relations, Nelson-Hall Publishers, Chicago.

Quattrone, G. A. and Jones, E. E. (1980). ‘The perception of variability within in-groups and out-groups: Implications for the law of small numbers’, Journal of Personality and Social Psychology, 38: 141-152.

Simon, B. and Brown, R. (1987). ‘Perceived intragroup homogeneity in minority-majority contexts’, Journal of Personality and Social Psychology, 53: 703-71 1.

Snyder, C. R. and Fromkin, H. L. (1980). Uniqueness the Human Pursuit of Difference, Plenum Press, New York.

Tajfel, H. (1978). The Social Psychology of Minorities, Minority Rights Group, London, (No. 7).

Tajfel, H., Billig, M. G., Bundy, R. P. and Flament, C. (1971). ‘Social categorization and intergroup behaviour’, European Journal of Social Psychology, 1: 149-1 78.

Tajfel, H. and Turner, C. J. (1986). ‘The social identity theory of intergroup behaviour’. In: Worchel, S. and Austin, W. G. (Eds) Psychology of Intergroup Relations, Nelson-Hall Publishers, Chicago.

Turner, J. C. (1978). ‘Social categorization and social discrimination’. In: Tajfel, H. (Ed.) Diferentiation Between Social Groups, Academic Press, London, pp. 101-168.

Turner, J. C. (1981). ‘The experimental social psychology of intergroup behaviour’. In: Turner, J. C. and Giles, H. (Eds) Intergroup Behaviour, Blackwell, Oxford, England, pp. 66-101.

Turner, J. C. (1982). ‘Towards a cognitive redefinition of the social group’. In: Tajfel, H. (Ed.) Social Identity and Intergroup Relations, Cambridge University Press, Cambridge,

Turner, J. C., Hogg, M. A., Oakes, P. J., Reicher, S. D. and Wetherell, M. S. (1987) Rediscover- ing the Social Group. A Self-categorization Theory, Basil Blackwell, Oxford.

van Knippenberg, A. and Ellemers, N. (1989). ‘Social identity and intergroup differentiation processes’. In: Stroebe, W. and Hewstone, M. (Eds) European Review of Social Psychology, Wiley, Chichester.

Wilder, D. A. (1984). ‘Predictions of belief homogeneity and similarity following social categorization’, British Journal of Social Psychology, 23: 323-333.

Wilder, D. A. (1986). ‘Social comparison: Implications for creation and reduction of intergroup bias’. In: Berkowitz, L. (Ed.) Advances in Experimental Social Psychology, Vol. 19, Academic Press, New York. pp. 291-355.

Psychology 13: 389-397.

23: 363-368.

pp. 1540.

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Social identity and perceived group homogeneity: Evidence for the ingroup homogeneity effect

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