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Copyright 1997 by the American Psychological Association, Inc. 0012-1649/97/53.00

Developmental Psychology 1997, Vol. 33, No. 5, 824-833

Measurement of Two Social Competence Aspects in Middle Childhood Ann-Margret Rydell, Berit Hagekull, and Gunilla Bohlin

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Uppsala University The psychometric properties of a rating measure for parents and teachers for social competence, conceptualized as social skills and behaviors, were studied. The rating measure was constructed from factor analyses on 4 samples of school-age children. Factor analyses identified 2 moderately correlated competence aspects, valid for both sexes and for children from varying socioeconomic backgrounds. The first factor, Prosocial Orientation, captured a style promoting positive social interactions; the second factor, Social Initiative, described initiative as opposed to withdrawal in social situations. Scales based on the 2 factors showed reliability in internal consistency and stability across 1 year, validity in interrater agreement concurrently and across 1 year, correspondence with observed peer behavior, and the capacity to discriminate between children of different peer status.

Children's adaptive functioning in their social environment, often referred to as the phenomenon of social competence, is of central importance to socioemotional development in middle childhood (e.g., Anderson & Messick, 1974; Cavell, 1990; Elliott & Gresham, 1993). Social competence has been studied in terms of peer status (Blechman, Tinsley, Carella, & McEnroe, 1985; Cauce, 1987; Howes, 1987) and as prosocial behaviors and interpersonal skills (Denham, Zahn-Waxler, Cummings, & Iannotti, 1991; Ladd & Goiter, 1988; Rosen, Furman, & Hartup, 1989; Tremblay, Vitaro, Gagnon, Piche, & Royer, 1992; Turner & Harris, 1984). As has been commented on by researchers in the field (e.g., Cavell, 1990; Dodge, Pettit, McClaskey, & Brown, 1986), the different social competence measures point to a lack of consensus concerning the definition of social competence.

A parallel can be drawn with work on children's maladaptive behavior. Reliable and widely used rating scales for behavior problems have been developed (e.g., Achenbach, Howell, Quay, & Conners, 1991; Conners, 1990; Quay & Peterson, 1987; Rutter, Tizard, & Whitmore, 1970) through which substantial knowledge about the phenomena has been generated. For example, a large number of studies have established the content of the problem behavior domain and identified dimensions of problem behavior (e.g., Achenbach et al., 1991; Garber, Quiggle, Panak, & Dodge, 1991; Venables et al., 1983). In comparison, competence studies addressing similar issues have been relatively few (Goodman, 1994; Matson, Rotatori, & Helsel, 1983; Tremblay et al., 1992). The content of the social competence construct, what skills and behaviors to measure, is obviously the first question in a measurement endeavor. Prosocial behaviors (helpfulness, generosity, and empathy) and resourcefulness in initiating activities in social interaction contexts gained the highest ratings by a number of psychologists in their Q-sort descriptions of a hypothetical, socially competent child (Waters, Noyes, Vaughn, & Ricks, 1985). In factor analyses on parent and teacher ratings of social competence, items reflecting prosocial behaviors and cooperation have loaded on social skills /prosocial factors (Goodman, 1994; Matson et al., 1983; Tremblay et al., 1992), whereas a conceptualization of social competence as social initiative and engagement in interactions (e.g., suggests activities, spectator vs. participant) was presented in a peer competence Q-sort scale for preschoolers (Waters, Wippman, & Sroufe, 1979). The salience of the above behaviors for peers also has been demonstrated in the peer status literature (for reviews, see Coie, Dodge, & Kupersmidt, 1990; Newcomb, Bukowski, & Pattee, 1993). Compared with children of average status, popular children exhibit high levels of prosocial behaviors and high levels of social interaction. The rejected status group is low in prosocial behavior compared with average children, but consistent differences regarding social initiative from the average group have not been reported. The neglected status group has been less easy to characterize, not consistently different from average children in either prosocial behavior or social avoidance (Coie et al., 1990; Newcomb et al., 1993).

Cavell (1990) pointed out that social competence definitions could be seen as referring to different levels of the phenomenon and that fruitful research questions concern how these levels relate to each other. A particularly valuable distinction was made between products of social functioning, such as peer status and self-esteem, and the child's actual social behavior (e.g., prosocial behaviors such as helping peers). In the present study, the Cavell distinction was adopted, and our social competence construct encompassed social skills and behaviors to the exclusion of social outcomes. Tn the social competence domain, evalu^'ons of behavior by the child's social partners, such as parents and teachers, play a central role in research, making the availability of easily administered instruments to assess individual differences a prime interest.

Ann-Margret Rydell, Berit Hagekull, and Gunilla Bohlin, Department of Psychology, Uppsala University, Uppsala, Sweden. This research was financially supported by the Bank of Sweden Tercentenary Ttoundation. We are indebted to Kerstin Bosson, Lilianne Eninger, Outi Lento, Lene Lindberg, Lena Manns, and Gunilla Stenberg for assistance with data collection. Correspondence concerning this article should be addressed to AnnMargret Rydell, Department of Psychology, Uppsala University, Box 1225, S-751 42 Uppsala, Sweden. Electronic mail may be sent via Internet to [email protected],se.

824

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MEASURING SOCIAL COMPETENCE

A second important question concerns the dimensionality of the competence domain. Differences in content between the aforementioned measures of competent social behavior raise the question whether items on prosodality/cooperation and on social participation/initiative reflect the same underlying dimension. In addition, the homogeneity of the rejected peer status group with regard to prosocial behaviors but not concerning social initiative (e.g., Newcomb et a l , 1993) suggests that prosocial behavior and social initiative do not always covary. So far, however, studies have shown social competence to be a unidimensional phenomenon. Factor analyses performed on items tapping problematic and prosocial behaviors but with no items on initiative have found prosocial items and items on cooperation and successful conflict handling to load on a single, separate factor (Goodman, 1994; Tremblay et al., 1992). In their social skills rating scale, Matson and colleagues (Matson et al., 1983) included a few social initiative items in an instrument containing items mainly capturing prosocial behaviors and found the initiative items to load on a single social skill factor together with the prosocial items. One possible explanation of the identification of one factor combining social initiative and prosocial behaviors may be that the initiative items (two items in a 20-item scale) provided insufficient coverage of social initiative behavior Thus, studies including a larger number of initiative items together with prosocial items would be needed to investigate the possibility of two competence aspects. Third, child behaviors that are salient to development are assumed to be visible to observers in different social contexts and to be relatively stable across time, at least in a short perspective. Modest informant agreement and stability have been demonstrated for prosocial behavior in middle childhood. Correlations between parent and teacher ratings of prosocial behavior have ranged from r = .14/.21 (for girls and boys, respectively; Tremblay et al., 1992) to r = .34 (Goodman, 1994). Tremblay and coworkers reported correlations of about .40 for teacherrated prosocial behavior of children between 6 and 7 years of age. Because ratings of children's social behavior could be questioned as being influenced by informants' biases and behavioral expectations, researchers should investigate the correspondence of rating measures to behavioral observations. Finally, peer status is a major social outcome in middle childhood, and salient social behaviors should be expected to discriminate between children of different peer status. In our research group, data on social competence conceptualized as skills and behaviors have been collected in a number of studies concerning child functioning in middle childhood. We used a questionnaire to be filled out by parents and teachers. In this article, data from four samples will be used to address measurement issues in the competence domain. By including several items on social initiative as well as items covering prosocial behaviors, cooperation, and conflict handling, we propose to investigate whether two competence aspects, one pertaining to prosocial behaviors and another pertaining to initiative and outgoingness in interactions, can be identified. The properties of the instrument regarding internal consistency, test-retest reliability in terms of short-term stability, validity in terms of interrater agreement at the same point in time and across time,

and relations to observed positive peer behaviors and to peer status will be examined.

Method Participants The pilot sample consisted of all 121 children (56.2% boys) in six elementary school classrooms in two urban and one rural working/lowermiddle-class school district in and near a small town close to Uppsala, Sweden. The majority, 85 children, were in Grade 2 (8 to 9 years old), 29 children were in Grade 3 (9 to 10 years old), and 7 children were in Grade 1 (7 to 8 years old). Because teachers were not expected to be reliable sources of information regarding the children's sociodemographic characteristics, further background information was not obtained. The Uppsala sample consisted of children from a longitudinal project in Uppsala, Sweden (for a description of the sample, see Hagekull & Bohlin, 1990). In 1985, we selected 123 families from the communal birth register (62% of the contacted mothers agreed to participate); mothers and infants were enrolled when one infant was 6 weeks of age. When the infants were 10 months old, fathers were invited to participate in the project, and 104 (85%) agreed to do so. The present data set encompassed the 96 children (50% boys) who remained in the study when the children were 7 to 8 years old (M = 7 years 11 months, SD = 4 months). Reasons for attrition over the 8 years have been illness or death in the family, moving out of the country, travels abroad, objections to interview-questionnaire content, and time shortage. The attrition group was not significantly different from the group who remained with respect to sex composition, parity, number of siblings, or paternal education at the start of the project (all x2 values < 2.40, ns). When the children were 8 years old, maternal age ranged from 29 to 52 years (M = 38 years 4 months, SD = 4 years 6 months). A minority, 4%, were only children, and most of the children (89%) lived in two-parent households. In 76% of the Uppsala families, one or both parents had a university degree or a college education, 10% had completed secondary school (12 years of school), 11 % had 2 years of vocational training on the high school level, and in 3% of the families, none of the parents had any training or schooling after compulsory school (9 years of school). During the spring semester in second grade, data from 95 children that remained in the study were secured. The provincial sample consisted of 118 children (93% of the available children; 51.7% boys), from six third-grade classrooms, living in and near a provincial Swedish town. Attrition was due to parents' wish that the child should not participate. The children were between 9 and 10 years old (M = 9 years 6 months SD = 4 months). The majority (87%) lived in two-parent families, and 6% were only children. In 28% of the families, one or both parents had a university degree or a college education, 14% had completed secondary school, 46% had vocational training, and in 12%, none of the parents had gone beyond compulsory school. The county sample consisted of 423 children (M age = 8 years 5 months SD = 3 years 3 months), constituting 71% of a randomly selected sample of children born in 1986 and living in a county in the middle of Sweden, According to Statistics Sweden (personal communication, October 1991), this county is representative of the country at large as regards sociodemographic composition. About half of the children (48%) were boys. The majority (89%) lived in two-parent families, and a minority, 3%, were only children. In 34% of these families, one or both parents had a university degree or a college education, 36% had completed secondary school, 25% had vocational training, and in 5%, none of the parents had gone beyond the compulsory school. Thus, this study comprises four samples of Swedish 7- to 10-yearold children, about half of them boys. All children were enrolled in

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mainstream classrooms. There were differences in educational background between the Uppsala, provincial, and county samples. Although the Uppsala sample had a large proportion of highly educated parents, as one might expect in a university area, the majority of the children in the provincial sample had parents with fairly short formal schooling. The county sample consisted mainly of families with a medium level of parental education (secondary school or vocational training).


Measures Social Competence Inventory. In 29 questionnaire items, social skills and behaviors indicative of empathy, including ability to decode emotions and thoughts of others, altruism, generosity, helpfulness, social participation, initiative taking, cooperation, and conflict handling were described. The item pool was composed to cover a broad range of social competence characteristics. Items were constructed on the basis of several prior instruments describing children's social competence (Goodman, 1994; Matson et al., 1983; Tremblay et aL, 1992; Waters et al., 1979, 1985). Five-step response scales were used with scale endpoints stated for each item (1 = doesn't apply at all, 5 = applies very well to the child). The middle steps were defined in;the general instruction to the respondents. Higher scores indicated higher competence. Observations of social behavior. Although not originally designed for validation of the Social Competence Inventory, observations of social behavior were available in the Uppsala sample and were used to assess correspondence between teacher ratings and observed behavior. A recording scheme for social behavior, the Social Behavior Checklist for preschool children (White & Watts, 1973) was modified to suit 7- to 8-year-old children. Five peer interaction items judged to be less relevant for school children than for preschoolers (e.g., imitates peer, competes for adult attention) were discarded, and 5 items measuring relevant behaviors not sufficiently covered by the original checklist (e.g., engaged in conflict, response to peer's initiative) were added. After testing the scheme on 10 children in the pilot sample, 15 items representing positive and negative behaviors with peers, and U items representing positive and negative behaviors with adults were included. Each child was observed during ordinary school activities for 22 five-min periods distributed across recess (8 periods), classroom activities (12 periods), and the lunch meal (2 periods). The behavioral items were recorded in terms of frequency (15 items) or in terms of duration (0 = not present, 5 = present for more than 4 mins; 11 items) during each observation period. Each item was averaged across observation periods and standardized. Scales were constructed on the basis of factor analyses, homogeneity estimates, and visual content analysis. In the present context, the scale measuring positive behaviors with peers, the Positive Peer Behavior scale (Cronbach's a = .61) will be reported (see Table 1 for display of

Table I Items in the Positive Peer Behavior Scale Item Postive affect towards peer Prosocial behaviors Initiative with peer Positive interaction with peer Responds to peer Solitary activity (reversed) Evidence of leadership

Characteristic behavior Smiles, touches friendly, verbalizes friendly Helps, comforts, shares Addresses, shows something, suggests activity Engaged in play, conversation, or work Answers, listens, joins Peer accepts suggestion, follows, listens, asks for direction

items). Estimates of interobserver agreement were obtained by pairing observers prior to, in the middle of, and at the end of data collection. Five children in the training phase and 13 children in the data collection phases were observed during a total of 163 five-min periods. Interrater agreement was high ( r = .98) for the Positive Peer Behavior scale, and there were no significant mean differences between independent observers. Peer status. Peer status was measured with a widely used sociometric method (Asher & Dodge, 1986). Each child was asked to nominate 3 classmates he or she wanted to play with and to indicate how much (1 = not at all to 5 = very much) he or she wanted to play with each child in the class. The not at all ranking is seen as equivalent to a negative nomination according to Asher and Dodge. A categorization in peer status groups was performed, following the procedure described by Asher and Dodge: For each child, a positive nomination (PN) score and a negative nomination (NN) score was computed, and the scores were standardized within each classroom. A social preference (SP) score was obtained by subtracting the standardized NN score from the standardized PN score, and a social impact (SI) score was obtained as the sum of the standardized PN and NN scores. The SP score and the SI score were then standardized within each classroom. Children with popular (SP score >1.0; PN score > 0 ; NN score < 0 ) , rejected (SP score < - 1 . 0 ; PN score < 0 ; NN score > 0 ) , neglected (SI < - 1 . 0 ; PN score < 0 ; NN score < 0 ) and average (SP and SI scores between —.5 and .5) peer status were identified.

Procedure Pilot sample. During the fall of 1992, six class teachers completed the Social Competence Inventory for each of their pupils and responded to questions about the child's gender and grade level. Uppsala sample. In the spring of 1993 when the children were 8 years old, 89 mothers and 7 fathers completed a mailed questionnaire containing the Social Competence Inventory and questions about maternal age, family composition, number of siblings, and parental education. During the same period, observations of social behavior in school (n 90) took place, and the class teachers (n = 92) responded to the Social Competence Inventory. Reasons for attrition in the school setting were that parents did not wish the child's school to be contacted or that the family lived abroad and information from the teacher could not be obtained. Five graduate psychology students were trained as observers of social behaviors. Most (93.3%) children were observed during 22 fiveminute periods, and no child was observed for less than 19 periods. Observations were distributed over 2 school days for each child, but for practical reasons, 15 children were observed during 1 day. In the spring of 1994 (children at age 9 years), mothers (« = 92), fathers (n = 15), and class teachers (n = 87) completed the Social Competence Inventory. Reasons for attrition were that the family wanted to leave the study, that one of the parents could not be contacted or did not return the questionnaire in spite of several reminders, and that the teacher was not contacted, according to the parent's wish, or did not return the questionnaire after reminders. Twelve of the 87 children had a new teacher in Grade 2. Provincial sample. During December 1993 to March 1994, parents were contacted and, after a written consent, 93 parents (83 mothers and 10 fathers) and the class teacher for 118 children completed the Social Competence Inventory. For parents, background questions (family composition, number of siblings, parental education) were included in the mailed questionnaire. Reasons for attrition were that parents agreed to the child's participation but did not return the parent questionnaire despite several reminders. During the same period, each child (n = 118) was interviewed by a senior psychology student. The interviews took


MEASURING SOCIAL COMPETENCE place in a separate room in the child's school. The child was assured of confidentiality at the start and close of the interview, which took approximately 1 hour. As part of the procedure, the child was asked to nominate the three classmates he or she most wanted to play with and then to indicate how much ( I = not at all; 5 = very very much) he or she wanted to play with each classmate presented on the class roster. County sample. During the fall of 1994, parents (402 mothers and 2( fathers) responded to a mailed questionnaire containing the Social Competence Inventory and questions about their child's gender, age, number of siblings, and about the mother's age, family composition, and parental education. In the following, ratings made by either mother or father (one rating per child) will be referred to as parental ratings. In the Uppsala sample, mothers and fathers made independent ratings when the children were 9 years old. Hence, maternal and paternal ratings will be analyzed separately for that age. The measures and data sources used in the different samples are displayed in Table 2.

Results Factor Analyses of the Social Competence Inventory and Construction of Scales Principal factor analyses with both orthogonal and oblique rotations were performed on the pilot-sample teacher ratings of the 29 social competence items. Three factors obtained eigenvalues greater than 1, but Cattell's scree test and the principle of simple structure (Gorsuch, 1974) favored a two-factor solution. Orthogonal and oblique rotations yielded the same factors, but the size of the interfactor correlation (r = .33) indicating 11% overlapping variance among the factors favored an oblique rotation. The oblique two-factor solution explained 65.2% of the variance. In the interpretation of factors, loadings greater than or equal to .40 were considered salient. In the first factor, high loading items reflected a social style that seemed to promote smooth social interactions. Descriptions of generosity, empathy, understanding of others, conflict handling, and helpfulness were

Table 2 Measures and Data Sources in Four Samples Sample and child age Pilot 8 to 9 years Uppsala 8 years

Teachers

121

Social Competence Inventory

Parents" Teachers Observers Mothers Fathers Teachers

96 92 90 92 75 87

Peer Nominations

Parents6 Teachers Classmates

93 118 118


Parents'"

423

Positive Peer Behavior Social Competence Inventory

Provincial 9 years


a

N


9 years

County 8 years

Data source

Measure

89 mothers, 7 fathers. fathers.

b

83 mothers, 10 fathers,

;

402 mothers, 21

827

included in this factor. This mainly unipolar factor was seen as a prosocial orientation aspect of social competence. Items in the second factor described both active initiatives in social interaction situations and withdrawal behaviors in such situations. This factor was interpreted as mirroring a bipolar social competence aspect of initiative taking versus social withdrawal. The factors were named Prosocial Orientation and Social Initiative, respectively. To cross-validate the selected factor solution from the pilot sample, teacher ratings from the Uppsala sample (8-year data) and teacher ratings from the provincial sample were pooled (n = 210). as were parental ratings from the Uppsala (8-year data) and from the provincial samples (n = 189). Factor analyses with oblique rotation were performed separately on the two pooled data sets, yielding modest to moderate interfactor correlations (r = .47 and r ~ .27) in the teacher and the parent data set, respectively. After matching factors on the basis of their highest loadings, Cattell's salient variable similarity index (s index; Cattell, Balcar, Horn, & Nesselroade, 1969) was used to compare the obtained factors with the factors from the pilot sample. The pilot sample factors proved stable in the teacher ratings for the Prosocial Orientation factor (s = .97, p < .001), and for the Social Initiative factor (s = .83, p < .001). The factors were also replicated in the parental ratings. Both the Prosocial Orientation factor (5 = .92, p < .001) and the Social Initiative factor (s = .78, p < .001) proved stable. Thus, in factor analytic procedures, the same two correlated aspects of social competence were demonstrated in three sets of data, consisting of both teacher and parent ratings of children from varying social backgrounds. Prior to the construction of scales, the pilot teacher ratings, the Uppsala teacher ratings (8 year), and the provincial teacher ratings (n = 331) were pooled, and two obliquely rotated factors, with the same content as in previous analyses, were obtained from the enlarged teacher data set. On the basis of the factor analyses on the pooled parent and the larger of the pooled teacher data sets, two scales were constructed. The criterion for including an item was a factor loading greater than or equal to .40 in both the parent and the teacher solution. Three items (helps peers with school work, makes contact easily with unfamiliar adults, and easily gets into conflicts with adults) failed to reach this criterion. Double loading items were considered to belong to the factor in which the item loaded higher. A fourth item (tries to boss peers around) was discarded on the grounds that the parent and the teacher solutions disagreed on which factor it should belong to. With this procedure, 25 items were identified as salient in both data sets and were retained as scale items (Table 3). Scales were created from unit-weighted scores by averaging items. The Prosocial Orientation scale consisted of 17 items, and the Social Initiative scale had 8 items. The alpha coefficient for the Prosocial Orientation scale, on the basis of parental ratings, amounted to .88, and the teacher-rated scale had an alpha of .94. The Social Initiative scale had alpha values of .75 (parents) and .91 (teachers). The correlation between parent scales was r = .36, and the corresponding figure for teacher scales was r = .53. A final test of factor pattern and scale homogeneity was performed on the parent ratings from the nation representative county sample (n = 423). Exploratory factor analyses with

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RYDELL, HAGEKULL, AND BOHLIN Table 3

Factor Pattern of Social Competence in Parent (n = 189) and Teacher (n — 331) Samples


Parents

Teachers

Item

FI

HI

FI

FII

Has capacity for generosity to peers Has capacity to be helping/altruistic Has capacity to sympathize with peers Criticizes peers Helpful with adults Helps peer tidy up/search for lost items Shares his/her belongings Good at preventing conflicts Comforts peer who is upset/sick Includes shy children in play Has ability to decode peers' feelings Tries to intervene in peer conflicts Gives compliments to peers Finds solution when in conflict Has the capacity to play/work well with peers Can give and take in interactions Shares peers' joy Leads play activities Socially withdrawn with peers Makes contact easily with unfamiliar children Hesitant with peers Spectator while others play Shy/hesitant with unfamiliar adults Suggests activities to peers Dominated by peers

.76 .70 .66 -.62 .57 .56 .54 .51 .50 .48 .48 .46 .45 .44 .43 ,42

-.03 -.05 -.03 -.15 -.01 .00 -.07 -.09 .16 .31 .06 .16 .03 .17 .37 .19 .16 .62 -.60 .56 -.52 -.51 -.50 .48 -.45

.85 .81 .82 -.62 .61 .66 .67 .68 .63 .54 .75 .57 .61 .67 .76 .70 .52 -.05 .02 .19 -.18 -.19 -.08 .25 .05

-.08 -.09 -.03 .51 .07 .07 -.04 .13 .22 .17 .06 .31 .25 .17 .02 .13 .08 .88 -.81 .65 -.71 -.65 -.63 .66 -.72

Note.

.41 -.29 .02 .25 -.16 -.20 -.08 -.12 .11

Boldface is used to identify items with factor loadings ^ .40. FI = Factor 1; FII = Factor 2.

scree tests, extraction of two and three factors, consideration of the factor correlation, and application of the criterion of psychological meaningfulness to the interpretation of factor solutions again favored the two-factor oblique solution with a Prosocial Orientation and a Social Initiative factor. The s indices for the Prosocial Orientation factor and the Social Initiative factor were .86 and .95, respectively, in comparison with the previously extracted teacher factors, and .91 and .95 compared with the factors from the pooled parent ratings, all significant (on the p < .001 level). In the county sample, the alpha coefficient for the Prosocial Orientation scale was .88 and for the Social Initiative scale it was .76, and the correlation between the scales was r = .42. To investigate the stability of the factor solution with regard to gender, separate analyses for boys and girls in the county sample were performed. The obliquely rotated two-factor solution (the interfactor correlation was r — .36 for girls and r = .42 for boys) was stable across gender (s — .90 for the Prosocial Orientation factor and s• = .91 for the Social Initiative factor, both p < .001). Thus, the factor pattern was once again confirmed, showing that social competence, with the present operationalization, could be conceptualized as two moderately correlated competence aspects and that the two-factor solution was applicable to both boys and girls.

Psychometric Properties of the Social Competence Scales Test-retest reliability. On the assumption of reasonable stability of social competence behaviors across a time period of 1

year, test-retest reliability could be tested in terms of stability in mothers' and teachers' social competence ratings with longitudinal data from the Uppsala sample. The correlation coefficients between ratings for children ages 8 and 9 years for the two competence aspects were substantial (see Table 4 ) . However, as the two scales were related to each other at both ages, some uncertainty regarding the interpretations of the coefficients is introduced. For example, part of the stability in prosocial orientation could be accounted for by the combined effect of prosocial orientation and social initiative ratings at 8 years on the prosocial orientation ratings at 9 years. To investigate the crosstime relations of each competence scale, controlling for its

Table 4 Test-Retest Correlations for Social Competence Ratings by Mothers and Teachers in the Uppsala Sample of Children Between 8 and 9 Years of Age Mothers" (n = 85)

Teachersb (n = 75)

Scale

r

r

Prosocial Orientation Social Initiative

77*** 70.***

a

Children rated by fathers at 8 years excluded, new teacher at 9 years excluded. ***» < .001.

.81*** b

Children rated by a

829



shared variance with the other competence scale at each age, hierarchical regression analyses were performed. As an example, the amount of variance in the 9-year mother Prosocial Orientation scale explained by 8-year prosocial orientation ratings, while controlling for social initiative ratings at ages 8 and 9 years, was estimated. In all four hierarchical regression analyses of 9-year ratings (mother prosocial orientation and social initiative ratings, and teacher prosocial orientation and social initiative ratings), the increment in explained variance in 9-year ratings was substantial when 8-year ratings of the same competence scale were added to 8- and 9-year ratings of the other competence scale. The increment in R2 varied between .27 and .53, and betas ranged from (3 = .58 to /? = .82, (all p values < .001). Thus, the stability coefficients were accounted for mainly by the agreement across 1 year of concerning child behaviors specifically captured by each scale and not by the shared variance of the two competence aspects. Validity. Validity of the social competence scales was explored in four sets of analyses. First, validity was estimated in terms of agreement between different raters at the same point in time. Second, the capacity of the scales to predict ratings across time with a new rater was studied. Third, agreement between ratings and observed behavior was estimated; finally, validity in terms of the capacity of the social competence scales to discriminate between peer status groups was explored. First, the agreement between parents and teachers was studied in correlational analyses in the provincial sample and in the Uppsala sample for children at age 8 years. In the latter sample, agreement between fathers and teachers was tested as well as mother-father concordance in children at age 9 years (see Table 5). The agreement was clearly significant for the Social Initiative scale in all comparisons (r = .37 to r = .47). For the Prosocial Orientation scale, agreement was significant in all comparisons in the Uppsala sample (r ~ .28 to r = .44), but parent-teacher agreement in the provincial sample was not significant (r — .17, p = .11). Second, the validity of the scales was studied in terms of relations between 8-year mother and teacher ratings and in 9year ratings by new raters (fathers and new teachers) in the Uppsala sample. There were significant cross-time relations in

all comparisons except teacher-parent comparisons for prosocial orientation. For prosocial orientation, mother 8-year ratings were related to 9-year father and 9-year teacher ratings, r (68) = .40/r(80) = .46, p < .001, but 8-year teacher ratings were not related to 9-year mother or 9-year father ratings, r(88) = .09/r(71) = .21, ns. For the 11 children who had changed teachers between ratings, and for whom teacher ratings were available in both Grade 1 and Grade 2, the teacher-teacher correlation was r = .49, ns. With regard to social initiative, all the cross-time correlations were significant, (r = .43/r = .45, p < .001 for mother-father and mother-teacher comparisons; r = .46, p < .001 and r — .38, p < .01 for teacher-mother and teacher-father comparisons; and r = .73, p < .01 for the teacher-teacher comparison). The small n value in the teacherteacher comparisons raised suspicions that the relatively high correlations (only about 10 units lower than cross-time correlations for same teachers; see Table 4) might be accounted for by a few outliers, but scatter plots proved this not to be the case. As a third test of validity, information in the Uppsala sample from the child's school setting at age 8 years was used to investigate the validity of the social competence scales in terms of concurrent relations to observations of peer behavior. Two sets of analyses were performed. First, relations between teacher ratings and the Positive Peer Behavior scale, which was based on observations of behavior, were assessed in correlational analyses. Both competence scales should be related to the observational measure, the Positive Peer Behavior scale containing items reflective of both prosocial and social initiative behaviors (see Table 1). This expectation was borne out as relations between both scales and the Positive Peer Behavior scale were significant (r- .29, p < .01 a n d r = .29, p < .01; for Prosocial Orientation and Social Initiative, respectively). Second, the capacity of the two scales to discriminate between observed prosocial and initiative behavior was investigated. In an effort to match observational items to the content of the two rating scales, two measures were constructed from items in the Positive Peer Behavior scale. The two observational items "positive affect" and "prosocial behaviors'1 were combined to yield a prosocial behavior measure corresponding to prosocial orientation (interobserver agree-

Table 5 Interrater Correlations of Social Competence Ratings Rater relation and scale

Sample

Parent/teacher Prosocial Orientation Social Initiative Parent/teacher Prosocial Orientation Social Initiative Mother/father Prosocial Orientation Social Initiative Father/teacher Prosocial Orientation Social Initiative

Provincial

**p < .01. ***p < .001.

Age (years) 93 .17 45***

Uppsala

92

.29** Uppsala

74 44*** 42***

Uppsala

71 .35** 37***


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RYDELL, HAGEKULL, AND BOHLIN

ment: r [18] = .97; no significant mean differences between independent observers). The three items "positive contacts," "solitary play" reversed, and "evidence of leadership" were combined to a measure conceptually corresponding to social initiative (interobserver agreement, r [18] = .87; no significant mean differences between independent observers). Multiple regression analyses were used to control for the shared influence of the two rating scales. Prosocial behavior was predicted by prosocial orientation ratings only (/? — .23, p < .05; /? for social initiative ratings was - . 0 3 , ns), and initiative behavior was significantly predicted by social competence ratings of social initiative only (/? = .30, p < .01; /? for prosocial orientation ratings was .06, ns). As a final validity estimate, information in the provincial sample from the child's school setting was used to test the expectation that teacher ratings of social competence should discriminate between peer.status groups on the basis of nominations from classmates. The popular group was contrasted with the rejected group, as one would expect the scales to discriminate between children with extreme peer status positions. The policy to contrast the average group with each nonaverage group (e.g., Newcomb et al., 1993) was also adopted. Thus, the peer status groups were compared on prosocial orientation and social initiative in analyses of variance (ANOVAs) with planned orthogonal comparisons, with F ratios for hypothesis testing of differences between the popular and the rejected groups, the average and the rejected groups, and the average and the neglected and the popular groups, respectively. As seen in Table 6, both rating scales differentiated popular from rejected children ( F = 22.16, p < .0001 and F = 13.66, p < .001 for prosocial orientation and social initiative, respectively). Prosocial orientation separated rejected from average children (F = 8.68, p < .01), and the difference between the average and the popular group was close to significance (F - 3.68, p = .06). Social initiative separated popular from average children (F = 4.62, p < .05), whereas rejected children were marginally different from average in this respect ( F = 2.81, p < .10). Discussion In this study, we have investigated the psychometric properties of a rating instrument for social competence. For replica-

tion purposes, four different samples were used. On the basis of factor analyses, two social competence aspects were identified, valid for both boys and girls, and a Prosocial Orientation scale and a Social Initiative scale were constructed. The scales showed good homogeneity in alpha coefficient analyses, testretest reliability in terms of stability across 1 year, and validity in terms of interrater agreement (mother-father and parentteacher), correspondence with observed social behavior with peers, and capacity to discriminate between peer status groups. One question in the present study concerned the relation between prosocial behavior and social initiative. Differences in content in existing competence measures (Goodman, 1994; Tremblay et al., 1992; Waters et al., 1979) as well as research in the peer status domain (Coie et alM 1990; Newcomb et al., 1993) suggested that these might constitute distinct competence aspects. These expectations were borne out, as our analyses identified one prosocial orientation and one social initiative aspect. The common denominator of our Prosocial Orientation factor could be seen as an inclination to behave smoothly in normal social interactions, coupled with abilities to perform adequately in more troublesome situations. In the Social Initiative factor, the items describe a social style of active participation (e.g., leads play activities, hesitant with peers reversed) but identify relatively few specific skills and behaviors. In this respect, it is similar to the conceptualization of peer competence by Waters and coworkers (Waters et al., 1979) and by Hagekull and Bohlin (1994). The two competence aspects were valid for both sexes and found in both parent and teacher ratings of 7to 10-year-old children from varying educational backgrounds. In the present study, more social initiative items were included in the item pool than have previously been used. The broader coverage of initiative behavior seems a reasonable explanation for the identification of two rather than one competence aspect, as has been found in earlier studies (Goodman, 1994; Matson et al., 1983; Tremblay et al., 1992). There was, however, a modest intercorrelation of the two scales. Thus, prosocial orientation and initiative should not be regarded as independent and unrelated competence dimensions but rather as reflecting two aspects with common variance within a broader social competence construct.

Table 6 Social Competence Scores Among Rejected, Neglected, Average, and Popular Children in the Provincial Sample Scale

Rejected {n = 16)

Neglected (n = 19)

Average in = 19)

Popular (n = 17)

2.93 0.94

3.79 0.70

3.65 0.80

4.11 0.39

3,70

7 gg***

3.32 1.13

3.42 1.06

3.84 0.85

4.49 0.38

3,70

5.83**

df

F

Prosocial Orientation

M SO Social Initiative M SD a

ab

ae

Significant difference between the rejected and the popular group. b Significant difference between the rejected and the average group. c Significant difference between the popular and average group. **/? < .01. ***p < .001.



Properties of the Social Competence Scales Test-retest reliability was estimated in terms of stability across 1 year, on the assumption that the phenomena under study are reasonably stable in a short perspective. Some change in children's social competence behavior between 8 and 9 years of age may be expected, however, and reliability estimated this way may involve error variance due to maturation. Still, the moderately high cross-time correlations for mother and teacher ratings, after controlling for the intercorrelation of the two scales at Time 1 and Time 2, exceeded the levels previously reported for teachers' prosocial ratings across 1 year (Tremblay et al., 1992) and should attest to acceptable test-retest reliability of the social competence scales. Agreement between raters, correspondence of ratings to observed behavior, and capacity to discriminate between peer status groups would point to the scales as valid measures of the social competence aspects. Concurrent rater agreement was consistent and substantial on the Social Initiative scale but was less consistent regarding the Prosocial Orientation scale, especially between parents and teachers. The relatively modest parentteacher correlations for prosocial behavior were similar to the amount of parent-teacher agreement found in prior studies (Goodman, 1994; Tremblay et al., 1992). Interrater agreement across 1 year was ascertained for both scales with two exceptions: teacher prosocial orientation ratings did not predict mother and father ratings a year later. Taken together, interrater concordance data point to somewhat different properties of the two scales in that prosocial orientation ratings were less consistent across context, concurrently, and across time than were social initiative ratings. A parallel could be drawn to parent-teacher agreement on the two broad-spectrum behavior problem dimensions, externalizing and internalizing problems (Achenbach & Edelbrock, 1978). Generally, parents and teachers have agreed fairly well (correlations around .45) about externalizing problems but considerably less about internalizing problems (Rydell, Dahl, & Sundelin, 1995; Stanger & Lewis, 1993; Verhulst & Akkerhuis, 1989). One explanation (cf. Verhulst & Akkerhuis, 1989) would be that externalizing problem behaviors are more obvious and easier to see for an observer compared with internalizing behaviors. Similarly, social initiative behaviors may be more visible, whereas prosocial orientation ratings require more judgment and inference, which would explain the difference in agreement for the two scales. In this context, it should be noted that teacher ratings in first grade did not predict mother and father prosocial orientation ratings 1 year later, whereas mothers' prosocial orientation ratings were predictive of teachers' ratings across 1 year. Perhaps, Ihe family environment may be a better arena for observations of the whole range of subtle prosocial behaviors than the school context, and mothers, knowing their children better, may be more observant of these behaviors than teachers are. Hypothetically, the teachers' better acquaintance with the children in second grade compared to first grade may have resulted in more accurate observations at Time 2 of the prosocial orientation behaviors that mothers had noticed already at Time 1. Besides indicating that the Social Initiative scale reflects behaviors that are easily noted by social partners, the more consis-

831

tent validity results for this scale also suggest that social initiative may be less influenced by context than are prosocial orientation behaviors. The level of social initiative may be a relatively situational stable characteristic of the child. Considering also that lack of withdrawal seems to be a general feature of the Social Initiative scale and that two items tap response to unfamiliar people, one may speculate about possible roots of this competence aspect in the temperamental dimension of behavioral inhibition (Kagan, 1989). Behavioral inhibition has been proposed by Rubin and coworkers (Rubin, Hymel, Mills, & Rose-Krasnor, 1991; Rubin, Coplan, Fox, & Calkins, 1995) as a possible antecedent to failure to develop social skills. Such suggestions can be tested only in longitudinal designs, but the content and the psychometric features of the Social Initiative scale indicate that an investigation of links from early inhibition might prove fruitful in efforts to uncover antecedents to social initiative in middle childhood. Both scales were related to positive peer behavior in the school setting. It is important to note that the differences in content of the two scales were reflected in differential relations to observed behaviors. With the observational items of the Positive Peer Behavior scale grouped to match the content of the rating scales, each rating scale predicted its corresponding behaviors only. The coefficients for the relations between ratings and observations were, however, modest in size. This may partly be due to the fact that the observations did not perfectly match the rating scales as they had not been primarily designed for validation purposes. Also, the relatively low (a = .61) internal consistency of the Positive Peer Behavior scale may have served to attenuate the size of the coefficients. Both competence scales discriminated between popular and rejected children. With regard to comparisons with the average group, our results were highly consistent with what one would expect from peer status research (Coie et al., 1990; Newcomb et al., 1993): The popular group was characterized mainly by high social initiative scores but was high also in prosocial orientation. In contrast, the main characteristic of the rejected group was low prosocial orientation scores, whereas the rejected children were only marginally lower than average in social initiative. The neglected group did not differ from the average group on either scale, which is consistent with Newcomb and coworkers' conclusion that neglected children are behaviorally discriminated from average children mainly by peer reports and observations. In adult reports, they seldom differ from average children. Thus, our results attest to the validity of the scales, as both had the capacity to discriminate between children that differ in peer acceptance, a variable considered crucial in development. Further, the value of assessing the two competence aspects separately is suggested by differential relations of the two scales to peer rejection and to peer popularity. Finally, the differences in content between the scales point to possible differential relations to other variables. For instance, empathy, emotional responses to others' emotional experiences (Bryant, 1982) could be assumed to be associated with the Prosocial Orientation scale, which has items on emotional support, rather than with social initiative. In a study by our research group (unpublished data), this prediction was borne out. Teenagers (« = 207) completed a self-report version of the Social

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KYDELL, HAGEKULL, AND BOHLIN

Competence Inventory and a Swedish adaptation for teenagers of Bryant's Empathy Index for Children (Bryant, 1982). Prosocial orientation was related to empathy (r = .53, p < .001), whereas there was no relation (r = .01, ns) between initiative and empathy. This finding constitutes some evidence for construct validity of the Prosocial Orientation scale,


Conclusion In the present article, two factor analytically based scales were found to represent the distinct but correlated competence aspects of prosocial orientation and social initiative. The psychometric properties of the established scales should permit researchers to use them to explore individual differences in children's social competence. The fruitfulness of having separate measures for these two areas, instead of a broad competence measure encompassing both, should be further tested in relations of the scales to other variables. The separation of the two aspects should prove to be of considerable theoretical value if differential relations to social outcomes and to antecedent factors were to be demonstrated. References Achenbach, T.M., & Edelbrock, C. S. (1978). The classification of child psychopathology: A review and analysis of empirical efforts. Psychological Bulletin, 86, 1275-1301. Achenbach, T. M., Howell, C. T., Quay, H. C , & Conners, C. K. (1991). National survey of problems and competencies among four- to sixteenyear-olds. Monograph of the Society for Research in Child Development, 56, (3, Serial No. 225), 1-119. Anderson, S., & Messick, S. (1974). Social competency in young children. Developmental Psychology, 10, 282-293. Asher, S., & Dodge, K. A. (1986). Identifying children who are rejected by their peers. Developmental Psychology, 22, 444-449. Blechman, E. A., Tinsley, B., Carella, E. T., & McEnroe, M. J. (1985). Childhood competence and behavior problems. Journal of Abnormal Psychology, 94, 70-77. Bryant, B. K. (1982). An index of empathy for children and adolescents. Child Development, 53, 413-425. Cattell, R. B., Balcar, K. R., Horn, J. L., & Nesselroade, J. R. (1969). Factor matching procedures: an improvement of the s index; with tables. Educational and Psychological Measurement, 29, 781-792. Cauce.A. M. (1987). School and peer competence in early adolescence: A test of domain-specific self-perceived competence. Developmental

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Received October 25, 1995 Revision received January 9, 1997 Accepted January 9, 1997 •

Call for Nominations The Publications and Communications Board has opened nominations for the editorships of Experimental and Clinical Psychophannacology, Journal of Experimental Psychology: Human Perception and Performance (JEPrHPP), Journal of Counseling Psychology, and Clinician's Research Digest for the years 2000-2005. Charles R. Schuster, PhD, Thomas H. Carr, PhD, Clara E. Hill, PhD, and Douglas K. Snyder, PhD, respectively, are the incumbent editors. Candidates should be members of APA and should be available to start receiving manuscripts in early 1999 to prepare for issues published in 2000. Please note that the P&C Board encourages participation by members of underrepresented groups in the publication process and would particularly welcome such nominees. Self-nominations are also encouraged. To nominate candidates, prepare a statement of one page or less in support of each candidate and send to Joe L. Martinez, Jr., PhD, for Experimental and Clinical Psychophannacology Lyle E. Bourne, Jr., PhD, for JEP:HPP David L. Rosenhan, PhD, for Journal of Counseling Psychology Richard M. Suinn, PhD, for Clinician's Research Digest

Send all nominations to the appropriate search committee at the following address: Karen Sellman, P&C Board Search Liaison Room 2004 American Psychological Association 750 First Street, NE Washington, DC 20002-4242 First review of nominations will begin December 8, 1997.