Child Development, September/October 2014, Volume 85, Number 5, Pages 2062–2073

Growth of Social Competence During the Preschool Years: A 3-Year Longitudinal Study Ant
onio J. Santos

Brian E. Vaughn

ISPA–University Institute

Auburn University

In^es Peceguina and Jo~ ao R. Daniel

Nana Shin

ISPA–University Institute

Ewha Womans University

This study examined the stability and growth over a 3-year period of individual differences in preschool children’s social competence, which was assessed in three domains: social engagement/motivation, profiles of behavior and personality attributes characteristic of socially competent young children, and peer acceptance. A total of 255 children (126 girls and 129 boys) participated in this study. Growth curve analyses demonstrated both stability and change with regard to social competence over early childhood. Social competence measures and latent variables were invariant over this time period, individual differences in social competence were largely stable from year to year, and significant increases over time were observed for the domain most closely reflective of specific personal attributes skills.

Over the past 40 years, early childhood peer groups have been recognized as important socialization contexts (e.g., Hartup, 1970, 1989; Rubin, Bukowski, and Laursen, 2009; Rubin, Bukowski, & Parker, 2006) and in some respects the impact of peer influences on behavior, relationships, and social adaptation more generally can be as important as the impact of parents and families (e.g., Harris, 1995, 2009). Because the majority of women in the paid labor force place their children in some form of group care (U.S. Bureau of Labor Statistics, 2010), nearly 60% of children from middle-income families in United States interact with peers before they enter kindergarten (Federal Interagency Forum on Child and Family Statistics, 2013). This social fact has motivated many different programs of research intended to characterize and measure the quality of social adaptation among children with extensive group care experiences (e.g., Vaughn et al., 2009) as well as to document the broader implications of early, extensive, nonparental care experiences (e.g.,

Data collection and preparation of this article has been supported in part by Grants POCTI/PSI/46739/2002, PTDC/PSI/ 66172/2006, and PEst-OE/PSI/UI0332/2011 from the Portuguese Foundation for Science and Technology, and by Grants SBR9514563, BCS99-83391, BCS01-26163, BCS06-23019, and BCS0843919 from the U.S. National Science Foundation. Correspondence concerning this article should be addressed to Ant
onio J. Santos, ISPA–University Institute, Rua Jardim do Tabaco 34, 1149-041 Lisboa, Portugal. Electronic mail may be sent to [email protected].

National Institute of Child Health and Human Development, 2006). It is generally accepted that social transactions with peers are the basis of friendships, social networks, and peer status for school-age children and adolescents (e.g., Hartup, 1970; Hay, Payne, & Chadwick, 2004; Parker & Asher, 1993; Rubin et al., 2006) and a growing body of evidence suggests that these kinds of effects are seen during early childhood, as well (e.g., Howes, 2009; Ladd, 2005; Santos, Vaughn, & Bost, 2008). Ladd (2005) reviewed a large literature indicating that the degree to which a child can be considered “successful” in the preschool peer group forecasts future social adjustment and academic achievement, as well as deviance and psychopathology (i.e., when children are less “successful”). Given these implications of peer experiences in early childhood, there has been renewed research interest on just what it means to be a “successful,” or “socially competent” preschool child (e.g., Bost, Vaughn, Washington, Cielinski, & Bradbard, 1998). This study is a component of a larger research program on social and emotional adaptation during early childhood and considers the longitudinal stability and growth of social competence over the preschool years. Following from the arguments in Waters and © 2014 The Authors Child Development © 2014 Society for Research in Child Development, Inc. All rights reserved. 0009-3920/2014/8505-0023 DOI: 10.1111/cdev.12246

Growth of Social Competence

Sroufe (1983) and from the results reported by Shin et al. (2011), we expected to find substantial stability of individual differences with respect to social competence over time and also expected to detect meaningful positive growth with respect to social competence (i.e., older children should be more socially competent than younger children). The larger research program was grounded, in part, in a conceptual framework regarding social competence and its growth over early childhood that was first articulated by Waters and Sroufe (1983), who argued that social competence was the central organizing construct of early childhood. They suggested that social competence summarizes children’s capacity to recruit personal and interpersonal resources in the service of achieving their goals in social groups, while maintaining a good developmental trajectory. This theoretical framework has been used to characterize and measure social competence in cross-sectional (e.g., Bost et al., 1998; Santos, Peceguina, Daniel, Shin, & Vaughn, 2013; Vaughn et al., 2009) and short-term longitudinal (e.g., Shin et al., 2011) studies of children in child-care settings. These studies have established that social competence can be construed as a multilevel (or hierarchical) latent construct that is related to domains of social engagement/social motivation, individual level traits, tendencies, and behaviors, and to peer acceptance. Each of these three domains captures important aspects of the Waters and Sroufe (1983) conceptual description of the social competence construct. Vaughn et al. (2009) believed that social engagement/social motivation served as the foundational indicator for social competence insofar as: (a) engagement provides opportunities to broaden and build behavioral and cognitive skills required for successful goal achievement within social contexts, (b) affords opportunities to discover the goals of interactive partners, and (c) provides interaction partners with critical information about the child’s relative value as a preferred playmate. They further reasoned that the frequency of initiated social engagement was prima facie evidence for social motivation. It has also been common in the social development literature covering early childhood to characterize social competence in terms of the content of social interactions and the child attributes associated with interaction contents (see Ladd, 2005, for a review of this literature). Vaughn et al. (2009) used Q-sort descriptions of children to assess a broad range of such interaction qualities and personal attributes. Q-data are quite useful for this kind of broad-band assessment (see Block & Block, 1980, for a discussion of the utility

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of Q-data). The specific Q-sort descriptions have the additional positive quality of having been used in prior research to characterize very social competent preschool-age children (e.g., Waters, Garber, Gornal, & Vaughn, 1983). Peer acceptance also has a long history as a social competence indicator in the social development literature (see Ladd, 2005; Rose-Krasnor, 1997, for reviews) and is typically operationalized in terms of peer sociometric choice data. Peer acceptance serves as an indicator of both likability and integration in the classroom peer group, which reflect the important social goal of establishing and maintaining alliances and positive relationships within salient social groups. Shin et al. (2011) also showed that individual differences were stable over consecutive academic years (i.e., 3–4 years of age); however, their sample could not be tested during the kindergarten year because individual children dispersed to many different kindergartens after preschool. Studying a sample from preschool to kindergarten would be of interest for both conceptual and empirical reasons. First, in the United States it is common that kindergarten marks a transition from a less to more structured experience of school curricula, which can challenge the child’s social and cognitive resources and may lead to a reorganization of those resources (see Ladd, 2005). However, in typical U.S. settings the child is challenged in several ways (e.g., move to a new institutional setting, changes in classroom peer group composition, and exposure to new curricula and teacher expectations for behavior). Second, having three waves of data spaced approximately equally over years affords possibilities of using statistical procedures (e.g., latent variable growth models) that cannot be used when measurement is completed at only two time points. In this study, longitudinal data were collected for children three times during early childhood (i.e., at ages 3, 4, and 5) for groups that remained in the same setting for both preschool and kindergarten, thus reducing the challenges due to setting and peer group composition, which allows us to consider only the shifts of teacher expectations and curriculum structure as the primary changes to which the child must adjust. Our intention was to test the cross-time stability of individual differences in social competence over this period and also to test the possibility of normative linear growth with respect to social competence in a sample of children recruited from preschool programs in Portugal. Finding coherence in the structure of the social competence construct, predictability of later social competence from earlier assessments, and growth

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Santos, Vaughn, Peceguina, Daniel, and Shin

over early childhood would constitute support for the Waters and Sroufe (1983) conceptualization of social competence as an organizing construct, and would also provide evidence for the cross-national validity of the hierarchical model, previously described by Santos et al. (2013) and Shin et al. (2011). Having a valid, temporally stable, and behaviorally based index of social competence would also facilitate future studies to directly examine the peer group as an agent of socialization during early childhood. The choice to conduct this study in Portugal is fortuitous for several reasons. First, the trend to increasing the numbers of children participating in group care parallels the experience in North America but started at a somewhat earlier point in time because women with children began entering the Portuguese labor force in large numbers during the 1970s and the government recognized the value of providing organized care by funding child-care programs starting at 3 years of age. Thus, 82.3% of Portuguese children attend a preschool program (GEPE, 2010). Moreover, the state instituted a program for training early childhood educators at the university level (typically a 5-year course of instruction and practical experiences), so the teacher cadre is well versed in pedagogical practices appropriate for young children and also receives training in early childhood development. Finally, it is the practice in Portugal for children to remain with the same group of peers throughout their years in preschool and kindergarten (with many remaining in the same group through primary school). These stable groups provide an ideal context to detect stabilities for the dimensions considered to be indicators of peer social competence, as well as examinations of normative changes over early childhood. Comparisons between Portuguese and U.S. samples reported previously (e.g., Santos et al., 2013) suggest that the social competence construct is measured equally well with the battery of measures described by Vaughn et al. (2009) and we did not expect to find marked differences in terms of model structure or cross-time stability, although stability of peer group membership could increase the magnitude of stability paths when compared to the sample described by Shin et al. (2011). The model of social competence adopted by Vaughn et al. (2009) includes the three indicator domains described above, with each domain being assessed using multiple measures. The instruments used as indicators for Peer Acceptance do not allow for positive or negative “growth” over time, because children at every age level make a fixed number of

choices that differ across instrument types but not across children. Therefore, the means for peer acceptance cannot change with increasing age unless the size of the social group changes. Of course, the distributions of positive choices across group members may change with age, but this would not be detected as normative growth (i.e., the means would be nearly identical at all age levels). The other two social competence domains are not constrained in this way; that is, children may increase (or decrease) their rates of engagement with peers and these changes can differ in trajectory depending on the quality of initiated interaction (e.g., positively toned interactions may increase as a function of age; Vaughn, 2001). There may also be age-related shifts in the presence and/or the expression of specific social behaviors and/or personality attributes that are reflected in the Q-sort scores for the profiles of social behavior and personality attributes domain. The data were collected over several annual waves of data collection in the research program on social adaptation mentioned above, between 2004 and 2011 in the region of Lisbon, Portugal. The sample was recruited from two preschools, both affiliated with elementary schools, serving a middle-class (by the standards of the local community) population from European ethnic backgrounds. The general model to be tested hypothesizes that the social competence construct will have a hierarchical structure with the three broad factors relevant to individual facets or domains that are subordinate to a general factor (i.e., Social Competence), and with the directly measured variables subordinate to the domain factors (Vaughn et al., 2009). On the basis of the findings of Shin et al. (2011) for an American sample observed over two time points, we hypothesized that the model would be invariant across age levels and that the second-order latent factor (Social Competence) would show significant and substantial crossage stability. Finally, we tested the profiles of behavior and personality attributes and the social engagement/motivation (latent factors for change over the 3-year age range studied). We hypothesized that positive growth would be observed for the profiles of behavior and personality attributes latent variable, but did not make a directed prediction concerning the social engagement/motivation factor.

Method Participants A total of 255 children (126 girls and 129 boys) participated in this study. They were recruited from

Growth of Social Competence

two preschools in Lisbon, Portugal (five different classrooms) and were observed in 3 consecutive years (i.e., as 3-, 4-, and 5-year-olds). All families were from European ethnic backgrounds and were middle to upper socioeconomic status in terms of their education levels, occupational titles, twoparent family status, number of children in household, and family incomes, in comparison to Portuguese standards. All assessments took place in the daycare centers. Children were observed in different settings (e.g., free play and group activities in the classroom, meals, playground, transitions between activities). Consent was obtained from school directors, teachers, and parents prior to data collection and child assent was elicited for all protocols requiring direct child involvement (e.g., sociometric interviews). Instruments Using the protocols described by Vaughn et al. (2009), social competence was operationalized with reference to three measurement families, representing three broad indicator domains: (a) Social Engagement/Motivation (SE/M), (b) Profiles of Behavioral and Psychological Attributes (PB/PA), and (c) Peer Acceptance (PA). Each domain was measured using two or three indicators, multiple observers collected data within each domain, and different teams collected data for each domain. Prior to data collection, each observer spent at least 2 hr in the classrooms to become familiar with the names of the children and also to allow the children to become familiar with him or her. Social Motivation and Engagement Rates of visual attention received, and positive and neutral interaction initiated were measured indicators of the SE/M measurement family. For interaction initiation, each child was observed for a 15-s interval. At the end of the interval, the observer recorded the identification codes of all the children with whom the focal child interacted. The affective tone of the interaction event was also recorded (i.e., the interaction was characterized as positive, neutral, or negative). To be coded as positive interaction, (a) one or both children had to clearly evidence positive affect during the social exchange and (b) the positive affect expression should not be accompanied by a negative affect expression from the interactive partner. Negative interactions were coded when one or both children expressed negative affect (e.g., anger, distress, fear, sadness) in a

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facial, vocal, or gestural mode, unless these expressions were displayed in the context of fantasy play. Social interactions that were recorded as neither positive nor negative were coded as neutral and included all the verbal and nonverbal exchanges that did not contain affect expression. Scores were the total frequencies of positive, neutral, and negative interactions initiated by the child, which were converted to rate scores by dividing the totals by the number of observation rounds for which the child had been present. These rate scores were then standardized within each classroom to adjust for children’s absences from the class during the observation period and for differences in the final number of rounds observed in each classroom. Only the standardized rate scores for positive and neutral interactions initiated (within classrooms) were retained for analysis (see Santos et al., 2013; Vaughn et al., 2009). Initiated negative interactions were not included in the study for two reasons. First, the peer relations literature suggests that negative interactions predict lower peer acceptance and higher peer rejection (Rubin et al., 2006), but it is also the case that preschool children with the highest rates of interaction overall tend to have the highest rates of negative interactions as well (Vaughn, Vollenweider, Bost, Azria-Evans, & Snider, 2003). Thus, adding negative interactions would most likely result in a less well-fitting hierarchical model. Observers made a minimum of 200 observations per child. Research assistants were trained to 80%+ exact agreement prior to data collection in live observations. The reliability of scores contributed by each observer for positive and neutral initiated interactions was calculated as the Spearman–Brown correlation between observers (completed prior to standardizaing the total scores). This correlation averaged .76 for neutral interactions and .63 for positive interactions. These levels of interrater agreement are consistent with previous reports using live observation protocols and have proven satisfactory for analysis purposes in prior reports (e.g., Santos et al., 2013). For visual attention received, the observer watched a particular child (the focal child) for a 6-s interval. After 6 s, codes identifying the children looked at by the child being observed were recorded as receiving 1 unit of visual attention each (i.e., a given child might be looked at twice in 6 s, but would only receive a score for 1 unit of visual attention for that interval). For each round, a target child was observed when his or her name appeared on the class roster and no child was observed twice

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before all children present that day were observed once. When the eye gaze direction was ambiguous, the recipient was recorded as receiving a “doubtful” occurrence (i.e., “?”). Doubtful occurrences were not considered in the computation of the child’s total score for looks received. Only 1 unit of visual regard from an observed child was credited to a single peer recipient for a given interval. As with the interaction protocol, observers completed a minimum of 200 rounds of visual attention observation in each classroom. The total scores for visual attention units received by a child were divided by the number of rounds that he or she was present, to adjust the final score for absences, and these scores were standardized within each classroom. Past research using these observation procedures has shown that observers rapidly attain agreement rates of 80% and above (exact agreement for recipient of a unit of visual attention), with only limited training periods (e.g., Bost et al., 1998; Vaughn et al., 2009), after a short period of observing live interactions. Observers for this study also reached the 80% rate within 4 hr of training. The reliability estimate (i.e., Spearman–Brown formula) for visual regard received total scores in this Portuguese sample (i.e., treating the sums for each child contributed by a single observer as an “item” in a twoitem scale) averaged .82, and ranged from .59 to .90 across all classrooms in the sample. These values for interrater agreement are consistent with prior research using this protocol for live observations and the resulting scores have proven adequate for analysis using latent variable statistical procedures (e.g., Santos et al., 2013). Profiles of Behavioral and Psychological Attributes Profiles of behavioral and psychological attributes were assessed using two Q-sets, the California Child Q-sort (CCQ; Block & Block, 1980) and the Preschool Q-sort (PQ; Bronson’s adaptation of a Q-sort originally used by Baumrind, 1967). Teams of two observers provided Q-sort descriptions after observing in a given classroom for a minimum of 20 hr. Observations were made in a variety of settings (e.g., small-group activities, meal times, free play indoors, outdoor play, etc.). Observers did not generally watch the same child at the same moment and frequently completed their observations on different days. When observations were completed, both observers described each participating child in the classroom with the CCQ-set (100 items) and PQ-set (72 items). The items were sorted into nine catego-

ries (1 = most atypical attributes and 9 = most typical attributes of the child), with a rectangular distribution. The Q-sort for a given child was subsequently correlated with the profile of a hypothetical child at the positive extreme for social competence, generated by aggregating the descriptions provided by experts on social development during early childhood (Waters, Noyes, Vaughn, & Ricks, 1985). Pearson’s correlation between a Q-sort for a given child, and the “criterion” sort for the construct was his or her “score” for that construct. During training, complete Q-sort descriptions were provided by each pair of observers (N = 8) for children who did not participate in this study. Agreement rates for the full set of items ranged between .71 and .90, for the PQ-set (M = .79), and between .69 and .90, for the CCQ-set (M = .77). Criterion-based scores were first averaged across observers for each Qsort and then the two final Q-sort criterion scores were averaged to derive the social competence composite score. These final Q-sort scores were standardized within classroom prior to further analyses. Peer Acceptance The Peer Acceptance domain was assessed using two different sociometric interviews: (a) positive nominations (McCandless & Marshall, 1957) and (b) paired comparisons (Vaughn & Waters, 1981). A research staff member individually interviewed each child outside the classroom in a quiet area. For the nomination task, children were presented with the set of photographs of all classmates and asked to name a peer that he or she especially likes to play with. The request was repeated two more times. As the child named the peers, the photographs were turned face down. Scores were the total number of choices received from peers, divided by the number of children making choices. For the paired comparison task, photographs of all the possible pairs within each classroom (i.e., N
(N 1)/2) were presented to the child, one pair at a time. For each pair, the child was asked to choose the peer with whom she or he especially likes to play. The pairs were randomly organized, and no child was seen twice before all other children were seen once. Each child’s photograph appeared the same number of times on the leftand right-hand sections of the dyad photographs. The acceptance score for this measure was the total number of choices received from peers, divided by the number of classmates who completed the task. Prior to the analyses, the scores for the two

Growth of Social Competence

sociometric measures were standardized within each classroom. Plan of Analyses All latent variable models were tested using Mplus (version 5.2; Muth
en & Muth
en, 2009). Each hypothesis was tested using latent variable methods. To test for measurement and structural invariance of the model for each age, a series of nested multigroup, confirmatory factor analyses (CFA) were computed. Cases with missing data were included in the latent variable analyses with full information maximum likelihood estimation. The first model was unconstrained and tested the fit of the hierarchical structure for each age level. This was followed by a test of invariance for the measurement model (i.e., were factor loadings for a given measured variable equivalent at all ages). The final test constrained both measurement and structural loadings to be equal. The next model examined was longitudinal and tested the year-toyear stability for the second-order factor (Social Competence). Finally, age-related changes in the social engagement/motivation and personality/ behavioral profiles domains were examined using latent variable growth curve models.

Results Model Equivalence Across Age Levels The unconstrained CFA tested whether the hierarchical model was a good fit to the data at each age level. All model fit statistics were acceptable (see Table 1). All measured variables had significant loadings on their first-order factor and the three first-order factors loaded significantly on the second-order factor for each age level. Standardized path weights for the measurement and structural paths and their associated error terms for the unconstrained model are presented in Figure 1. This constitutes the baseline model against which the subsequent models are compared. Table 1 also presents the fit statistics and model comparison values for the analyses constraining the measurement and structural pathways as well as intercepts to equivalence. Model 1 tested the equivalence of measurement loadings across age levels and this model was not significantly different from the baseline model, using the difference in chi-squares as the criterion. Model 2 constrained both the measurement and structural loadings in the model to equality and this model did not differ significantly from

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Model 1. Following recommendations made by Little, Card, Slegers, and Ledford (2007) a third model constrained measurement loadings, structural loadings, and intercepts to equivalence. Model 3 was not significantly different from Model 2. These results indicate that the same social competence construct was measured at each age level and justify our subsequent tests of social competence stability at the group level and analyses of change in the domain scores over time. Stability of Social Competence From 3 to 5 Years The next analysis tested the longitudinal stability of the SC construct across age levels (i.e., the path between social competence at Time x at Time x + 1 was estimated for each age-level change). The fit statistics for the model were acceptable, v2(df = 147) = 198.29, p < .003, v2(df = 1.35), normed fit index (NFI) = .90, nonnormed fit index (NNFI) = .95, comparative fit index (CFI) = .97, root mean square error of approximation (RMSEA) = .04. Standardized path weights for both measured and latent variables, as well as disturbance variances (numeric values over arrow points) for the first- and second-order factors are presented in Figure 2. Consistent with our hypotheses, the path from social competence at age 3 to social competence at age 4 and from social competence at age 4 to social competence at age 5 were large and significant, bs = .89 and .74, ps < .001, respectively. Thus, individual differences with respect to social competence that were observed at 3 years of age were largely reproduced at age 4 and the age 4 individual differences were also reproduced at age 5. Growth of Social Competence From 3 to 5 Years of Age Social competence, as measured using our battery of assessments, shows significant stability over early childhood; however, we would expect also that social competence should increase with age, because older preschoolers have had opportunities to broaden and build their skill sets as a function of their experiences with peers. As noted above, only the PB/PA and SE/M domains in our model could show measureable growth over time. The means and standard deviations for the measured variables used as indicators for these domains are presented in Table 2. For completeness, the means and standard deviations for the sociometric data are also included in Table 2. A visual examination of the values in Table 2 suggests small directional changes for the two Q-sort scores (i.e., increases across age

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Santos, Vaughn, Peceguina, Daniel, and Shin

Table 1 Factorial Invariance Across Years: Goodness-of-Fit Statistics

Unconstrained model Model 1: Measurement loadings Model 2: Structural loadings Model 3: Intercepts

v2

df

v2/df

NFI

NNFI

CFI

RMSEA

Δv2

AIC

55.35** 63.93* 71.64** 72.51

33 41 45 59

1.68 1.56 1.59 1.23

.96 .95 .95 .95

.96 .96 .96 .99

.98 .98 .98 .99

.03 .03 .03 .02

8.58 7.70 .88

199.35 191.93 191.64 164.51

Note. NNFI = nonnormed fit index; CFI = comparative fit index; RMSEA = root mean square error of approximation; AIC = Akaike’s information criterion. *p < .05. **p < .01. Four Year Olds

Three Year Olds SC

.49 .76

CCQ

.34

.71

.91 .17

.73 PQ

.47

.82

.79

.61

Visual Initiate Attention Positive ..38

.62

.96

.33

.50

.66 Initiate Neutral

Nominations

.56

.35

.67

.85 Paired Comparisons

.85

CCQ

.28

.66

Peer Acceptance

SE/M

PB/PA

.59

.81

.57

.09

Peer Acceptance

SE/M

PB/PA

.82

SC

.71

.56

Visual Initiate Attention Positive

PQ

.28

.50

.68

.71

.70

.93

Initiate Neutral

Nominations

Paired Comparisons

.49

.51

.14

Five Year Olds SC

.89 .21

CCQ

.12

.83 PQ

.31

.59 Peer Acceptance

SE/M

PB/PA

.94

.64

.81 .34

.87

.55

Visual Initiate Attention Positive

.25

.70

.69 Initiate Neutral

.52

.77

.88

Nominations

Paired Comparisons

.44

.23

Figure 1. Testing the invariance of the model across age levels. SC = Social Competence; PB/PA = Profiles of Behavior and Personality Attributes; SE/M = Social Engagement/Motivation; CCQ = California Child Q-sort; PQ = Preschool Q-sort.

levels) as well as increases for the initiated interaction variables (but not visual attention received) from the 3- to 4-year age level. However, from the 4- to 5-year levels, values for all three indicators of the SE/M domain decline. This change was not anticipated a priori and we return to this result in the Discussion. As anticipated by the arguments above, no age-related changes in means were observed for the sociometric variables. To test our hypothesis that social competence should increase over time, linear growth models were

tested separately for the PB/PA latent factor and for the SE/M latent factor. The general linear growth model with multiple indicators tested for both domains is presented in Figure 3 with the explicit hypothesis being that the slope term would be positive and significantly different from zero. The unconditional model fit the observed data adequately for the PB/PA domain, v2(df = 11) = 19.22, p < .05; CFI = .98, NNFI = .98, RMSEA = .053, standardized root mean square residual (SRMR) = .034, with a mean for the slope term of .03, p < .001, indicating a

Growth of Social Competence Age 3

Age 4

Age 5

.203 .893

SC .567 .678

.905 .602 .181

PB/PA .813

Q1

SE/M

.742 .812 Q2

.637

VR

PA .816

.761

.835

IN

PCS

Q1

.688

Q2

.527

SE/M

VR

SC

.352

PA .726

.722

NS

.741

.861

..559 .663 IP

.949 .100

PB/PA .616

.260 SC

.420

2069

.805

PB/PA .856

.939

IN

SE/M

PCS

Q1

PA .746

.880

NS

.568

.832

.560 .686 IP

.884 .657 .219

Q2

.909

.552 .690

VR

IP

NS

IN

PCS

Figure 2. Longitudinal stability of social competence from year to year. SC = Social Competence; PB/PA = Profiles of Behavior and Personality Attributes; SE/M = Social Engagement/Motivation; PA = Peer Acceptance; Q1 = California Child Q-sort; Q2 = Preschool Q-sort; VR = visual attention received; IP = initiated positive interactions; IN = initiated neutral interactions; NS = nominations sociometric; PCS = paired comparisons sociometric.

Table 2 Descriptive Statistics for Measured Indicators of Social Competence at Three Age Levels

Indicator CCQ PQ Visual attention Initiate positive Initiate neutral Positive nominations Paired comparisons

Age 3 M (SD) 0.09 0.09 0.58 0.09 0.34 0.14 9.6

(0.19) (0.23) (0.24) (0.08) (0.18) (0.11) (2.3)

Age 4 M (SD) 0.12 0.13 0.58 0.13 0.38 0.15 9.9

(0.19) (0.22) (0.27) (0.10) (0.16) (0.11) (2.4)

Intercept

Age 5 M (SD) 0.17 0.16 0.48 0.10 0.35 0.15 9.98

(0.17) (0.28) (0.25) (0.07) (0.14) (0.11) (2.7)

Note. Values in table are not standardized. Q-sort values are partial correlations (social desirability controlled). Visual attention and interaction values are rate scores (events per unit of observation). Sociometric values are averages (total number of sociometric choices received divided by the number of children making choices). CCQ = California Child Q-sort; PQ = Preschool Q-sort.

small but significant incremental increase in the scores for this domain over time, equivalent to approximately 34% of a standard deviation increase in the level of social competence for each year that the child was in the preschool. Moreover, the negative correlation between the intercept and slope terms in the model (r = .59) indicates that children with lower initial scores tended to gain more in subsequent years. However, the unconditional growth model did not fit the data for the SE/M domain, v2(df = 33) = 263.82, p < .001; CFI = .75, NNFI = .73, RMSEA = .16, SRMR = .16.

Discussion We suggested at the outset that indicators of social competence should be observable for children in these Portuguese preschools and that individual dif-

1

1

Slope

1

DOMAIN 11

V11

0

1

2

DOMAIN 21

V21

V12

DOMAIN 31

V22

V31

V32

-.008 (-.590)

1 .006

Intercept 0.00 (0.00)

1

Slope 0.03 (0.43)

0 1

1

.012 PB/PA

2

-.006

PB/PA

CCQ1

PQ1

CCQ2

PQ2

CCQ3

.008

.011

.009

.010

.019

PB/PA

PQ3 .01

Figure 3. Values in parentheses for intercept and slope terms are standard deviations. PB/PA = Profiles of Behavior and Personality Attributes; CCQ = California Child Q-sort criterion score for social competence; PQ = Preschool Q-sort criterion score for social competence.

ferences with respect to social competence should show stability from year to year, perhaps especially when membership in the group is stable over consecutive years, as was true for this sample. Our results are consistent with these assumptions. The social competence construct was adequately measured in children as young as 3 years of age and revealed a hierarchical structure that proved invariant over the 3 years of assessment. That is, the measurement and the meaning of the social competence construct did not change from 3 to 5 years of age.

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Furthermore, our analysis of associations between social competence at Time 1 (age 3) and at Time 2 (age 4), as well as between social competence measured at Times 2 and 3, were substantial and significant. Thus, variability with respect to social competence in the 1st year that the group was formed accounted for 2-year variability, and variability in the 2nd year accounted for variability in the 3rd year. These findings are consistent with conclusions offered by Campbell, Lamb, and Hwang (2000) to the effect that individual differences in children’s social skills begin to stabilize at about 3.5 years of age (also see Howes, 1988). The results are also consistent with assumptions of intraperson coherence that are central to the Waters and Sroufe (1983) model of competence. Our findings suggest that this intraperson coherence is not necessarily vulnerable to the stresses of transition (to kindergarten) that are associated with changes in teacher expectations and classroom curricula, although it remains to be determined whether and how institutional and class composition modifications may induce changes with respect to child social competence. Our stability results replicate and extend findings from the two-wave longitudinal study reported by Shin et al. (2011). Children in the Shin et al. study had been assessed using the same set of social competence indicators employed here and also showed significant cross-age stability for the global social competence composite, although at a somewhat lower magnitude (b = .74) than we observed (b = .89). It is possible that the stability of group membership in Portuguese preschool classrooms contributed to this difference, but the path coefficients from age 3 to age 4 were significant in both samples. Thus, while stable group composition may increase estimates of stability for social competence, this does not appear to be the primary source of stability. There are relatively few longitudinal studies of social competence focused on the period of early childhood (but see Howes, 1988), and we were not able to find a single study that assessed children in 3 consecutive years using the same measures at each age. In part, this reflects the structure of early childhood education and child day care in North America. Groups formed when children are about 3 years of age may (but often do not) remain in the same facility as 4-year-olds, but enter kindergarten in a different institution (i.e., elementary school) after their fifth birthdays. Consequently, most published longitudinal studies concerning children in this age range used multiple observations in a sin-

gle academic year (e.g., Roseth, Pellegrini, Bohn, Van Ryzin, & Vance, 2007) or over 2 consecutive years (e.g., Howes, 1988; Shin et al., 2011). Many studies that assess social competence during early childhood are concerned with its antecedents in parent–child interaction histories and children’s relationship representations (e.g., Denham, Renwick, & Holt, 1991; Sroufe, Egeland, Carlson, & Collins, 2005). The central premise of these studies has been that child social competence is a consequence of these interaction/relationship histories. However, the magnitude of associations in these studies often is modest and/or restricted to one or the other sex (e.g., Denham et al., 1991) and it is rare that indicators of child social competence are directly assessed in the context of peer interactions. This study stands in contrast to most others in this regard. Our definition of social competence (i.e., success in accomplishing personal goals in the social context of the peer group) is broad compared to definitions used in other studies and our measures are based on direct observations of behavior and sociometric interviews with child participants. We view our measurement model as representing social competence from the point of view of the children themselves, rather than from an adult-centered perspective. Furthermore, no single domain of measurement adequately captures the range of what our social competence construct implies. It is the latent aggregate of multiple domains that serves as our primary social competence indicator and it was this indicator that showed substantial cross-age stability. Thus, while it may be that the initial levels of child social competence are due, in part, to child/adult transaction histories, we suggest that the stability of child social competence seen here and in our other studies (e.g., Shin et al., 2011) arises from group processes involving the children themselves. We were also interested in the growth/change in social competence and, because data were obtained over 3 consecutive years, we could use more modern statistical techniques to evaluate growth than was possible in the Shin et al. (2011) report. Two of the social competence indicator domains included measures that could show normative “growth” over this range of ages but only one of these latent variables yielded results consistent with a linear growth hypothesis. Children’s scores for PB/PA increased by about one third of a standard deviation each year, indicating that individual child profiles became more similar to the hypothetical “most socially competent preschool child” (Waters et al., 1983) over time.

Growth of Social Competence

The SE/M latent factor did not show a comparable annual increase. This finding was not anticipated, in part because the associations between this factor and the other latent factors were significant and substantial at all age levels. In retrospect, this pattern of change may be explained by changes to the structure of the curriculum that were implemented in the oldest preschool classrooms. Beginning at age 5, children were assigned “seat work” (e.g., number and letter worksheets) in preparation for their transition to elementary school, and their opportunities for freely chosen activities and play partners were reduced. Moreover, teachers supervised the older children more closely to assure that their attention remained on the assigned work, rather than on peers (even so, children received visual attention from peers on approximately 48% of observation intervals and initiated interactions in at least 44% of observation intervals). These curricular changes may account for the observed declines in visual attention received from peers and in the initiated interaction categories from 4 to 5 years of age (see Table 2). Our growth curve analyses not only highlight strengths of our model and approach but also identify some of its limitations. First, the Waters and Sroufe (1983) model does not define social competence in terms of specific attributes of children, their goals, or the specific content of their achievements. Rather, it defines social competence as the flexible application of available resources (both personal and interpersonal) in service of goal attainment, however the children themselves might define those goals. This implies that some of children’s goals may not be considered to be conventionally “prosocial” and that the means used to attain some goals may involve coercive and/or “antisocial” behaviors (e.g., Hawley, 2002; Roseth et al., 2007). Socially competent children may, at times, behave aggressively but the key idea is flexibility in the use of such tactics. A child who was inflexible in the choice of aggressive or otherwise coercive tactics for goal achievement would not be considered socially competent by this definition. Our measurement model reflects this definitional approach by the use of broadband assessments that capture a wide range of behavioral tactics and strategies relevant to goal selection and attainment. Moreover, these indicators show moderate to strong positive associations with each other and with the global SC latent variable at each age level and the longitudinal associations for social competence are quite strong. At the same time, we recognize that our indicators are just that; they are not themselves “socially

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competent acts.” Peer acceptance is an indicator of social competence because it has numerous correlates in the tactics and personal attributes that lead to successful goal achievement, but peer acceptance per se is a group property insofar as it reflects the evaluations of peers concerning the relative value of the target child as a playmate (Hartup, 1970; Rubin et al., 2006). Likewise, receiving visual attention is an indicator of SC because peer attention is associated with many behavioral attributes supportive of goal achievement and social integration (e.g., Vaughn & Martino, 1988; Waters et al., 1983) and not necessarily because preschool children have receiving attention from peers as a goal. In addition, receiving visual attention depends on the choices of peers and not on the rate of attending to peers oneself. Thus, as with peer acceptance, receiving attention is a group level process. Of our indicators, the Q-sort descriptions come the closest to describing goals and the tactics a child might employ in their achievement. Interestingly, PB/PA is the only measured domain for which we found significant positive growth over time. Apparently, children improve in their abilities to act in the service of goal achievement as a function of age (and/or experience in their stable groups) and the greatest amount of change occurs among the children demonstrating lower levels of skill in the previous year. We also recognize that the data reported here have an inherent nestedness, in the sense that behaviors are nested within children over time and children are nested within classrooms (also over time). We did not use mixed (or multilevel) models to analyze these data for several reasons, the most critical of which was the fact that with so few classrooms at each age level the study was underpowered to detect any but the largest of effects. As we continue this program of research and increase the number of classrooms, we intend to test for potential classroom level effects, but at this point we acknowledge that our interpretations are limited by the possibility that undetected classroom effects were influencing the results. To conclude, this study demonstrates both stability and change with regard to social competence over early childhood. Our findings support the conceptual model of Waters and Sroufe (1983) and add to the empirical base for that model provided by previous studies of preschool-age children. Because the model adopts a broadband definition of social competence that yields a valid and reliable index of the social competence construct, it should be useful to investigators studying both the antecedents and

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the consequences of early childhood social competence with peers.

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Vaughn, B. E., Vollenweider, M., Bost, K. K., Azria-Evans, M. R., & Snider, J. B. (2003). Negative interactions and social competence for preschool children in two samples: Reconsidering the interpretation of aggressive behavior for young children. Merrill-Palmer Quarterly, 49, 245–278. doi:10.1353/mpq.2003.0017 Vaughn, B. E., & Waters, E. (1981). Attention structure, sociometric status, and dominance: Interrelations, behavioral correlates, and relationships to social competence. Developmental Psychology, 17, 275–288. doi:10. 1037/0012-1649.17.3.275 Waters, E., Garber, J., Gornal, M., & Vaughn, B. E. (1983). Q-sort correlates of social competence. Developmental Psychology, 19, 550–560. doi:10.1037/0012-1649. 19.4.550 Waters, E., Noyes, D., Vaughn, B. E., & Ricks, M. (1985). Social competence and self-esteem: A Q-sort analysis of conceptual and empirical similarities between related constructs. Developmental Psychology, 21, 508–522. doi:10.1037/0012-1649.21.3.508 Waters, E., & Sroufe, L. A. (1983). Social competence as a developmental construct. Developmental Review, 3, 79–97. doi:10.1016/0273-2297(83)90010-2