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Temperamental, Parental, and Contextual Contributors to Early-Emerging Internalizing Problems: A New Integrative Analysis Approach

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Title: Temperamental, Parental, and Contextual Contributors to Early-Emerging Internalizing Problems: A New Integrative Analysis Approach
Language: English
Authors: Mills, Rosemary S. L., Hastings, Paul D., Helm, Jonathan, Serbin, Lisa A., Etezadi, Jamshid, Stack, Dale M., Schwartzman, Alex E., Li, Hai Hong
Source: Social Development. May 2012 21(2):229-253.
Availability: Wiley-Blackwell. 350 Main Street, Malden, MA 02148. Tel: 800-835-6770; Tel: 781-388-8598; Fax: 781-388-8232; e-mail: cs-journals@wiley.com; Web site: http://www.wiley.com/WileyCDA/
Peer Reviewed: Y
Page Count: 25
Publication Date: 2012
Document Type: Journal Articles
Reports - Research
Descriptors: Structural Equation Models, Parenting Styles, Inhibition, Risk, Personality, Meta Analysis, Disadvantaged, Behavior Problems, Children, Mothers, Rejection (Psychology), Stress Variables, Correlation, Context Effect, Social Development
Geographic Terms: Canada
DOI: 10.1111/j.1467-9507.2011.00629.x
ISSN: 0961-205X
Abstract: This study evaluated a comprehensive model of factors associated with internalizing problems (IP) in early childhood, hypothesizing direct, mediated, and moderated pathways linking child temperamental inhibition, maternal overcontrol and rejection, and contextual stressors to IP. In a novel approach, three samples were integrated to form a large sample (N = 500) of Canadian children (2-6 years; M = 3.95 years; SD = 0.80). Items tapping into the same constructs across samples were used to create parallel measures of inhibited temperament, maternal positive, critical, and punitive parenting, maternal negative emotionality, family socioeconomic and structural stressors, and child's IP. Multiple-groups structural equation modeling indicated that associations were invariant across samples and did not differ for boys and girls. Child inhibition, less positive and more critical parenting, maternal negative emotionality, and family socioeconomic disadvantage were found to have direct associations with IP. In addition, maternal negative emotionality was associated with IP through more critical parenting, and both maternal negative emotionality and socioeconomic stress were associated with IP through less positive parenting. Results highlight the multiple independent and cumulative risk factors for early IP and demonstrate the power of integrating data across developmental studies.
Abstractor: As Provided
Number of References: 80
Entry Date: 2012
Accession Number: EJ961581
Database: ERIC
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  Value: <anid>AN0074102842;bsp01may.12;2019Jun26.08:49;v2.2.500</anid> <title id="AN0074102842-1">Temperamental, Parental, and Contextual Contributors to Early-emerging Internalizing Problems: A New Integrative Analysis Approach. </title> <p>This study evaluated a comprehensive model of factors associated with internalizing problems (IP) in early childhood, hypothesizing direct, mediated, and moderated pathways linking child temperamental inhibition, maternal overcontrol and rejection, and contextual stressors to IP. In a novel approach, three samples were integrated to form a large sample (N = 500) of Canadian children (2–6 years; M = 3.95 years; SD =.80). Items tapping into the same constructs across samples were used to create parallel measures of inhibited temperament, maternal positive, critical, and punitive parenting, maternal negative emotionality, family socioeconomic and structural stressors, and child's IP. Multiple‐groups structural equation modeling indicated that associations were invariant across samples and did not differ for boys and girls. Child inhibition, less positive and more critical parenting, maternal negative emotionality, and family socioeconomic disadvantage were found to have direct associations with IP. In addition, maternal negative emotionality was associated with IP through more critical parenting, and both maternal negative emotionality and socioeconomic stress were associated with IP through less positive parenting. Results highlight the multiple independent and cumulative risk factors for early IP and demonstrate the power of integrating data across developmental studies.</p> <p>Keywords: internalizing; temperament; parenting; early childhood</p> <p>Internalizing problems (IP) are a common type of maladjustment in childhood, affecting 14–18 percent of young people (Zahn‐Waxler, Klimes‐Dougan, & Slattery, 2000). Early childhood is a common period of onset, with anxiety problems in particular becoming evident during the transition to childcare or school. Early‐emerging IP are moderately stable from childhood onward (Majcher & Pollack, 1996), and are likely to be detrimental to development (Hammen & Rudolph, 2003), underscoring the importance of early identification and improved understanding of how these problems develop to advance prevention and treatment (Hirshfeld‐Becker & Biederman, 2002). Numerous factors in children and their environment contribute to these problems, including inhibited temperament, parental overcontrol and negativity, and environmental stressors (e.g., Rapee, 2001). How these factors interact to increase the risk of IP and the processes through which these problems develop are not well understood, however, due in part to the lack of research in which multiple factors have been examined simultaneously in early childhood. Such research requires studies in which children are assessed comprehensively on factors both internal and external to the child. The present study addressed this need using a novel cross‐study method in which three independent samples were integrated at early childhood (e.g., Curran & Hussong, 2009).</p> <p>Bioecological models suggest that children's development occurs in nested multilevel contexts, from biological to cultural, with risk factors occurring across levels (e.g., Bronfenbrenner & Ceci, 1994). IP most likely originate from the interactive effects of inhibited temperament reflecting high physiological reactivity attributable to predisposing genetic vulnerabilities, experiences of parental overcontrol and negativity, and exposure to stressful familial, economic, and social disadvantages (Zahn‐Waxler et al., 2000). Children are increasingly likely to develop IP when more of these multilevel risk factors are present and less likely to develop them if risk factors are offset by the presence of protective factors (Ashford, Smit, van Lier, Cuijpers, & Koot, 2008). In the present study, we tested a model postulating pathways to IP from child temperamental inhibition; from parenting practices that are likely to increase children's anxiety, especially if they are highly inhibited and hence more susceptible to anxiety; and from parental negative emotionality and family stressors that are likely to interfere with effective parenting.</p> <p>Twin and family studies have revealed shared genetic risk for temperamental inhibition, anxiety, and depression (e.g., Bolton et al., 2006; Boomsma, van Beijsterveldt, & Hudziak, 2005). Evidence linking physiological reactivity to anxiety problems supports the idea that inhibition may contribute to the development of IP by affecting the ability to cope with anxiety. Some children are physiologically prone to states of anxiety due to over‐activation of brain systems involved in stress responses. At the behavioral level, these brain systems activate regulatory processes designed to lower arousal, evidenced by distress in the face of novelty and avoidance of unfamiliar situations (e.g., Kagan & Snidman, 1999). Distress decreases after withdrawal, reinforcing avoidant coping. However, withdrawal also decreases opportunities to learn or to practice more effective coping, increasing the likelihood of future anxious reactions through a positive feedback loop in which anxious avoidance maintains or increases arousal and the sense of low control. Thus, more inhibited children may show more IP in early childhood.</p> <p>Factors external to the child may contribute also to children's IP. Parents are likely to contribute by responding to the child in ways that reinforce the child's distress. Both inappropriate behavioral control (e.g., overdirective and punitive) and psychological control (e.g., rejection and criticism) might contribute to the development of IP by undermining the child's sense of self‐agency and self‐efficacy and increasing anxiety and distress (e.g., Barber, 1996; Rapee, 2001). Meta‐analyses of research on relations between parenting and childhood IP (McLeod, Weisz, & Wood, 2007; McLeod, Wood, & Weisz, 2007), indicated that child depression was associated with parental rejection and to a lesser extent with parental overcontrol (overinvolvement) whereas anxiety was associated with parental overcontrol and to a lesser extent with parental rejection. A handful of prospective longitudinal studies suggest the same conclusion (e.g., Bayer, Sanson, & Hemphill, 2006; Letcher, Smart, Sanson, & Toumbourou, 2009), although young children's inhibition also contributes reciprocally to these negative parenting dimensions (Rubin, Nelson, Hastings, & Asendorpf, 1999). Studies also point to substantial inverse relations between positive parenting (warm engagement) and both IP and negative parenting (Bayer et al., 2006; Coplan, Arbeau, & Armer, 2008). However, it is unclear whether the presence of negative parenting and the absence of positive parenting are independently associated with children's IP or whether they are at opposite ends of a single continuum. Less positive, more critical, and/or more punitively controlling parents may have children with more IP. Given inhibited children's greater susceptibility to parental influences (e.g., Gallagher, 2002), the links between IP and parenting dimensions may be stronger among more inhibited children.</p> <p>Parental behavior may be the final common pathway through which other risk factors contribute to children's IP. Parents' personal characteristics may be an important risk factor for IP (Rubin & Mills, 1991), that operates through the mechanism of parent behavior. Psychologically vulnerable parents are more likely to resort to ineffective parenting practices during challenging childcare interactions and may direct feelings of frustration or distress toward their children. Emotionally negative parents are more likely to engage in overcontrolling and critical or rejecting parenting (Coplan et al., 2008; Mills et al., 2007). Further, parenting behavior likely mediates relations between parental emotionality and children's IP (Elgar, Mills, Waschbusch, Brownridge, & McGrath, 2007). Thus, parental negative emotionality may be associated with children's IP, and less positive, more critical, and/or more punitive parenting may mediate the association.</p> <p>Parental behavior may also be the pathway through which family risk factors contribute to children's IP. Family socioeconomic disadvantages (e.g., low income and low level of education) and challenging structural characteristics (e.g., single parenthood and large family size) are well‐established family risk factors in children's development (Conger & Dogan, 2007; Weinraub, Horvath, & Gringlas, 2002). Their effects on children have been attributed at least partly to the stresses they put upon parents leading to effects on the quality of parenting (Bronfenbrenner & Ceci, 1994). Economic hardship undermines parents' ability to respond nurturantly to their children (Conger & Conger, 2002). Single‐parent and blended families are at higher risk of parent–child conflict, leading to deterioration in the quality of parenting (e.g., Weinraub et al., 2002). The more children there are in the family, the more parents' financial, psychological, and time resources are likely to be strained (Henderson, Hetherington, Mekos, & Reiss, 1996), and larger family size has been linked to less individual attention to and more differential treatment of children, less positive parenting, and more autocratic discipline (e.g., Jenkins, Rasbash, & O'Connor, 2003). The more young children there are in the family, the greater may be the strain on resources (e.g., Harris, Raley, & Rindfuss, 2002). Thus, contexts of greater family socioecon‐omic stress (lower socioeconomic status) and structural stress (single‐parent family, less than two biological parents, and more young children in the family) may be associated with children's IP, and low positive, more critical, and/or more punitive parenting may mediate these relations.</p> <p>In summary, the present study evaluated a comprehensive, ecological model hypothesizing direct, mediated, and moderated pathways linking child temperamental inhibition, maternal overcontrol and rejection, and contextual stressors to IP in early childhood. We predicted that: (<reflink idref="bib1" id="ref1">1</reflink>) more inhibited children would show more IP in early childhood; (<reflink idref="bib2" id="ref2">2</reflink>) less positive, more critical, and/or more punitively controlling parents would have children with more IP; (<reflink idref="bib3" id="ref3">3</reflink>) the links between IP and parenting dimensions would be stronger among more inhibited children; (<reflink idref="bib4" id="ref4">4</reflink>) parental negative emotionality would be associated with children's IP, and less positive, more critical, and/or more punitive parenting would mediate the association; and (<reflink idref="bib5" id="ref5">5</reflink>) contexts of greater family socioeconomic and structural stress would be associated with children's IP, and low positive, more critical, and/or more punitive parenting would mediate these relations.</p> <p>We evaluated the model using integrative data analysis (e.g., Curran & Hussong, 2009), a new technique in which raw data from two or more existing independent samples are integrated, and models are fit directly to the integrated dataset. The advantages of the technique include increased statistical power, greater heterogeneity in sample demographic characteristics, the ability to test more complex models than would be possible within any single sample, and the potential to replicate findings across the samples. There are threats to parsimony associated with pooling datasets with different sample characteristics, measurements, and response scales. However, the examination of sample differences allows a richer understanding of the generalizability of the results as well as the accuracy of current theoretical notions. The present study is the first to use this approach to evaluate a comprehensive model of the factors associated with IP in early childhood. A large and regionally, socioeconomically, and linguistically diverse sample of young Canadian children was obtained by integrating three samples: a sample at familial and sociodemographic risk (sample 1; Campisi, Serbin, Stack, Ledingham, & Schwartzman, 2009), a sample of children varying in risk for anxiety problems (sample 2; Daycare and Preschool Adjustment Study, e.g., Hastings et al., 2008), and a sample of families recruited from a representative community group (sample 3; Shame in Childhood Study, e.g., Mills, Imm, Walling, & Weiler, 2008). To maximize the size of the integrated sample, assessments of parenting focused on mothers.</p> <hd id="AN0074102842-2">Method</hd> <p></p> <hd id="AN0074102842-3">Participants</hd> <p>Participants were 523 Canadian children ranging from 2.1 to 6.1 years of age (<emph>M</emph>= 3.95 years; <emph>SD</emph>=.80) obtained by integrating three samples. Given that there is no evidence for age differences in risk factors associated with IP in early childhood, the sample comprised all children up to kindergarten age, including seven kindergarten children in sample 1 who turned six before being assessed late in the school year. Twenty‐three children with missing values on most of the key variables were excluded (7 in sample 1 and 16 in sample 3), leaving a final integrated sample of 500. Demographic characteristics are shown in Table 1. On average, children in samples 1 and 3 were 7 months older than those in sample 2, <emph>t</emph>(<reflink idref="bib257" id="ref6">257</reflink>) = 4.82, <emph>p</emph> <.001, and <emph>t</emph>(<reflink idref="bib372" id="ref7">372</reflink>) = 11.23, <emph>p</emph> <.001, respectively; children in samples 1 and 3 did not differ in age, <emph>t</emph>(<reflink idref="bib365" id="ref8">365</reflink>) =.09, NS. Mothers in sample 1 were significantly younger than those in sample 2, <emph>t</emph>(<reflink idref="bib255" id="ref9">255</reflink>) = 8.59, <emph>p</emph> <.001, and in sample 3, <emph>t</emph>(<reflink idref="bib364" id="ref10">364</reflink>) = 8.39, <emph>p</emph> <.001; mothers in samples 2 and 3 did not differ in age, <emph>t</emph>(<reflink idref="bib369" id="ref11">369</reflink>) =.03. Proportionately, more mothers had a low level of education (high school or less) in sample 1 (67 percent) than in sample 2 (12 percent) or sample 3 (16 percent), χ<sups>2</sups>(<reflink idref="bib8" id="ref12">8</reflink>, N = 500) = 213.87, <emph>p</emph> <.001, and proportionately more mothers were in a low income family (below $30 000/year) in sample 1 (40 percent) than in sample 2 (11 percent) or in sample 3 (16 percent), χ<sups>2</sups>(<reflink idref="bib12" id="ref13">12</reflink>, N = 487) = 73.01, <emph>p</emph> <.001. Details on the three samples follow.</p> <p>1 Demographic Characteristics of the Individual and Integrated Samples</p> <p> <ephtml> <table><thead valign="bottom"><tr><th>Characteristics</th><th>Sample 1 (N <italic>=</italic> 126)</th><th>Sample 2 (N <italic>=</italic> 133)</th><th>Sample 3 (N <italic>=</italic> 241)</th><th>Integrated (N <italic>=</italic> 500)</th></tr><tr><th>Percent or <italic>M</italic></th><th>N or <italic>SD</italic></th><th>Percent or <italic>M</italic></th><th>N or <italic>SD</italic></th><th>Percent or <italic>M</italic></th><th>N or <italic>SD</italic></th><th>Percent or <italic>M</italic></th><th>N or <italic>SD</italic></th></tr></thead><tbody valign="top"><tr><td>Child age (years)</td><td>4.11</td><td>1.23</td><td>3.50</td><td>.76</td><td>4.10</td><td>.26</td><td>3.95</td><td>.80</td></tr><tr><td>Child gender</td><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Boys</td><td>48.4%</td><td>61</td><td>45.9%</td><td>61</td><td>54.4%</td><td>131</td><td>50.6%</td><td>253</td></tr><tr><td> Girls</td><td>51.6%</td><td>65</td><td>54.1%</td><td>72</td><td>45.6%</td><td>110</td><td>49.4%</td><td>247</td></tr><tr><td> Maternal age</td><td>30.69</td><td>3.31</td><td>35.32</td><td>5.11</td><td>34.98</td><td>5.62</td><td>34.14</td><td>5.40</td></tr><tr><td>Maternal education</td><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Below high school</td><td>27.0%</td><td>34</td><td>.0%</td><td>0</td><td>.4%</td><td>1</td><td>7.0%</td><td>35</td></tr><tr><td> Completed high school</td><td>39.7%</td><td>50</td><td>12.0%</td><td>16</td><td>15.4%</td><td>37</td><td>20.6%</td><td>103</td></tr><tr><td> Community college</td><td>24.6%</td><td>31</td><td>26.3%</td><td>35</td><td>48.1%</td><td>116</td><td>36.4%</td><td>182</td></tr><tr><td> Undergraduate degree</td><td>7.9%</td><td>10</td><td>42.1%</td><td>56</td><td>32.8%</td><td>79</td><td>29.0%</td><td>145</td></tr><tr><td> Graduate/professional degree</td><td>.8%</td><td>1</td><td>19.5%</td><td>26</td><td>3.3%</td><td>8</td><td>7.0%</td><td>35</td></tr><tr><td>Family income ($)</td><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> 0–10 000</td><td>7.1%</td><td>9</td><td>.0%</td><td>0</td><td>.9%</td><td>2</td><td>2.3%</td><td>11</td></tr><tr><td> 10 001–20 000</td><td>16.7%</td><td>21</td><td>6.2%</td><td>8</td><td>8.6%</td><td>20</td><td>10.1%</td><td>49</td></tr><tr><td> 20 001–30 000</td><td>15.9%</td><td>20</td><td>4.7%</td><td>6</td><td>7.3%</td><td>17</td><td>8.8%</td><td>43</td></tr><tr><td> 30 001–40 000</td><td>10.3%</td><td>13</td><td>9.3%</td><td>12</td><td>10.3%</td><td>24</td><td>10.1%</td><td>49</td></tr><tr><td> 40 001–60 000</td><td>28.6%</td><td>36</td><td>17.8%</td><td>23</td><td>26.3%</td><td>61</td><td>24.6%</td><td>120</td></tr><tr><td> 60 001–74 999</td><td>12.7%</td><td>16</td><td>14.0%</td><td>18</td><td>14.2%</td><td>33</td><td>13.8%</td><td>67</td></tr><tr><td> 75 000 or more</td><td>8.7%</td><td>11</td><td>48.1%</td><td>62</td><td>32.3%</td><td>75</td><td>30.4%</td><td>148</td></tr></tbody></table> </ephtml> </p> <p> <bold>Sample 1 (N <emph>=</emph> 126). </bold> The Concordia Longitudinal Risk Project is a study of long‐term trajectories of childhood social withdrawal and aggression. The original sample included over 1770 elementary‐school Francophone (French‐speaking) children living in predominantly lower income neighborhoods of Montreal in 1976–1978. As the original participants (now in their thirties and forties) became parents, assessments of their children were conducted. Of over 700 representative participants in the most recent phase of the ongoing Concordia project, 126 had children of an appropriate age for the current study and agreed to participate with their child (acceptance rate was approximately 78 percent of eligible families). Participating parents did not differ significantly from either the ongoing longitudinal sample or the original Concordia sample of 1770 in their sociodemographic or behavioral characteristics. The 126 participants involved in the present study consisted of those second‐generation children for whom the target variables were assessed in early childhood and their mothers. Families were 95 percent White and 5 percent other ethnicities (Hispanic and African). Assessments were done in home when the children were 4–6 years old, at which time, mothers completed instruments used in the present study.</p> <p> <bold>Sample 2 (N <emph>=</emph> 133). </bold> The Daycare and Preschool Adjustment Study comprised 133 families in greater metropolitan Montreal. Targeted advertisements were used to recruit children ranging from 'low risk' to 'at risk' for early anxiety problems. By mother report of behavior problems (Achenbach & Rescorla, 2001), 42 children were in the borderline‐clinical to clinical range for IP for their age whereas 48 were less than 1 <emph>SD</emph> above, and 43 were at or below the norm for their age. Children ranged from 2.1 to 4.9 years old. The sample was 70 percent White and 30 percent other ethnicities (Asian, Caribbean, African, Indian, Hispanic, and North African). Half the families were Anglophone (first language is English), one or both parents were Francophone (first language is French) in 37 families, and one or both parents were Allophone (first language is neither English nor French) in 33 families. Mothers completed instruments used for the current study during four assessments over approximately 7 months, involving a telephone interview, a home visit, a mailed questionnaire, and a laboratory visit.</p> <p> <bold>Sample 3 (N <emph>=</emph> 241). </bold> The Shame in Childhood Study is a community sample of 253 English‐speaking from three‐ to four‐year‐old children and their parents. Families with young children born between June 1, 1999 and May 31, 2000 were recruited through a letter of invitation sent to 3500 families drawn randomly from all 6358 families residing in Winnipeg, a mid‐size city (population 600 000) in the geographic center of Canada. Details were provided about the study, and families were invited to return an enclosed stamped return postcard if they had a healthy child and wished to participate. Of 257 families volunteering to participate, four withdrew shortly thereafter. Parents predominantly were White (94 percent), were of European ancestry (74 percent), were at least second‐generation Canadian (78 percent), and reported feeling very Canadian (94 percent rated themselves 8 or higher on a 10‐point scale). Mothers completed instruments used for the current study during an assessment that involved a mailed questionnaire followed shortly by a laboratory visit.</p> <p> <bold>Between‐sample Heterogeneity. </bold> The samples varied considerably in method of recruitment, with the result that they differed in children's risk for IP. Children in sample 1 were at elevated risk of internalizing and/or externalizing problems by virtue of having parents at high risk, children in sample 2 ranged from low to at risk of anxiety based on screening scores, and children in sample 3 were recruited from the community and thus were at low risk of adjustment problems. By increasing the heterogeneity of the pooled sample, these differences increased the potential generalizability of the results, but it was also possible that the higher risk for IP in samples 1 and 2 would contribute disproportionately to the findings. For this reason, in the analyses described below, we examined sample differences before pooling across samples and tested the measurement and structural models with a multiple‐groups approach.</p> <hd id="AN0074102842-4">Measures</hd> <p>A central challenge in integrative data analysis, given that the measures are likely to differ across samples, is to develop appropriate measures of the constructs and to establish measurement invariance across samples. In the present study, in all samples, the same instrument was used to assess IP, and the same information was gathered about family socioeconomic and structural characteristics. In two of the three samples, the same instruments were used to assess child temperament and maternal emotionality. The three samples used different instruments to measure parenting, all of which have been used and validated in previous research. Where the instruments differed across samples, parallel measures were created by selecting items that, although not identical across samples, reflected the same construct and had common scales. Where the items were on different scales (4‐ or 5‐point scale vs. 7‐point scale), values of the smaller scale were converted to those of the larger scale using graduated constants to preserve distributional properties (1 = 1, 2 = 2.5, 3 = 4, 4 = 5.5, 5 = 7; 0 = 1, 1 = 3, 2 = 5, and 3 = 7). The use of rescaled original units was preferable to a transformation (e.g., z‐transformation) because the latter can change distributions of data, and thus, relations among variables can be altered (Cudeck, 1989).</p> <p>For child temperament, parenting, and maternal emotionality, we (five of the coauthors with cumulatively many decades of experience in measurement and assessment of child and parent characteristics) began by identifying candidate parallel items in the instruments used with each sample on the basis of item content. Item and scale statistics were examined in iterative processes to identify sets of relevant items, with the highest possible reliability indices, computed for the integrated sample using data centered within each constituent sample. Centering served to control for between‐group variance by ensuring that variances would range around zero. This method of item selection produced subsets of items with the highest reliability across the samples while accounting for between‐sample differences. Selected items were then rigorously tested with their original sample means included [multiple‐group confirmatory factor analysis (CFA)] to ensure unidimensionality across the samples, as described in the <emph>Results</emph> section. Thus, the parallel measures were sets of items that reflected the same construct as indicated by acceptable reliability in the integrated sample. The instruments and procedures used to create parallel measures across samples are described below. The items selected for each measure are shown in Table 2 (see Table 3 for alpha coefficients in the three samples and the integrated sample).</p> <p>2 Instruments and Items Used to Create Parallel Measures Across Samples</p> <p> <ephtml> <table><thead valign="bottom"><tr><th /><th>Sample 1</th><th>Sample 2</th><th>Sample 3</th></tr></thead><tbody valign="top"><tr><td>Temperament</td><td /><td /><td /></tr><tr><td> Inhibition</td><td>Prefers playing with others R</td><td>Prefers to watch than join</td><td>Prefers to watch than join</td></tr><tr><td>Very sociable R</td><td>At ease with almost anyone R</td><td>At ease with almost anyone R</td></tr><tr><td>Makes friends easily R</td><td>Acts friendly and outgoing R</td><td>Acts friendly and outgoing R</td></tr><tr><td>Very friendly with strangers R</td><td>Joins others quickly R</td><td>Joins others quickly R</td></tr><tr><td>Takes long time to warm up</td><td>Nervous talking to strangers</td><td>Nervous talking to new adults</td></tr><tr><td>Tends to be shy</td><td>Acts shy around new people</td><td>Acts shy around new people</td></tr><tr><td>Parenting</td><td /><td /><td /></tr><tr><td> Positive</td><td>PSI‐expected warmer feelings R</td><td>Have warm intimate times</td><td>Have warm intimate times</td></tr><tr><td>PSI (I feel unliked R/unappreciated R)</td><td>I give comfort and understanding</td><td>I give comfort and understanding</td></tr><tr><td>PDI‐likely to discuss situation</td><td>(Reason with/RCE‐find out why)</td><td>(Give reasons/emphasize reasons)</td></tr><tr><td> Critical</td><td>Insult child or say mean things</td><td>(Tease/RCE‐call silly/RCE‐angry)</td><td>Scold and criticize</td></tr><tr><td>Get on child's back/curse</td><td>Scolding makes child improve</td><td>Scold when falls short</td></tr><tr><td>(Hold grudge/things build up)</td><td>(Ashamed/often angry)</td><td>PCS (silent/unfriendly/disappointed)</td></tr><tr><td> Punitive</td><td>Child can see I'm upset</td><td>When angry, I show it</td><td>Explode in anger</td></tr><tr><td>Speak to my child calmly R</td><td>RCE‐yell at my child</td><td>Yell or shout</td></tr><tr><td>(Spank grab hit/PDI‐physical)</td><td>(Physical/conflict/RCE‐punish)</td><td>(Spank/grab/slap/shove)</td></tr><tr><td>Parent emotionality</td><td /><td /><td /></tr><tr><td> Negative</td><td>Many things annoy me</td><td>Irritable</td><td>Many things annoy me</td></tr><tr><td>When I get scared, I panic</td><td>Calm R</td><td>When I get scared, I panic</td></tr><tr><td>Fewer fears than most R</td><td>Afraid</td><td>Fewer fears than most R</td></tr><tr><td>Get emotionally upset easily</td><td>Upset</td><td>Get emotionally upset easily</td></tr><tr><td>Takes a lot to make me mad R</td><td>Angry</td><td>Takes a lot to make me mad R</td></tr><tr><td>Events make me fretful</td><td>Nervous</td><td>Events make me fretful</td></tr><tr><td>Frequently get distressed</td><td>Distressed</td><td>Frequently get distressed</td></tr><tr><td>Easily frightened</td><td>Frightened</td><td>Easily frightened</td></tr></tbody></table> </ephtml> </p> <p>1 <emph>Notes</emph>: Parenting was assessed in sample 1 mainly by the parenting scale (PS; Arnold, O'Leary, Wolff, & Acker, 1993), in sample 2, mainly by the child rearing practices report (CRPR) Q‐sort (Block, 1981), and in sample 3, mainly by the parenting styles and dimensions questionnaire (PSDQ; Robinson, Mandleco, Olsen, & Hart, 2001). Where a specific item was drawn from one of the following measures, the source is indicated beside the item: RCE = responses to children's emotions (Hastings & De, 2008); PCS = psychological control scale (Olsen et al., 2002); PDI = parenting dimensions inventory (Slater & Power, 1987); PSI = parenting stress index‐short form (Abidin, 1995). Response scales ranged from 1 to 7 for the PS and CRPR, from 1 to 5 for the PSI, RCE, PSDQ, and PCS, and from 0 to 3 for the PDI. Composite items that were created through averaging are enclosed in parentheses, with component items in the composite separated by slashes. R = reverse‐scored. Items with parallel content across the instruments were used to create measures of <emph>Positive Parenting</emph> (in sample 1, 3 of 12 parent–child dysfunctional interaction items from the PSI and all five discuss items from the PDI; in sample 2, 2 of 7 accept items from the CRPR and 2 of 3 reward items from the RCE; in sample 3, 2 of 5 warmth/support and 2 of 5 reasoning/induction items from the PSDQ), <emph>Critical Parenting</emph> (in sample 1, 5 of 10 overreactivity items from the PS; in sample 2, 2 of 5 reject and 2 of 16 authoritarian items from the CRPR and 1 of 3 punish and 1 of 3 magnify items from the RCE; in sample 3, 2 of 4 verbal hostility items from the PSDQ and 3 love withdrawal/guilt induction items from the PCS), and <emph>Punitive Parenting</emph> (in sample 1, 3 of 10 overreactivity items from the PS and all 5 physical items from the PDI; in sample 2, 3 of 16 authoritarian items from the CRPR and 1 of 3 punish items and 1 of 3 magnify items from the RCE; in sample 3, 2 of 4 verbal hostility and all four physical coercion items from the PSDQ).</p> <p>3 Descriptive Statistics and Reliabilities of the Measures by Sample</p> <p> <ephtml> <table><thead valign="bottom"><tr><th>Variables</th><th>Sample 1</th><th>Sample 2</th><th>Sample 3</th><th>Integrated sample</th></tr><tr><th><italic>M</italic> or %</th><th><italic>SD</italic> or N</th><th>α</th><th><italic>M</italic> or %</th><th><italic>SD</italic> or N</th><th>α</th><th><italic>M</italic> or %</th><th><italic>SD</italic> or N</th><th>α</th><th><italic>M</italic> or %</th><th><italic>SD</italic> or N</th><th>α</th></tr></thead><tbody valign="top"><tr><td>Child temperament</td><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Inhibition</td><td>2.92</td><td>1.03</td><td>.75</td><td>3.59</td><td>1.37</td><td>.85</td><td>3.40</td><td>1.37</td><td>.85</td><td>3.33</td><td>1.31</td><td>.83</td></tr><tr><td>Maternal parenting</td><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Positive</td><td>6.18</td><td>1.01</td><td>.66</td><td>6.29</td><td>.66</td><td>.47</td><td>5.93</td><td>.69</td><td>.45</td><td>6.09</td><td>.79</td><td>.54</td></tr><tr><td> Critical</td><td>2.03</td><td>.60</td><td>.43</td><td>1.55</td><td>.65</td><td>.58</td><td>2.39</td><td>.79</td><td>.52</td><td>2.08</td><td>.79</td><td>.51</td></tr><tr><td> Punitive</td><td>3.26</td><td>1.01</td><td>.53</td><td>3.27</td><td>.87</td><td>.54</td><td>2.48</td><td>.77</td><td>.71</td><td>2.89</td><td>.95</td><td>.61</td></tr><tr><td>Maternal emotionality</td><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Negative emotionality</td><td>2.19</td><td>.69</td><td>.79</td><td>2.23</td><td>.70</td><td>.82</td><td>2.33</td><td>.70</td><td>.78</td><td>2.27</td><td>.70</td><td>.77</td></tr><tr><td>Family socioeconomic stress</td><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Mother's education</td><td>2.16</td><td>.94</td><td /><td>3.69</td><td>.92</td><td /><td>3.23</td><td>.76</td><td /><td>3.08</td><td>1.03</td><td /></tr><tr><td> Father's education</td><td>2.14</td><td>.85</td><td /><td>3.58</td><td>.99</td><td /><td>3.10</td><td>.89</td><td /><td>3.00</td><td>1.05</td><td /></tr><tr><td> Occupational prestige</td><td>40.94</td><td>12.09</td><td /><td>54.07</td><td>12.15</td><td /><td>51.93</td><td>11.31</td><td /><td>49.78</td><td>12.79</td><td /></tr><tr><td> Family income</td><td>4.10</td><td>1.75</td><td /><td>5.73</td><td>1.54</td><td /><td>5.25</td><td>1.64</td><td /><td>5.08</td><td>1.75</td><td /></tr><tr><td> Family socioeconomic stress (<italic>z</italic>)</td><td>−.05</td><td>.75</td><td>.67</td><td>−.02</td><td>.75</td><td>.68</td><td>−.01</td><td>.79</td><td>.78</td><td>−.02</td><td>.77</td><td>.74</td></tr><tr><td>Family structural stress</td><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Dual‐ vs. single‐parent family</td><td>(81.0)</td><td>(102)</td><td /><td>(82.7)</td><td>(110)</td><td /><td>(89.6)</td><td>(216)</td><td /><td>(85.6)</td><td>(428)</td><td /></tr><tr><td> Two vs. fewer biological parents</td><td>(73.8)</td><td>(93)</td><td /><td>(78.2)</td><td>(104)</td><td /><td>(83.4)</td><td>(201)</td><td /><td>(79.6)</td><td>(398)</td><td /></tr><tr><td> Number of children under the age of seven</td><td>1.65</td><td>.65</td><td /><td>1.60</td><td>.65</td><td /><td>1.97</td><td>.73</td><td /><td>1.79</td><td>.71</td><td /></tr><tr><td>Child internalizing problems</td><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> Internalizing problemsa</td><td>52.40</td><td>8.49</td><td>.85/.70</td><td>54.47</td><td>10.21</td><td>.85</td><td>51.11</td><td>9.15</td><td>.81</td><td>52.33</td><td>9.37</td><td /></tr></tbody></table> </ephtml> </p> <ulist> <item>2 <emph>Notes</emph>: For the integrated sample, Cronbach's alpha reliabilities were computed using centered data to control for between‐group variance.</item> <item>3 a Means and standard deviations are for the <emph>T</emph> scores; sample 1 alphas are for the 2–3 and 4–8 versions, respectively, of the child behavior checklist.</item> </ulist> <p> <bold>Child Temperament. </bold> To assess child temperament, parents completed the children's behavior questionnaire (CBQ; Rothbart, Ahadi, Hershey, & Fisher, 2001), in samples 2 and 3, and the emotionality, activity, and sociability (EAS) temperament survey for children (Buss & Plomin, 1984), in sample 1. For the CBQ, parents rate descriptions of children's reactions to situations on a 7‐point scale from 1 (<emph>extremely untrue</emph>) to 7 (<emph>extremely true</emph>). For the EAS, parents rate items using a 5‐point scale from 1 (<emph>not characteristic or typical</emph>) to 5 (<emph>very characteristic or typical</emph>). Six items equivalent in meaning across instruments (6 of 12 shyness items from the CBQ; all 5 shyness and 1 of 5 sociability items from the EAS) were used to create a parallel measure of <emph>inhibited temperament</emph>[e.g., prefers to watch than join, at ease with almost anyone (reversed), acts shy around new people]. EAS items were rescaled converting 5‐point to 7‐point scales.</p> <p> <bold>Parenting. </bold> Mothers' parenting styles and practices were assessed using the instruments and items shown in Table 2. Mothers rated the likelihood or frequency of using a parenting strategy or indicated how descriptive the item was of their parenting practices. To increase similarity, some items within an instrument for one sample were averaged to form a single composite item that cohered more than any of its constituent items with the items on the instruments for the other samples. Individual and composite items were used to create three‐item parallel measures of <emph>positive</emph>, <emph>critical</emph>, and <emph>punitive</emph> parenting. Items originally scored on 4‐ or 5‐point scales were rescaled to 7‐point scales. Parenting scores were computed as the mean of item ratings. The relatively low alphas (see Table 3) are likely due to the small number of items in each measure necessitated by rigorous item selection. The composites reflect parcels, and they are very beneficial empirically. Their use allows for the inclusion of information from multiple items without requiring extra parameters within the model (Bagozzi & Edwards, 1998), a greater ratio of common‐to‐unique variance (Little, Cunningham, Shahar, & Widaman, 2002), and are most likely to capture the true relations among constructs of interest (Rushton, Brainerd, & Pressley, 1983).</p> <p> <bold>Maternal Negative Emotionality. </bold> Two self‐report scales were used to assess mothers' emotionality: the EAS temperament survey for adults (Buss & Plomin, 1984) in samples 1 and 3, and the positive and negative affect scales (PANAS; Watson, Clark, & Tellegen, 1988) in sample 2. The EAS for adults required mothers to rate items describing emotions and behaviors on scales ranging from 1 (<emph>not characteristic or typical of you</emph>) to 5 (<emph>very characteristic or typical of you</emph>). The PANAS included a list of affect terms (e.g., bashful, nervous, and happy) that parents rated as describing their own feelings in recent weeks on scales from 1 (<emph>not at all</emph>) to 5 (<emph>extremely</emph>). Eight items with parallel content across the two instruments (3 of 4 distress, 3 of 4 fear, and 2 of 4 anger items from the EAS; eight matching emotion words from the PANAS) were used to create a measure of <emph>negative emotionality</emph> (e.g., many things annoy me and get emotionally upset easily).</p> <p> <bold>Family Socioeconomic Stress. </bold> Family stress associated with socioeconomic status was assessed from four components: mother's education, father's education, highest occupational prestige in the family, and family income. Occupational prestige was scored using a recently updated version of the standard international occupational prestige scale (SIOPS; Treiman, 1977). The SIOPS was originally constructed with reference to the International Standard Classification of Occupation (ISCO68), an international measure of occupational codes and prestige criteria, to obtain a cross‐national comparative measure of occupational status. The original SIOPS has been updated to refer to the ISCO88 (Hakim, 1998). The occupations listed in the SIOPS range from the highest scores given to professionals (e.g., lawyers, physicians, and chief executive officers) to the lowest given to those who hold such occupations as domestic laborers, manufacturers, and farmhands.</p> <p>In all three samples, mothers were asked to report the highest levels of education completed by themselves and their children's fathers (in two‐parent households). Education levels were converted into 5‐point scales ranging from 1 (<emph>did not complete high school</emph>) to 5 (<emph>attained graduate or professional degree</emph>, e.g., MD and PhD). Mothers were also asked to report the annual total family income before taxes. Family income was converted into a 7‐point scale ranging from 1 (<emph>$0–$10 000</emph>) to 7 (<emph>$75 000 or more</emph>). A principal components analysis indicated that a single‐factor solution was appropriate for the four component measures (mother's education, father's education, highest SIOPS score in the family, and family income), eigenvalue = 2.23, variance accounted for = 56 percent. An overall measure of <emph>family socioeconomic stress</emph> for each sample was created by averaging the four measures after transforming each of them into a standard score (<emph>z</emph>) and reversing each scale so that higher values would reflect greater stress.</p> <p> <bold>Family Structural Stress. </bold> Family stress associated with structural characteristics was assessed by three indices: dual‐ vs. single‐parent family, two vs. fewer biological parents, and number of children under the age of seven. Preliminary analyses indicated that number of children under the age of seven did not cohere with either of the other two indices; therefore, rather than form a single index of structural stress, the three variables were examined separately in the model.</p> <p> <bold>Child IP. </bold> IP were assessed by two versions of the Achenbach child behavior checklist, the child behavior checklist (CBCL) and the Achenbach system of empirically based assessment (ASEBA CBCL) using the age 2–3 and 4–18 versions in samples 2 and 3 (ASEBA CBCL; Achenbach & Rescorla, 2001). The CBCL is the most widely used tool for assessing IP. It has good reliability and validity. Across CBCL versions, the IP scale comprised more than 30 items rated from 0 (<emph>not true as far as you know</emph>) to 2 (<emph>very true or often true</emph>). Total scores were obtained by summing mothers' ratings of the items; to account for changes in measurement over time, <emph>T</emph> scores derived from population norms were used. To check whether the use of <emph>T</emph> scores would obscure gender differences, we performed analyses using mean item scores. No significant gender differences within or across samples were found.</p> <hd id="AN0074102842-5">Results</hd> <p></p> <hd id="AN0074102842-6">Preliminary Analyses</hd> <p>An examination of missing values revealed that for the integrated sample about 2 percent of data had missing values with the exception of one variable, maternal negative emotionality, which had 9 percent missing values due to a delay in adding it to the sample 3 protocol. Given the moderate amount of missing data for maternal negative emotionality, it was decided to perform the main analyses using an imputation approach (Tabachnick & Fidell, 2007). Before imputation, 58 of the 500 cases had one or two missing values, and these data were assumed to be missing at random. To investigate this assumption, we used expectation maximization (EM) imputation with the complete dataset, as recommended by Widaman (2006) for small amounts of missingness. The imputation method uses all available data to estimate a new dataset with imputed values for those that originally had missing values. Next, the model of interest is estimated with the newly imputed dataset. Maximum likelihood (ML) estimation simply uses all available data from the original dataset to estimate the model of interest (Allison, 2002). We compared the results using EM with those obtained from ML, and we found that the parameter estimates under the two approaches were almost identical. The similarity of the results suggested that the more rigorous treatment of the missingness via EM was not necessary. Therefore, we used ML to estimate the final model and assumed that the missing values were missing at random.</p> <p>For the most part, variables were normally distributed. Although some variables showed skewness or kurtosis, multivariate normality was not substantially violated and did not affect the analysis. However, one case in sample 2 was excluded from the structural analysis owing to an unusual impact on multivariate kurtosis. Therefore, N = 499 in the final, reported model.</p> <hd id="AN0074102842-7">Test of the Measurement Model</hd> <p>To assess construct validity of the measures, a CFA model was fitted to each sample (Anderson & Gerbing, 1988). This measurement model constructs the latent variables from their corresponding manifest variable indicators and then correlates all of the latent variables. Good fit of the measurement model within each sample indicates a good construction of latent variables for each sample (Anderson & Gerbing, 1988). We evaluated model fit with the robust method (Byrne, 2006) using the comparative fit index (CFI) and root mean square error of approximation (RMSEA), which is not overly sensitive to sample size (Bentler, 2004). By convention, CFI values of.95 or above and RMSEA values of.06 or below are indicative of a good empirical fit (Tabachnick & Fidell, 2007). The analyses yielded CFI values of.969,.984, and.924 and RMSEA values of.023,.019, and.040 for samples 1, 2, and 3, respectively. These indices were within acceptable ranges, indicating reasonably good fit and providing evidence for factorial validity of the measures.</p> <p>Next, we performed a multiple‐group CFA model with the integrated sample to evaluate the extent to which the measures functioned similarly in relation to their underlying constructs across samples. Typically, multi‐group comparisons require constraints among the factor loadings as well as the correlations among the latent variables (Meredith, 1993). However, in the present context, in which the latent variables for the three samples were derived from different sets of items, invariance of the factor loadings would not be expected and was not required (Bentler, 2004; Byrne, Shavelson, & Muthén, 1989). Because the latent variables were the constructs of interest, we were concerned with invariance of the correlations among the latent variables but not with invariance of the factor loadings. If the correlation pattern for the latent variables is invariant across the samples, then the latent variables can be assumed to represent the same constructs for each of the three samples (Nesselroade, Gerstorf, Hardy, & Ram, 2007). It should be noted that two of the latent variable correlations were found to be problematic. Firstly, a significant difference was found for the correlation between parent emotionality and punitive parenting for samples 2 and 3 (χ<sups>2</sups>= 5.03, <emph>df</emph>= 1, <emph>p</emph>=.025). Secondly, a marginally significant difference was found for the correlation between family socioeconomic stress and critical parenting for samples 2 and 3 (χ<sups>2</sups>= 3.661, <emph>df</emph>= 1, <emph>p</emph>=.056). However, given the large sample size, there was a high amount of power for detecting differences among these correlations. Thus, the fit of the measurement model was used to assess the invariance of the correlations among the latent variables. The model with the constraints across all correlations yielded a CFI of.952 and a RMSEA of.031 across the three samples (χ<sups>2</sups>= 55.09, <emph>df</emph>= 51, <emph>p</emph>=.32), indicating a very good fit of the model. The fit of the model with all correlations freely estimated across the samples was very poor (χ<sups>2</sups>= 529.904, <emph>df</emph>= 129, <emph>p</emph> <.0001). The fit for the restricted model was much better than the fit of the free model (Δχ<sups>2</sups>= 507.814, <emph>df</emph>= 78, <emph>p</emph> <.0001). This finding suggests that it is reasonable to assume that the differences in the two problematic correlations are negligible and that the indicators within all samples are modeling the same latent variables (Nesselroade et al., 2007). Furthermore, when correlations were constrained to be direct effects in the hypothesized model, no significant differences were detected across the samples.</p> <hd id="AN0074102842-8">Preparation for the Structural Model</hd> <p>Given that each set of manifest variables loaded onto one latent factor, and these latent factors had a similar correlation pattern across samples, we next focused our attention on the reliability of the manifest variables across the samples. Because each set of variables loaded onto a single factor, we created a composite score for each set of variables. Simple averaging was used to create the composite score rather than a weighted score, which would be sample‐dependent (e.g., Cohen, 1990; Wainer, 1976). The use of single composite indicators for the latent variables within the measurement model allowed for a simple method of accounting for unreliability in the set of indicators (Kline, 2005). In this method, the uniqueness is defined as the total variability of the manifest variable X (or σ<subs>x</subs><sups>2</sups>) multiplied by the proportion of unreliability of the manifest variables (1 − r<subs>xx</subs>) for the particular construct of interest (where r<subs>xx</subs> represents Cronbach's alpha for variable X). Using these constraints, ξ becomes the remaining proportion of variability in X, or the part of the variability in X that is common across all the manifest variables.</p> <p>In essence, this approach extracts the part variability in the set of items that is estimated to be reliable and models only this reliable portion of the variance at the structural level. To incorporate this within our model, we defined the residual variance for each of the composites as the proportion of variance in the composite that was estimated to be unreliable (one minus Cronbach's alpha). X can be thought of as one particular composite within our model, ε represents the proportion of the variability of the manifest variables that is estimated to be unreliable, and ξ represents the proportion that is reliable. Therefore, the unreliability of each set of manifest variables from each sample can be accounted for in the measurement model. Because only the reliable part in each sample will go to the structural level, we know that the low reliabilities of the sets of manifest variables will not adversely affect the robustness of the model.</p> <p>Cronbach's alpha was used to estimate the reliability for each set of manifest variables (Bedeian, Day, & Kelloway, 1997). Cronbach's alpha was chosen because it can be used to estimate the reliability of a composite score for a set of variables (McDonald, 1999). Once Cronbach's alpha had been calculated for each set of manifest variables, the composites that summarized each set were decomposed within the measurement model into common and uncommon variance as described above. This decomposition ensured that relations at the structural level were based on reliable variance rather than random noise within the data.</p> <p>In summary, our approach to estimating the measurement model allowed different levels of reliability across each of the samples, as well as a direct specification of reliability based on Cronbach's alpha. Given that the sets of manifest variables were each unidimensional (good fit of a one‐factor CFA for each set of manifest variables), we knew that each set could be summarized via a composite score. To account for unreliability of the composite score, we extracted the proportion of variance that was estimated to be reliable and used only that portion at the structural level. This method ensures that the relations at the structural level reflect only reliable measurements of the manifest variables rather than a mixture of reliable and unreliable portions of variability.</p> <hd id="AN0074102842-9">Sample Differences</hd> <p>Prior to investigating the structural model, sample differences, as well as correlations among the measured variables, computed as composite indicators, were investigated. Descriptive statistics and coefficient alpha reliabilities for the variables are shown in Table 3. To examine mean differences among the three samples, a series of one‐way analyses of variance was performed; with Bonferroni's correction applied to follow‐up comparisons, only sample differences at <emph>p</emph> <.001 are reported. The analyses revealed numerous differences among the samples. Children in sample 1 scored lower on <emph>temperamental inhibition</emph> than those in samples 2 and 3, <emph>t</emph>(<reflink idref="bib253" id="ref14">253</reflink>) =−4.37, <emph>t</emph>(<reflink idref="bib362" id="ref15">362</reflink>) =−3.40, respectively; the latter two samples did not differ significantly. Mothers in sample 3 rated themselves as less <emph>punitive</emph> than those in samples 1 and 2, <emph>t</emph>(<reflink idref="bib365" id="ref16">365</reflink>) =−8.27, <emph>t</emph>(<reflink idref="bib372" id="ref17">372</reflink>) =−9.06, respectively; the latter two samples did not differ significantly on punitive parenting. Mothers in sample 3 rated themselves more <emph>critical</emph> than those in sample 1, <emph>t</emph>(<reflink idref="bib365" id="ref18">365</reflink>) = 4.37, who in turn rated themselves more critical than those in sample 2, <emph>t</emph>(<reflink idref="bib257" id="ref19">257</reflink>) = 6.19. Mothers in sample 2 rated themselves more <emph>positive</emph> than those in sample 3, <emph>t</emph>(<reflink idref="bib372" id="ref20">372</reflink>) = 4.90, but not those in sample 1, <emph>t</emph>(<reflink idref="bib257" id="ref21">257</reflink>) = 1.01, NS; there was no significant difference between mothers in samples 1 and 3, <emph>t</emph>(<reflink idref="bib365" id="ref22">365</reflink>) = 3.02, <emph>p</emph>=.003 (below the Bonferroni criterion). <emph>Family socioeconomic stress</emph> was higher in sample 1 than in sample 3, as indicated by lower scores on all four components, smallest <emph>t</emph>(<reflink idref="bib356" id="ref23">356</reflink>) = 6.19, and higher in sample 3 than in sample 2 on mother's and father's education, smaller <emph>t</emph>(<reflink idref="bib344" id="ref24">344</reflink>) = 4.60. <emph>Family structural stress</emph> did not differ significantly across samples except with respect to the number of children under the age of seven, which was higher in sample 3 than in samples 1 and 2, <emph>t</emph>(<reflink idref="bib365" id="ref25">365</reflink>) = 4.13, <emph>t</emph>(<reflink idref="bib372" id="ref26">372</reflink>) = 4.85, respectively, which did not differ significantly. Finally, mothers in sample 2 reported more IP in their children than did mothers in sample 3, <emph>t</emph>(<reflink idref="bib371" id="ref27">371</reflink>) = 3.26; sample 1 did not differ significantly from samples 2 or 3. In summary, the samples differed on several of the measured variables. Levels of IP were highest in sample 2 and lowest in sample 3, possibly due to sample differences in the recruitment of children at risk for IP. Sample 2 mothers showed more positive and less critical parenting than mothers in the other samples, possibly related to lower family stress in sample 2.</p> <hd id="AN0074102842-10">Correlations among Variables</hd> <p>Zero‐order correlations among the composite variables are summarized in Table 4 for the integrated sample. Due to the age range of the sample and research indicating that child age is negatively associated with positive parenting (Brody, Stoneman, & Burke, 1987; Stevenson, Leavitt, Thompson, & Roach, 1988), child age was included to determine whether it should be considered as a covariate in the analysis. As shown in Table 4, children with more IP were significantly more inhibited, <emph>r</emph>=.27, <emph>p</emph> <.001. They had mothers whose parenting was less positive, <emph>r</emph>=−.13, <emph>p</emph> <.01, more critical, <emph>r</emph>=.16, <emph>p</emph> <.001, and more punitive, <emph>r</emph>=.14, <emph>p</emph> <.01, and whose emotionality was more negative, <emph>r</emph>=.25, <emph>p</emph> <.001; and they lived in families under greater socioeconomic stress, <emph>r</emph>=.09, <emph>p</emph> <.05, with fewer than two parents, <emph>r</emph><subs>s</subs>=−.10, <emph>p</emph> <.05, or with fewer than two biological parents, <emph>r</emph><subs>s</subs>=−.11, <emph>p</emph> <.01. Child inhibition was not related to maternal parenting or negative emotionality. Mothers who reported more negative emotionality used parenting practices that were less positive, <emph>r</emph>=−.20, and more critical, <emph>r</emph>=.23, or more punitive, <emph>r</emph>=.33, all <emph>p</emph> <.001. Mothers experiencing more socioeconomic stress were less positive, <emph>r</emph>=−.09, or more punitive, <emph>r</emph>=.11, <emph>p</emph> <.05. Positive parenting was negatively related to both critical parenting, <emph>r</emph>=−.10, <emph>p</emph> <.05, and punitive parenting, <emph>r</emph>=−.28, <emph>p</emph> <.001, which were positively related to each other, <emph>r</emph>=.33, <emph>p</emph> <.001. Child age was not substantially related to other variables.</p> <p>4 Zero‐order Correlations among Variables in the Integrated Sample (N = 500)</p> <p> <ephtml> <table><thead valign="bottom"><tr><th /><th>1.</th><th>2.</th><th>3.</th><th>4.</th><th>5.</th><th>6.</th><th>7.</th><th>8.</th><th>9.</th><th>10.</th></tr></thead><tbody valign="top"><tr><td> 1. Child age</td><td /><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> 2. Inhibited</td><td>−.02</td><td /><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> 3. Positive</td><td>−.07</td><td>−.01</td><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> 4. Critical</td><td>.02</td><td>.01</td><td>−.10*</td><td /><td /><td /><td /><td /><td /><td /></tr><tr><td> 5. Punitive</td><td>.04</td><td>−.05</td><td>−.28***</td><td>.34***</td><td /><td /><td /><td /><td /><td /></tr><tr><td> 6. Negative</td><td>−.02</td><td>−.01</td><td>−.20***</td><td>.23***</td><td>.33***</td><td /><td /><td /><td /><td /></tr><tr><td> 7. Family socioeconomic stress</td><td>.00</td><td>−.11*</td><td>−.09*</td><td>−.03</td><td>.11*</td><td>.05</td><td /><td /><td /><td /></tr><tr><td> 8. Dual‐ vs. single‐parent familya</td><td>.00</td><td>.06</td><td>.07</td><td>.07</td><td>−.04</td><td>−.03</td><td>−.33***</td><td /><td /><td /></tr><tr><td> 9. Two vs. fewer biological parentsa</td><td>−.03</td><td>.05</td><td>.08</td><td>.01</td><td>−.03</td><td>−.01</td><td>−.36***</td><td>.81***</td><td /><td /></tr><tr><td>10. Number of children under the age of seven</td><td>.04</td><td>.06</td><td>−.08</td><td>.08</td><td>.12**</td><td>−.02</td><td>−.13**</td><td>.14***</td><td>.16***</td><td /></tr><tr><td>11. Internalizing problems</td><td>.03</td><td>.27***</td><td>−.13**</td><td>.16***</td><td>.14**</td><td>.25***</td><td>.09*</td><td>−.10*</td><td>−.11**</td><td>−.02</td></tr></tbody></table> </ephtml> </p> <ulist> <item>4 <emph>Notes</emph>: <sups>a</sups> Associations with these variables were examined using Spearman rank‐order correlations.</item> <item>5 * <emph>p</emph> <.05, ** <emph>p</emph> <.01, *** <emph>p</emph> <.001.</item> </ulist> <hd id="AN0074102842-11">Test of the Hypothesized Model</hd> <p>The model postulated pathways to IP from inhibition; from parenting practices that are likely to increase children's difficulty regulating anxiety reactions, especially if they are highly inhibited and hence more prone to anxiety reactions; from negative emotionality in the parent that interferes with effective parenting; and from family socioeconomic disadvantage and structural stressors (single‐parent family, fewer than two biological parents, and having more young children) that are likely to interfere with effective parenting. The model was tested with EQS 6.1 (Multivariate Software, Encino, CA, USA). To avoid potential bias caused by violations of normality in assessing model adequacy, we used maximum‐likelihood estimation with and without the robust option. The robust option corrects for violations of the assumption of normality (Chou, Bentler, & Satorra, 1991). The two methods yielded similar results. We evaluated model fit with the robust method using model chi‐square, CFI, and RMSEA.</p> <p>We tested the model using path analysis with a multiple‐groups approach that evaluated the equality of the model structure across samples. Although child age did not have significant zero‐order correlations with other variables, given its known association with less positive parenting, it was examined as a covariate in the model and retained due to its relations with parenting. Firstly, the model was tested in each of the three samples without any constraints on the parameters (sample 1: χ<sups>2</sups>(<reflink idref="bib48" id="ref28">48</reflink>) = 63.9, <emph>p</emph>=.06, CFI =.969, RMSEA =.023; sample 2: χ<sups>2</sups>(<reflink idref="bib48" id="ref29">48</reflink>) = 61.04, <emph>p</emph>=.10, CFI =.984, RMSEA =.019; and sample 3: χ<sups>2</sups>(<reflink idref="bib88" id="ref30">88</reflink>) = 131.82, <emph>p</emph>=.002, CFI =.924, RMSEA =.04). These models included the specifications for the residual variance as described earlier (Kline, 2005). Because all CFI values were larger than.95 and all RMSEA values were smaller than.05, these models were judged to be very good. Next, we constrained the path coefficients to be equal across the three groups. This model showed equally good fit, χ<sups>2</sups>(<reflink idref="bib51" id="ref31">51</reflink>) = 55.09, <emph>p</emph>=.32, CFI =.991, RMSEA =.022. The hypothesis that associations between IP and the three parenting indices would be stronger for more inhibited children was tested by including parenting × temperament interaction terms in the model. All were nonsignificant, indicating no moderation effects by inhibition, <emph>b</emph>=.47,.11, and 1.23 for the interaction with positive, critical, and punitive parenting, respectively, all <emph>p</emph> >.40. Fit indices for the non‐constrained model were χ<sups>2</sups>(<reflink idref="bib96" id="ref32">96</reflink>) = 192.92, <emph>p</emph> <.01, CFI =.96, RMSEA =.08, and those for the constrained model (all interaction path coefficients constrained to zero) were χ<sups>2</sups>(<reflink idref="bib105" id="ref33">105</reflink>) = 180.54, <emph>p</emph> <.01, CFI =.96, RMSEA =.10. No significant difference in model fit could be detected between the constrained and non‐constrained models (Δχ<sups>2</sups>(<reflink idref="bib9" id="ref34">9</reflink>) = 12.37, <emph>p</emph>=.19). These results suggest that effects of parenting on IP were not moderated by temperamental inhibition; therefore, these moderating effects were excluded from the final model. To examine reciprocal relations between critical parenting and IP and between positive parenting and IP, we performed two separate analyses including the inverse instead of the hypothesized path in the model. The coefficient for the reverse path from IP to critical parenting was nonsignificant [<emph>b</emph>=.01, <emph>p</emph>=.15; χ<sups>2</sups>(<reflink idref="bib38" id="ref35">38</reflink>) = 46.14, <emph>p</emph>=.17, CFI =.98, RMSEA =.03], and the coefficient for the path from IP to positive parenting was also nonsignificant [<emph>b</emph>=−.01, <emph>p</emph>=.10; χ<sups>2</sups>(<reflink idref="bib38" id="ref36">38</reflink>) = 45.17, <emph>p</emph>=.20, CFI =.98, RMSEA =.03]. On the basis of these findings, reciprocal relations between parenting dimensions and IP were excluded from the final model. Finally, additional multiple‐group analyses revealed no path coefficient differences as a function of dual‐ vs. single‐parent family [Δχ<sups>2</sups>(<reflink idref="bib18" id="ref37">18</reflink>) = 15.24, <emph>p</emph>=.65], two vs. fewer biological parents [Δχ<sups>2</sups>(<reflink idref="bib18" id="ref38">18</reflink>) = 17.97, <emph>p</emph>=.46] or child gender [Δχ<sups>2</sups>(<reflink idref="bib18" id="ref39">18</reflink>) = 16.14, <emph>p</emph>=.58]. For each of these three analyses, constraints were placed on all of the path coefficients across the two specified groups. The large <emph>p</emph> values (i.e., >.05) indicate no significant difference while constraining these paths across the groups; therefore, there was no detectable difference among the relationships of the latent variables as a function of group membership. Fit statistics for the non‐constrained models were χ<sups>2</sups>(<reflink idref="bib10" id="ref40">10</reflink>) = 6.64, <emph>p</emph>=.78, CFI = 1.0 and RMSEA =.001 for dual‐ vs. single‐parent family; χ<sups>2</sups>(<reflink idref="bib10" id="ref41">10</reflink>) = 12.04, <emph>p</emph>=.36, CFI =.99 and RMSEA =.035 for two vs. fewer biological parents; and χ<sups>2</sups>(<reflink idref="bib10" id="ref42">10</reflink>) = 11.78, <emph>p</emph>=.78, CFI =.99 and RMSEA =.03 for boys vs. girls. Therefore, these variables were excluded from the final model.</p> <p>The results of the final model are shown in Figure 1. Independent variables in the model were allowed to freely covary. Covariances for each pair of independent variables were estimated to measure the level of dependence not accounted for by the direct effects in the model. The covariances for each sample are not provided in Figure 1 because they differ across samples. Because multiple‐groups analysis constrains the unstandardized coefficients to be equal across groups, it does not yield a common set of standardized coefficients (Kline, 2005). For this reason, the unstandardized solution is shown in the figure, which provides coefficients for the significant paths. Coefficients reflect, for each unit increase in a predictor, the expected unit increase or decrease in the predicted variable, given that the other parameters remain unchanged. Note that the unstandardized coefficients for prediction of parenting variables are smaller in magnitude than those for prediction of IP due to differences in the measurement scales for these variables (1–7 for parenting and 29–76 for IP).</p> <p>Graph: 1 Unstandardized Path Coefficients Depicting Direct and Indirect pathways Linking Family, Parent, and Child Variables to Internalizing Problems in Early Childhood. Notes: Significant paths are represented by solid lines and nonsignificant paths by dashed lines. * p <.05, ** p <.01, *** p <.001.</p> <p>As Figure 1 shows, most hypothesized relations were significant. Unstandardized coefficients indicate strong direct predictive relations between IP and child inhibition (<emph>b</emph>= 2.64, <emph>p</emph> <.01), maternal low positive (<emph>b</emph>=−1.98, <emph>p</emph>=.02) and high critical (<emph>b</emph>= 3.20, <emph>p</emph>=.02) parenting, maternal negative emotionality (<emph>b</emph>= 3.22, <emph>p</emph> <.01), and family socioeconomic stress (<emph>b</emph>= 2.25, <emph>p</emph> <.01). Maternal punitive parenting was strongly predicted by maternal negative emotionality (<emph>b</emph>=.50, <emph>p</emph> <.01) and family socioeconomic (<emph>b</emph>=.19, <emph>p</emph> <.01) and structural (children under the age of seven) (<emph>b</emph>=.17, <emph>p</emph> <.01) stress; however, there was no association between children's IP and punitive parenting over and above the associations with less positive and more critical parenting. There were pathways to IP from negative emotionality and family structural stress (having more children under the age of 7) via less positive parenting (<emph>b</emph>=−.29, <emph>p</emph> <.01, and <emph>b</emph>=−.12, <emph>p</emph> <.01, respectively) and more critical parenting (<emph>b</emph>=.31, <emph>p</emph> <.01, and <emph>b</emph>=.09, <emph>p</emph>=.02, respectively), and from socioeconomic stress via less positive parenting (<emph>b</emph>=−.12, <emph>p</emph> <.01).</p> <p> <bold>Mediation by Parenting. </bold> The hypothesis that maternal negative emotionality and family stress would be associated with IP via their impact on parenting was tested by examining indirect effects estimated as the product of the path coefficients linking the independent variable to the mediator and linking the mediator to the dependent variable. Indirect effects are shown in Figure 1 by the significant paths to IP from maternal negative emotionality and from family structural stress (having more children under the age of seven) through low positive and high critical parenting, and from family socioeconomic stress through low positive parenting. To test the significance of mediation, we used bootstrapping procedures (MacKinnon, Lockwood, Hoffman, West, & Sheets, 2002; Shrout & Bolger, 2002), which involve taking multiple random samples of observations from the original sample, with replacement, and analyzing them to provide a distribution of parameter estimates. Using EQS, we created bootstrap samples and computed confidence intervals (CI) to assess the significance of the indirect effects. Distributions of parameter estimates tend to be skewed (Bollen & Stine, 1990; Lockwood & MacKinnon, 1998; MacKinnon et al., 2002); therefore, bias‐corrected confidence intervals were computed (Efron & Tibshirani, 1993; MacKinnon, Lockwood, & Williams, 2004). Maternal negative emotionality was associated with IP through high critical parenting (95 percent CI:.121–3.308, <emph>p</emph>=.03) and through low positive parenting (90 percent CI:.029–1.375, <emph>p</emph>=.08). Family socioeconomic stress was associated with IP through low positive parenting (95 percent CI:.007–.909, <emph>p</emph>=.045).</p> <hd id="AN0074102842-12">Discussion</hd> <p>The purpose of this study was to test an ecological model of the development of early‐emerging IP, postulating pathways to IP from temperamental inhibition, from maternal parenting practices that are likely to increase children's difficulty regulating anxiety reactions, from maternal negative emotionality that interferes with effective parenting, and from family socioeconomic and structural sources of stress that are likely to interfere with effective parenting. Using a form of integrative data analysis (e.g., Curran & Hussong, 2009), three independent samples were integrated to create a large and heterogeneous sample of children assessed on these multilevel factors. Structural analysis performed with a multiple‐groups approach revealed that associations were invariant across samples. Child inhibition, less positive and more critical parenting, maternal negative emotionality, and family socioeconomic disadvantage were found to have direct associations with IP. In addition, there was some evidence for mediation by parenting. Maternal negative emotionality was associated with IP through more critical parenting, and both negative emotionality and socioeconomic stress were associated with IP through less positive parenting. Reciprocal relations between IP and parenting dimensions were nonsignificant. These findings provide the first evidence in a large and heterogeneous sample that early‐emerging IP may be understood from an ecological perspective as involving multilevel factors and that the impact of more distal factors operates through effects on parenting.</p> <p>In keeping with the model, most of the key factors hypothesized to contribute to IP were found to be directly associated with IP. Each of these multilevel factors may increase the risk of IP in early childhood, and they may do so in an additive fashion. The finding of a direct link between inhibition and IP is consistent with theory and research identifying inhibited temperament as a vulnerability factor in the development of IP (e.g., Rapee, 2001). The finding that inhibition was relatively independent of the parenting and family contextual factors that were associated with IP could be interpreted as evidence for the dispositional or genetic basis of this temperament trait (Biederman et al., 1990; Rosenbaum et al., 1993), and for the notion that inhibited temperament and parenting may not be strongly related in the absence of cumulative risk (e.g., Paulussen‐Hoogeboom, Stams, Hermanns, & Peetsma, 2007). Temperamental inhibition is thought to reflect the behavioral manifestation of underlying biological difficulties with emotion regulation (Rubin, Coplan, Fox, & Calkins, 1995). The development of emotion regulation is fundamental to young children's ability to manage their stress reactions. Inhibited children's greater physiological reactivity to novel and social stressors and resultant avoidant coping strategies leave them prone to heightened anxiety and IP. This risk may be independent of any effect inhibited children's reactivity has on parenting, at least in the context of low cumulative risk.</p> <p>Also consistent with current theory and research (Coplan et al., 2008; Hastings, Nuselovici, Rubin, & Cheah, 2010), early IP were directly associated with less positive and more critical maternal parenting. During early childhood, children lacking external support to develop emotion regulation skills due to low parental warmth or engagement or high parental rejection or negativity will be more prone to states of anxiety and less able to manage these states. Past research has linked punitive parenting to IP (McLeod, Wood, et al., 2007), and although such was the case in the present study when considering their zero‐order correlation, the relation was not significant once the contribution of critical parenting was taken into account. Given the association found between punitive and critical parenting, possibly owing to the element of hostility common to these parenting dimensions, critical parenting might reflect the aspect of emotionally negative parenting specifically salient for children's IP more strongly than punitive parenting. These findings support models suggesting that psychological control (behaviors such as criticism that undermine self‐efficacy) is an important risk factor for IP (e.g., Barber, 1996).</p> <p>The association between IP and low positive and high critical parenting is consistent with recent research linking IP not only to negative parenting but also to a lack of positive parenting (Bayer et al., 2006; Coplan et al., 2008). Either or both the absence of positive responses and the presence of negative responses may play a role in the mechanisms mediating the effect of parenting on children's risk of developing IP. Parenting practices that foster understanding of rules and show respect for child autonomy should protect against IP by contributing to the development of self‐regulation and a sense of competence and self‐efficacy. Conversely, parenting practices that convey criticism and rejection of the child should increase the risk of IP by contributing to poor self‐regulation and a sense of incompetence. Research with multiple time points is needed to determine whether parenting does in fact have a lasting influence on the development of IP and whether it occurs through these processes. Such research would also permit analysis of the reciprocal influences that may occur between parenting and child inhibition.</p> <p>Family stress and parent characteristics showed both direct and indirect associations with IP via parenting. IP were predicted by socioeconomic stress directly and indirectly through less positive parenting and by maternal negative emotionality directly and indirectly through more critical parenting, associations reflecting significant mediation by parenting. These associations provide evidence that family stressors may be linked to young children's IP through multiple processes, in accord with ecological models (Bronfenbrenner & Ceci, 1994), that posit a chain of reciprocating influences leading from socioeconomic factors at the family level to the psychological resources of parents, the quality of parenting, and children's psychological adjustment (Conger & Dogan, 2007; Rubin & Mills, 1991). Thus, family and parent factors may contribute to the development of children's IP through multiple processes, including parenting practices, genetic transmission of emotionality from parent to child, or other learning processes.</p> <p>In summary, child temperamental inhibition, family socioeconomic status, maternal negative emotionality, and maternal low positive and high critical parenting appeared to be key independent factors in early‐emerging IP, and maternal negative emotionality appeared to have both direct and indirect associations with IP through critical parenting. Interestingly, there were no moderating effects of inhibition, possibly reflecting the relative lack of consistency in studies of the interactive contributions of temperament and parenting to adjustment (Bates & Pettit, 2007). This study's failure to find significant moderating effects cannot be attributed to a lack of power, inasmuch as the integrated sample afforded excellent power to detect even small effect sizes. Rather, these results suggest that, in early childhood, separate individual factors may be sufficient by themselves to increase or decrease the risk of IP. To the extent that this is so, it would appear that developmental trajectories for IP may have multiple highly specific starting points, in accord with the principal of equifinality. Whether problems tend to persist, worsen, or improve thereafter, however, may depend on the cumulative and interactive effects of multiple risk and protective factors. To test this idea, research is needed to identify the factors associated with persisting, worsening, or improving trajectories of IP.</p> <p>There are limitations of the study that must be considered. Most of the measures were constructed by selecting items that were parallel in content due to the use of different measurement instruments in the three separate samples. Measurement error was likely higher than would have been the case with identical measures, and as such, the findings may represent an underestimate of the relations that exist. Nevertheless, as described in the <emph>Preparation for the Structural Model</emph> subsection of the <emph>Results</emph> section, this limitation is mitigated by the use of structural equation modeling techniques designed to account for measurement error by incorporating the individual sample reliabilities in the model. The results of the analyses supported most of the hypotheses and showed equality of the model across samples, providing replication of the findings across three independent samples. Another limitation is the use of reports from mothers only, which may have increased the degree of correspondence between them. Although the use of a single source of information may not entirely account for the pattern of relations that were found (e.g., consistent with theory, IP were predicted more strongly by critical than by punitive parenting), future work should replicate the findings using multiple raters and different methods of measurement.</p> <p>Also limiting the study is its cross‐sectional design, which necessitates caution in interpreting the causal directions reflected by the present results. The findings are consistent with theory and research suggesting that temperament, parenting, and family context contribute to the development of IP. Moreover, reciprocal relations between IP and parenting dimensions were nonsignificant. However, the possibility remains that IP are a cause as well as a consequence of less positive or more critical parenting. Due to their deference and appearance of vulnerability, internalizing children may elicit parental over‐ or under‐reactions, either of which could interfere with the development of emotion regulation skills and, therefore, contribute to anxiety. Thus, an important research direction is to examine trajectories of development across childhood. Such research is also needed to determine which aspects of children's psychological functioning serve as the mechanisms by which internal dispositional vulnerabilities and external relational and contextual influences become manifest as IP.</p> <p>Despite these limitations, the present study has important implications. Firstly, it demonstrates the power and efficacy of integrative data analysis and the advantages that can be afforded to researchers by combining complementary samples to investigate complex developmental questions. Secondly, the findings raise the possibility that single risk factors could be starting points for the development of IP. Single risk factors may lead to the worsening of IP through ripple effects that elevate other risk factors compounding the child's vulnerability. Thirdly, the findings highlight the need for more investigation of the role parenting may play in the development of IP. Parenting practices may be a key link in the chain of effects leading to or away from young children's IP, with family stressors affecting parenting practices both by decreasing positive parenting and by increasing negative parenting. Whether the associations found in the present study are in fact attributable to these processes awaits longitudinal analysis.</p> <hd id="AN0074102842-13">Acknowledgments</hd> <p>This research was supported by Canadian Institutes of Health Research grant MOP‐74642 awarded to the team, and by grants from Health Canada (Child and Youth Division), awarded to Lisa Serbin and colleagues from the Concordia Longitudinal Risk Project, from Social Sciences and Humanities Research Council of Canada and the Fonds de la Recherche en Santé du Québec grants awarded to Paul Hastings for the Daycare and Preschool Adjustment Study, and from Canadian Institutes of Health Research grant MOP‐57670 to Rosemary Mills for the Shame in Childhood Study. We thank the children and parents who made this research possible and gratefully acknowledge the assistance of Farriola Ladha, Claude Senneville, Nadine Girouard, and Bobbi Walling. We also thank Keith Widaman for statistical consultation.</p> <p>Publisher's Note, added on 21 September 2012 following initial online publication on 24 August 2011:</p> <p>Due to an error during the production process, the ordering of the author names in this article was changed.</p> <p>The original and correct author ordering is Mills, Hastings, Serbin, Etezadi, Stack, Schwartzman, Li and Helm, but the published version incorrectly listed the name ordering as Mills, Hastings, Helm, Serbin, Etezadi, Stack, Schwartzman and Li.</p> <p>The publisher would like to apologise for this error.</p> <ref id="AN0074102842-14"> <title> Footnotes </title> <blist> <bibl id="bib1" idref="ref1" type="bt">1</bibl> <bibtext> This research was supported by Canadian Institutes of Health Research grant MOP‐74642 awarded to the team, and by grants from Health Canada (Child and Youth Division), awarded to Lisa Serbin and colleagues from the Concordia Longitudinal Risk Project, from Social Sciences and Humanities Research Council of Canada and the Fonds de la Recherche en Santé du Québec grants awarded to Paul Hastings for the Daycare and Preschool Adjustment Study, and from Canadian Institutes of Health Research grant MOP‐57670 to Rosemary Mills for the Shame in Childhood Study. We thank the children and parents who made this research possible and gratefully acknowledge the assistance of Farriola Ladha, Claude Senneville, Nadine Girouard, and Bobbi Walling. We also thank Keith Widaman for statistical consultation.</bibtext> </blist> </ref> <ref id="AN0074102842-15"> <title> References </title> <blist> <bibtext> Abidin, R. R. (1995). Parenting stress index: Manual (3rd ed.). Odessa, FL: Psychological Assessment Resources.</bibtext> </blist> <blist> <bibl id="bib2" idref="ref2" type="bt">2</bibl> <bibtext> Achenbach, T. M. (1991). Manual for the child behavior checklist/4–18 and 1991 profile. Burlington, VT: University of Vermont, Department of Psychiatry.</bibtext> </blist> <blist> <bibl id="bib3" idref="ref3" type="bt">3</bibl> <bibtext> Achenbach, T. M. (1992). Manual for the child behavior checklist/2–3 and 1992 profile. 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  Data: Temperamental, Parental, and Contextual Contributors to Early-Emerging Internalizing Problems: A New Integrative Analysis Approach
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  Data: <searchLink fieldCode="AR" term="%22Mills%2C+Rosemary+S%2E+L%2E%22">Mills, Rosemary S. L.</searchLink><br /><searchLink fieldCode="AR" term="%22Hastings%2C+Paul+D%2E%22">Hastings, Paul D.</searchLink><br /><searchLink fieldCode="AR" term="%22Helm%2C+Jonathan%22">Helm, Jonathan</searchLink><br /><searchLink fieldCode="AR" term="%22Serbin%2C+Lisa+A%2E%22">Serbin, Lisa A.</searchLink><br /><searchLink fieldCode="AR" term="%22Etezadi%2C+Jamshid%22">Etezadi, Jamshid</searchLink><br /><searchLink fieldCode="AR" term="%22Stack%2C+Dale+M%2E%22">Stack, Dale M.</searchLink><br /><searchLink fieldCode="AR" term="%22Schwartzman%2C+Alex+E%2E%22">Schwartzman, Alex E.</searchLink><br /><searchLink fieldCode="AR" term="%22Li%2C+Hai+Hong%22">Li, Hai Hong</searchLink>
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  Data: <searchLink fieldCode="SO" term="%22Social+Development%22"><i>Social Development</i></searchLink>. May 2012 21(2):229-253.
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  Data: Wiley-Blackwell. 350 Main Street, Malden, MA 02148. Tel: 800-835-6770; Tel: 781-388-8598; Fax: 781-388-8232; e-mail: cs-journals@wiley.com; Web site: http://www.wiley.com/WileyCDA/
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  Data: <searchLink fieldCode="DE" term="%22Structural+Equation+Models%22">Structural Equation Models</searchLink><br /><searchLink fieldCode="DE" term="%22Parenting+Styles%22">Parenting Styles</searchLink><br /><searchLink fieldCode="DE" term="%22Inhibition%22">Inhibition</searchLink><br /><searchLink fieldCode="DE" term="%22Risk%22">Risk</searchLink><br /><searchLink fieldCode="DE" term="%22Personality%22">Personality</searchLink><br /><searchLink fieldCode="DE" term="%22Meta+Analysis%22">Meta Analysis</searchLink><br /><searchLink fieldCode="DE" term="%22Disadvantaged%22">Disadvantaged</searchLink><br /><searchLink fieldCode="DE" term="%22Behavior+Problems%22">Behavior Problems</searchLink><br /><searchLink fieldCode="DE" term="%22Children%22">Children</searchLink><br /><searchLink fieldCode="DE" term="%22Mothers%22">Mothers</searchLink><br /><searchLink fieldCode="DE" term="%22Rejection+%28Psychology%29%22">Rejection (Psychology)</searchLink><br /><searchLink fieldCode="DE" term="%22Stress+Variables%22">Stress Variables</searchLink><br /><searchLink fieldCode="DE" term="%22Correlation%22">Correlation</searchLink><br /><searchLink fieldCode="DE" term="%22Context+Effect%22">Context Effect</searchLink><br /><searchLink fieldCode="DE" term="%22Social+Development%22">Social Development</searchLink>
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  Data: <searchLink fieldCode="DE" term="%22Canada%22">Canada</searchLink>
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  Data: 10.1111/j.1467-9507.2011.00629.x
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  Data: 0961-205X
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  Data: This study evaluated a comprehensive model of factors associated with internalizing problems (IP) in early childhood, hypothesizing direct, mediated, and moderated pathways linking child temperamental inhibition, maternal overcontrol and rejection, and contextual stressors to IP. In a novel approach, three samples were integrated to form a large sample (N = 500) of Canadian children (2-6 years; M = 3.95 years; SD = 0.80). Items tapping into the same constructs across samples were used to create parallel measures of inhibited temperament, maternal positive, critical, and punitive parenting, maternal negative emotionality, family socioeconomic and structural stressors, and child's IP. Multiple-groups structural equation modeling indicated that associations were invariant across samples and did not differ for boys and girls. Child inhibition, less positive and more critical parenting, maternal negative emotionality, and family socioeconomic disadvantage were found to have direct associations with IP. In addition, maternal negative emotionality was associated with IP through more critical parenting, and both maternal negative emotionality and socioeconomic stress were associated with IP through less positive parenting. Results highlight the multiple independent and cumulative risk factors for early IP and demonstrate the power of integrating data across developmental studies.
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  Data: 80
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  Data: 2012
– Name: AN
  Label: Accession Number
  Group: ID
  Data: EJ961581
PLink https://search.ebscohost.com/login.aspx?direct=true&site=eds-live&db=eric&AN=EJ961581
RecordInfo BibRecord:
  BibEntity:
    Identifiers:
      – Type: doi
        Value: 10.1111/j.1467-9507.2011.00629.x
    Languages:
      – Text: English
    PhysicalDescription:
      Pagination:
        PageCount: 25
        StartPage: 229
    Subjects:
      – SubjectFull: Structural Equation Models
        Type: general
      – SubjectFull: Parenting Styles
        Type: general
      – SubjectFull: Inhibition
        Type: general
      – SubjectFull: Risk
        Type: general
      – SubjectFull: Personality
        Type: general
      – SubjectFull: Meta Analysis
        Type: general
      – SubjectFull: Disadvantaged
        Type: general
      – SubjectFull: Behavior Problems
        Type: general
      – SubjectFull: Children
        Type: general
      – SubjectFull: Mothers
        Type: general
      – SubjectFull: Rejection (Psychology)
        Type: general
      – SubjectFull: Stress Variables
        Type: general
      – SubjectFull: Correlation
        Type: general
      – SubjectFull: Context Effect
        Type: general
      – SubjectFull: Social Development
        Type: general
      – SubjectFull: Canada
        Type: general
    Titles:
      – TitleFull: Temperamental, Parental, and Contextual Contributors to Early-Emerging Internalizing Problems: A New Integrative Analysis Approach
        Type: main
  BibRelationships:
    HasContributorRelationships:
      – PersonEntity:
          Name:
            NameFull: Mills, Rosemary S. L.
      – PersonEntity:
          Name:
            NameFull: Hastings, Paul D.
      – PersonEntity:
          Name:
            NameFull: Helm, Jonathan
      – PersonEntity:
          Name:
            NameFull: Serbin, Lisa A.
      – PersonEntity:
          Name:
            NameFull: Etezadi, Jamshid
      – PersonEntity:
          Name:
            NameFull: Stack, Dale M.
      – PersonEntity:
          Name:
            NameFull: Schwartzman, Alex E.
      – PersonEntity:
          Name:
            NameFull: Li, Hai Hong
    IsPartOfRelationships:
      – BibEntity:
          Dates:
            – D: 01
              M: 05
              Type: published
              Y: 2012
          Identifiers:
            – Type: issn-print
              Value: 0961-205X
          Numbering:
            – Type: volume
              Value: 21
            – Type: issue
              Value: 2
          Titles:
            – TitleFull: Social Development
              Type: main
ResultId 1