Child Development, xxxx 2014, Volume 00, Number 0, Pages 1–16

Associations Between Family Structure Change and Child Behavior Problems: The Moderating Effect of Family Income Rebecca M. Ryan

Amy Claessens

Georgetown University

University of Chicago

Anna J. Markowitz Georgetown University

This study investigated conditions under which family structure matters most for child well-being. Using data from the Children of the National Longitudinal Survey of Youth (n = 3,936), a national sample of U.S. families, it was estimated how changes in family structure related to changes in children’s behavior between age 3 and 12 separately by household income level to determine whether associations depended on families’ resources. Early changes in family structure, particularly from a two-biological-parent to single-parent family, predicted increases in behavior problems more than later changes, and movements into single and stepparent families mattered more for children of higher versus lower income parents. Results suggest that for children of higher income parents, moving into a stepfamily may improve, not undermine, behavior.

Between 1960 and 1980, rates of divorce and nonmarital childbirth rose precipitously in the United States, and the latter increase continued unabated until 2008 (Martin, Hamilton, Ventura, Osterman, & Mathews, 2013; U.S. Census Bureau, 2006). These trends mean most children in the United States will experience one or more changes in family structure, from a two-biological-parent to single-parent family, or from either of these family structures into a stepparent family (Andersson, 2002; Bumpass & Lu, 2000). There is public and policy concern over these trends because children who have experienced family change experience poorer child cognitive and behavioral outcomes than those from intact families (Sigle-Rushton & McLanahan, 2004). Public policy attempts to reduce incidence of family change or ameliorate its apparent effects take three broad approaches (a) promoting marriage, (b) promoting father involvement, and (c) reducing economic This work was supported by an Emerging Scholars Small Grant to the first author awarded by the University of Wisconsin’s Institute for Research on Poverty. The authors thank Katherine Magnuson, Daniel Meyer, Marcia Carlson, Timothy Smeeding, and all the members of the University of Wisconsin Institute for Research on Poverty’s Emerging Scholars Small Grant Program 2012–2013 for their invaluable intellectual guidance. We also thank four anonymous reviewers for their helpful comments. Correspondence concerning this article should be addressed to Rebecca M. Ryan, Department of Psychology, Georgetown University, 3700 O Street, NW306 White Gravenor, Washington, DC 20057. Electronic mail may be sent to [email protected].

strain among single-parent families, for example, by ordering and monitoring child support payments. These policies assume that associations between family change and child development are as strong —or even stronger—in poor or near-poor families as they are in higher income families. With far higher rates of family instability, low-income families are the implicit targets of many of these policies. The present study tests this assumption by investigating the conditions under which family structure changes matter most to child well-being. Using longitudinal data on children’s behavior problems between birth and age 12, we ask if changes in family structure relate to changes in children’s behavioral outcomes differently for children born into low-, moderate-, and relatively high-income families. Comparing associations across income levels can tell us which group, if any, is most impacted by family changes, and in doing so illuminate the wisdom of targeting, explicitly or implicitly, marriage promotion and responsible fatherhood programs to low-income families. We also ask if the associations between family structure changes and child’s behavior vary by child age at the time of disruption. Previous research has found that the © 2014 The Authors Child Development © 2014 Society for Research in Child Development, Inc. All rights reserved. 0009-3920/2014/xxxx-xxxx DOI: 10.1111/cdev.12283

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effects of family instability on children’s behavior are strongest during the first 5 years of life (Cavanagh & Huston, 2008; Ryan & Claessens, 2013). However, no study has explored whether this pattern is consistent across income levels. Finally, we use a longitudinal approach that compares children to themselves over time, rather than comparing children across families, a strategy that reduces the influence of unobserved differences among families that may bias associations between family structure changes and children’s behavior. Family Instability and Family Income There is public and policy concern about the rise in family instability in the United States in part because family instability has been linked with poorer child outcomes, particularly in the behavioral domain (Cavanagh & Huston, 2008; Cooper, Osborne, Beck, & McLanahan, 2011; Fomby & Cherlin, 2007; Osborne & McLanahan, 2007; Waldfogel, Craigie, & Brooks-Gunn, 2010). Children who have experienced parental divorce have more behavioral problems, both internalizing and externalizing, than those in intact families, with effect sizes ranging between .22 and .33 (see Amato, 2001, for meta-analysis and review). Living in a stepparent or blended family has similar albeit smaller and less consistent associations with child behavior (Coleman, Ganong, & Fine, 2000; Ram & Hou, 2003; Thomson, Hanson, & McLanahan, 1994). Overall, these studies indicate that children who have experienced any kind of family change fare worse behaviorally than those in stable, two biological parent families. Policy efforts to promote marriage and encourage fathers’ parenting and monetary involvement primarily target low-income, disadvantaged families because rates of nonmarital childbirth and family instability are far higher in this population (Cherlin, 2005; Dion, 2005). This targeting, however, assumes that average associations between family instability and child behavior apply to low-income families. No prior study has directly tested this assumption. However, previous work finds that family instability has a larger negative impact on White children than African American children in terms of behavioral and cognitive outcomes (Amato & Keith, 1991; Dunifon & Kowaleski-Jones, 2002; Fomby & Cherlin, 2007). To the extent that White children come from more affluent households on average than African American children, it is reasonable to hypothesize that associations between family instability and child behavior would be

weaker, not stronger, for children in low-income families. Life course theory offers a perspective as to why family instability may matter less for children in low-income families than those with higher incomes. First, single-parent and stepparent families are more common among lower income families (McLanahan, 2004); thus, parents and children in them may perceive changes into these structures as more normative, more predictable, and, thus, less stressful (Elder & Shanahan, 2006; Maier & Seligman, 1976). A less stressful change should do less to alter parenting behavior and child well-being. Fathers in low-income families might contribute fewer economic resources to the home than higher income fathers. Divorce and separation, therefore, may not reduce the economic resources in the home as much at the lower end of the income distribution. Moreover, fathers in low-income families may spend less time on average interacting with their children (Guryan, Hurst, & Kearney, 2008) and have more emotionally strained relationships with their partners (Elder & Shanahan, 2006). Thus, their departure may also not undermine parenting and emotional resources in the home as much as the departure of a higher income father. By contrast, research on the association between income and child development suggest family instability may matter more for children in low-income than high-income families. Associations between changes in income and child outcomes have been found to be much larger, and in some cases reserved, for those at the lowest end of the income distribution (Dearing, McCartney, & Taylor, 2001; Duncan, Ziol-Guest, & Kalil, 2010). Reductions in economic resources have been shown to account for as much as half of the association between family change and child outcomes (McLanahan & Sandefur, 1994; Ram & Hou, 2003). If changes in economic resources account in part for links between family change and child development, and changes in economic resources matter more for those with less, family change might impact children in low-income families more than those in higher income families. Studies examining the effects of poverty on families’ socioemotional functioning suggest the same hypothesis. Family stress theory posits that economic hardship triggers emotional distress in parents, which can hinder a parent’s ability to be supportive, sensitive, and consistent with children and thus undermine child development (McLoyd, 1990). Income instability and economic hardship do predict higher levels of anxiety and depression among mothers of young children (Dearing, Taylor,

Family Change and Child Behavior

& McCartney, 2004). Parents in poverty—or at lowincome levels generally—may thus have weaker emotional resources with which to cope with the emotional strain of family change than those with higher incomes. If so, parent–child interactions, and thus child well-being, may suffer more as a result of family change in these families. It is possible that the associations between family change and child behavior, and thus the moderating impact of family income, will vary by type of change. Transitions into a single-parent family from a two-biological-parent family are theoretically detrimental to child well-being because children may lose crucial economic and emotional resources. Transitions into a stepparent family, by contrast, may undermine child well-being because the new social parent reorganizes family roles and relationships in ways that are stressful to children (Cavanagh & Huston, 2008; Fomby & Cherlin, 2007; Wu & Martinson, 1993). However, social parents may also increase economic and emotional resources at a crucial time in development. Indeed, in a previous study with the data used here, Ryan and Claessens (2013) found that once the negative impact of divorce or separation was held constant, movement into a stepparent family during middle childhood predicted reductions in children’s behavior problems relative to staying in a single-parent family. In the current article, we investigate whether these associations vary by family income. It is possible that moving into a single-parent family disrupts child behavior more in higher versus lower income families because higher income fathers offer more economic resources than their lower income counterparts, and thus children lose more in terms of these resources when unions dissolve. Likewise, moving into a stepparent family may benefit children in higher income families more than those with lower income families, because mothers in higher income families are more likely to repartner with men whose economic resources resemble their own (Jacobs & Furstenberg, 1986). These new partners, thus, may elevate the economic and emotional stability of the family more than stepfathers in lower income families. In these scenarios, family structure changes would impact children in higher income families more than those from lower income families for better and worse depending on the type of family transition. Alternatively, these patterns by type of change may obtain more strongly in low-income than highincome families. If changes in income matter more for those with less, then the exit of a biological father and entrance of a social father would harm

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and benefit, respectively, children’s developmental trajectories more in low- versus high-income families (Dearing et al., 2001; Duncan et al., 2010). Moreover, to the extent that low-income mothers have fewer emotional resources than higher income mothers (Dearing et al., 2004), they may be more negatively impacted by relationship dissolution, and more positively impacted by a new union, than their higher income counterparts. Timing of Family Change Life course theory and developmental research suggest that a family change during the first 5 years should alter children’s developmental trajectories more than a change experienced later because children are most dependent on the family context and most sensitive to contextual influences during this time (Elder & Shanahan, 2006; Shonkoff & Phillips, 2000). Children may also be most vulnerable to changes during the first 3 years as they undergo crucial brain development and form fundamental attachments with parents, processes that establish developmental trajectories that are mutable but increasingly difficult to change over time (Ainsworth, Blehar, Waters, & Wall, 1978; Bowlby, 1982). Although much of this research focuses on 0 to 3, the sensitive period in child development likely extends through the first 5 years (Shonkoff & Phillips, 2000). Indeed, previous research has found that family instability is most strongly associated with later behavioral outcomes when it is experienced between birth and age 5 (Cavanagh & Huston, 2008; Ryan & Claessens, 2013). We expect to find that early changes matter more than later changes, as in previous studies, but we explore whether the primacy of early instability obtains only or more strongly among families at certain income levels. Specifically, we anticipate that early changes will matter most within the income level or levels most impacted by family structure changes. We explore this question by estimating distinct associations between changes in family structure and changes in children’s behavior at four different developmental periods—early childhood (birth to age 3), preschool (age 4–5), middle childhood (age 6–9), and preadolescence (age 10–12)— and comparing associations across family income levels. Addressing Selection Bias All studies of family structure must address the fact that parents select into family structures based

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on characteristics that could influence child development. It is possible, for example, that parental characteristics such as poor emotional or behavioral health, low human capital, and interpersonal problems influence family disruption as well as parenting and children’s home environments, thus inducing a spurious association between family change and child well-being (Ginther & Pollack, 2004). Prior studies address this issue by controlling for a rich set of observable parent and child characteristics that may covary with family instability and child outcomes. However, no list of covariates can control for all potential confounding factors, particularly those that are difficult to observe. To account for both observable and unobservable characteristics, studies use models in which changes in family structure predict changes in child outcomes (e.g., Aughinbaugh, Pierret, & Rothstein, 2005; Dunifon & Kowaleski-Jones, 2002; Foster & Kalil, 2007; Magnuson & Berger, 2009; Ram & Hou, 2003), thus reducing the effect of stable child and family characteristics that could confound associations. The present study also uses change models; however, following Ryan and Claessens (2013), we use a longitudinal approach that estimates the effects of changes in family structure on changes in children’s behavior separately in early childhood, preschool, middle childhood, and preadolescence. This approach allows us to address the issue of selection bias while investigating the timing effects described above. The Present Study This study investigates one central research question: Do the effects of family structure changes vary by family income level at the time of the child’s birth? We do not offer directional hypotheses about the moderating effect of income because different theories predict different moderating effects. Life course theory suggest family changes should impact children in low-income families less than those in higher income families, a hypothesis supported by differences in family structure impacts by race/ethnicity. However, family stress theory, and the literature on family income and child outcomes, suggests family changes should impact children in low-income families more than those in highincome families. We also ask whether the effects of family structure changes are strongest when experienced during the first 5 years of life and whether these timing effects obtain across income levels. We answer these questions using a longitudinal data analytic approach that distinguishes not only

between family changes at four different developmental periods but also reduces the influence of family characteristics that could bias family change —child behavior associations.

Method Data and Sample Data were drawn from the children of the National Longitudinal Survey of Youth (NLSY). The NLSY is a nationally representative survey of youth who were aged 14–21 when interviewed in 1979 and were reinterviewed annually until 1994 and biennially thereafter. Beginning in 1986, the NLSY began following the children of the female NLSY respondents to assess their health, development, and overall well-being. Mothers were then interviewed about their children biennially. Because the NLSY gathered longitudinal information on mothers’ family structures, economic conditions, socioemotional well-being, and a range of other characteristics, the NLSY is ideally suited to investigating linkages between family environments and child developmental trajectories (Center for Human Resource Research, 2009). The analytic sample consists of children observed between birth and 12, pooled across waves from 1986 to 2008. Because children of mothers who were aged between 14 and 21 in 1979 had to be at least 12 years old by 2008, the sample is not nationally representative and contains an oversample of young, low-income mothers. It nonetheless provides a large, socioeconomically diverse sample with which to estimate associations between family change and child well-being across income levels. Data on children’s behavioral outcomes were first collected when children were between 3 and 4 years old (3/4) and again when children were between 5 and 6 years old (5/6), 7 and 8 years old (7/8), 9 and 10 years old (9/10), and 11 and 12 years old (11/ 12). To be included, children need to have data on behavioral outcomes for at least three time points between those ages. The analytic sample is further restricted to families with valid data on family structure at the time of the child’s birth and for each subsequent wave of data collection through age 12, yielding a final sample of 3,936 children. Some cases were missing data on important variables, such as income before the child’s birth and mothers’ socioemotional well-being. Missingness ranged from less than 1% for mother’s locus of control to 25% for income prior to birth. We followed Von Hippel’s (2007) recommendation to multiply

Family Change and Child Behavior

impute independent but not dependent variables. Although we did not impute the outcomes, they were included in the models used to impute the other variables (Von Hippel, 2007). Multiple imputation was conducted using the ICE command in Stata 12.0 (Royston, 2007; StataCorp LP, 2009), which is based on a regressionswitching protocol using chained equations. Following conventional guidelines (Graham, 2009), 30 imputed data sets were generated and coefficients and standard errors combined across data sets using the PROC MIANALYZE command for regression analyses in SAS 9.2 (SAS Institute Inc., 2008). Note, because income at the time of the child’s birth—our moderator—was imputed prior to dividing the sample into income groups, Ns for the subgroup models vary slightly across imputations.

Measures Family Structure Change At each interview the NLSY collected data from the child’s biological mother on her current household structure, including her marital status, whether the child’s biological father lived in the home, and whether a romantic partner who was not the child’s biological father lived in the home. We used this information to determine whether a family structure change occurred between intervals and the type of change. Any change in the mothers’ relationship status was considered a family structure change regardless of her marital status (e.g., the entrance of a romantic partner into the home was considered a move into a stepparent family even if the mother remained unmarried). We created family structure change variables for five intervals beginning with birth to ages 3–4. We then combined types to compare the most common changes and changes of greatest interest. These were changes from (a) two biological parents to a single mother, (b) two biological parents/stepparent family into a(nother) stepparent family, (c) a single mother into a stepparent family, or (d) any other change (stepparent family to two biological parents, blended family to single mother, single mother to two biological parents). Appendix S1 in the online Supporting Information displays frequencies of these changes during each change period for the full analytic sample and for children across family income levels.

5

Income Level To determine whether associations varied by family income, analyses were run separately by household income level in the 3 years prior to the child’s birth. The average over 3 years was used rather than birth year income alone to account for measurement error. It was important to use family income measured before the child’s birth to ensure that family structure changes did not impact the household income measure. Because changes in family structure can impact household income, they could partially account for links between family structure and child behavior, serving as mediators, but not moderators, of the association between family structure and child behavior. We compared families below 200% of the federal poverty line (FPL), the federal definition of low income (Addy & Wight, 2012); families living between 200% and 300% of the federal poverty line, as a measure of middle income; and families living above 300% of poverty, as a measure of high income. We ran supplementary analyses comparing families at or below 100% of the federal poverty line and at or below 185% FPL as the low-income group; however, findings were not substantially different using these specifications. Between 44% and 45% of the analytic sample was low income, 21%–22% of the sample was moderate income, and 34%–35% was high income before the child’s birth across imputations. Child Behavior Children’s behavioral outcomes were assessed using the total raw scores on the Behavior Problems Index (BPI; Zill & Peterson, 1986), a 28-item mother-reported measure of the frequency, range, and type of behavior problems the child displays. Many items were derived from the Achenbach Child Behavior Checklist and other child behavior scales. Responses were dichotomized to indicate the presence or absence of problem behaviors and then summed, yielding a score range of 0–28. The scale ranges and items were consistent across child age allowing for comparisons over time. Note, all analyses were also run separately for the internalizing and externalizing subscales of the BPI. However, because findings did not differ meaningfully by type of behavioral problem, scales were combined for the sake of parsimony. Descriptive statistics for child behavior problems are displayed in Table 1.

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Table 1 Descriptive Statistics by Income Level at Child’s Birth Full sample

BPI score Age 3/4 Age 5/6 Age 7/8 Age 9/10 Age 11/12 Child characteristics Race/ethnicity White Hispanic Black Male Low birthweight Firstborn Maternal and family characteristics Mother’s highest grade Mother’s age at child’s birth Maternal locus of control Maternal self-esteem Maternal AFQT Income in birth year Unmarried at birth Z-scored HOME score

Low income

Medium income

High income

M

SD

M

SD

M

SD

M

SD

8.22 8.12 8.33 8.16 7.55

5.42 5.76 6.13 6.29 6.04

9.17 9.13 9.42 9.35 8.68

5.74 6.23 6.50 6.86 6.55

8.00 7.71 7.90 7.73 7.13

5.38 5.46 5.94 5.99 5.85

7.13 7.07 7.22 6.90 6.37

4.76 5.06 5.48 5.36 5.13

52.28 20.38 27.34 51.30 7.73 31.22 12.98 28.1 8.78 21.98 39.98 4,3664 26.45 0.005

32.68 25.60 41.72 51.90 9.67 18.14 2.35 3.69 2.38 4.11 28.47 7,4159 0.99

11.88 27.41 9.32 20.91 24.95 1,9185 49.14 0.33

59.08 19.33 21.59 51.70 6.25 30.26 2.09 3.74 2.32 4.05 22.32 1,2525 1.08

13.09 27.96 8.72 22.19 42.95 38,660 12.84 0.13

73.53 14.25 12.22 50.26 6.11 48.79 1.94 3.66 2.28 3.99 26.12 13,392 0.88

14.35 29.08 8.10 23.25 57.65 7,8485 5.15 0.36

2.16 3.42 2.36 3.88 26.13 11,7002 0.8

Note. Data are drawn from the National Longitudinal Survey of Youth Maternal and Child Supplement. BPI = Behavior Problems Index; AFQT = Armed Forces Qualification Test; HOME = Home Observation for Measurement of the Environment.

Child Age Data on children’s behavioral outcomes were gathered at ages 3/4, 5/6, 7/8, 9/10, and 11/12. Data on family structures were gathered at these age intervals as well as at birth and ages 1–2. The first two intervals span the developmental periods of infancy and toddlerhood (birth and ages 1–2), the third and fourth intervals span the preschool years (ages 3/4 and 5/6), the fifth and sixth intervals span middle childhood (ages 7/8 and 9/10), and the seventh interval spans preadolescence (ages 11/12). The dummy variables used to capture changes between these intervals in analyses are described next.

Covariates We control for maternal characteristics that could influence family structure, parenting behavior, and home environment measured at or before the child’s birth. These include mothers’ highest grade completed and age at child’s birth. They also

include mothers’ self-esteem, measured with the 10item Rosenberg (1965) Self-Esteem Scale, and locus of control, measured with a 23-item version of the Rotter (1966) Locus of Control Scale in which higher scores indicate more external locus of control. The Self-Esteem Scale is widely used and has high internal validity. The Locus of Control scale correlates well with self-esteem, and education, but the alpha of the scale is low for the whole NLSY cohort (a = 0.36). As a measure of cognitive skill, we included mothers’ Armed Forces Qualification Test scores, a scale drawn from the mothers’ scores on the full Armed Services Vocational Aptitude Battery that focused on their arithmetic reasoning, word knowledge, paragraph comprehension, and numerical operations. We also controlled for household income in the birth year, and the Home Observation for Measurement of the Environment–Short Form (HOME–SF) in the year of the child’s birth (Caldwell & Bradley, 1984), a widely used selfreport and observational measure of the quantity and quality of cognitive stimulation and emotional support the child receives in the home. We control

Family Change and Child Behavior

for whether the mother was married at the time of the child’s birth to account for differences in child developmental trajectories by marital status. We also control for child characteristics that are likely related to both levels and changes in outcomes, including child race/ethnicity, low birthweight, firstborn child, and child gender. Descriptive statistics for all covariates are displayed in Table 1.

To determine associations between family structure and children’s outcomes at different ages, we estimate piecewise hierarchical linear models (HLM; Dunifon, 2005; Magnuson, 2007; Raudenbush, 2001; Raudenbush & Bryk, 2002; Singer, 1998). We use an HLM approach, rather than a standard regression model, because longitudinal models allows us to examine how changes in family structure predict changes in children’s behavior, an approach that reduces the influence of stable child and family characteristics that could confound associations. Piecewise HLM measures time in discrete intervals between child assessments, rather than as a continuous measure, providing a uniquely flexible way to estimate effects of family changes separately for each age interval. In these models, changes in family structure during the early and concurrent intervals are used to predict concurrent and later changes in child outcomes. The Level 1 model predicting BPI scores is depicted below.

þ p4ij 11=12tij þ etij

changes to adjust for family changes before age 3/4. The early changes are measured as four indicators for (a) changes from a two biological to single-parent family, (b) any two parent into a stepparent family, (c) single into a stepparent family, or (d) some other change, with no early change as the reference. The slopes are modeled as a function of these same changes in family structure, early and concurrent. p0ij ¼ b00j þ b01j ðSINGLE0ij  INTACT01ij Þ

Analytic Strategy

BPItij ¼ p0ij þ p1ij 5=6tij þ p2ij 7=8tij þ p3ij 9=10tij

7

ð1Þ

In this equation, the total BPI score of child i in family j at time t is a function of the child’s initial level of BPI at age 3/4, p0ij, and a series of timerelated dummy variables for each child assessment. In this formulation, 5/6tij takes on a value of zero for the first observation and then a value of one at each subsequent observation, and 7/8tij, 9/10tij, and 11/12tij follow this pattern. Thus, the change in child i’s behavior problems between ages 3/4 and 5/6 is p1ij, between ages 5/6 and 7/8 is p2ij, between ages 7/8 and 9/10 is p3ij, and p4ij represents the change in child i’s scores between ages 9/ 10 and 11/12. In the Level 2 models, children’s BPI scores at age 3/4 (the Level 1 intercept) and BPI changes over the four intervals (the Level 1 slopes) are taken as outcomes. The intercept is modeled as a function of early (birth to age 3/4) family structure

þ b02j ðSTEP0ij  2PARENTS01ij Þ þ b03j ðSTEP0ij  SINGLE01ijÞ

ð2aÞ

þ b04j ðOTHER0ij  OTHER01ij Þ þ b05j FAM01ij þ b06j CHILD01ij þ q0ij p1ij ¼ b10j þ b11j ðSINGLE1ij  INTACT0ij Þ þ b12j ðSTEP1ij  2PARENTS0ij Þ þ b13j ðSTEP1ij  SINGLE0ij Þ þ b14j ðOTHER1ij  OTHER0ij Þ þ b15j ðSINGLE0ij  INTACT01ij Þ

ð2bÞ

þ b16j ðSTEP0ij  2PARENTS01ij Þ þ b17j ðSTEP0ij  SINGLE01ij Þ þ b18j ðOTHER0ij  OTHER01ij Þ þ b19j FAM01ij þ b20j CHILD01ij þ q1ij Our interest is in the pattern of coefficients on growth or changes in BPI during the four intervals: p1i, p2i, p3i, and p4i. In Equation 2b, the change in BPI between ages 3/4 and 5/6, p1i, is predicted by a concurrent change in family structure, where b11j to b14j reflect coefficients estimating the effects of different family changes experienced between ages 3/4 and 5/6 (vs. no change in family structure), and by an early change in family structure, where b15j to b18j reflect changes in family structure between birth and age 3/4. Equations for p2i, p3i, and p4i proceed in this way. All covariates, including marital birth status, are controlled at the intercept and each time slope. Therefore, the reference group for changes during the preschool period are families who experienced no change in the preschool period, no change in the early period (because early changes are entered at all slopes), and were married at birth. By comparing the concurrent coefficients across equations (b11j–b14j), we can identify the developmental periods during which children are most impacted by a change in family structure in the short term. By comparing early change coefficients across equations (b15j–b18j), we can identify whether

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and when the effect of an early change emerges in childhood and whether those effects sustain, increase, or decrease over time. In additional models, we examine the lagged effects of changes experienced after the early period (between ages 3/4 and 5/6, etc.) that have significant concurrent effects on children’s behavior. We run all models separately for low-, moderate, and high-income families. By comparing the family change coefficients across models, we can assess the relative impact of family change on children outcomes across family income level. We formally test differences between these coefficients using a postestimation t test (Gujarati, 1995). A 90% confidence level is used for these tests because small subsample cell sizes substantially weaken our statistical power. Finally, to account for the clustering of children in families, the slopes from the Level 2 models are modeled as outcomes at Level 3. b00j ¼ c000 þ t00j for all b00j through b47j

ð3Þ

Results Table 2 displays results of models predicting children’s BPI scores at age 3/4 and changes in BPI scores over each change interval with early and concurrent family structure changes used as Level 2 predictors. Results are displayed separately for children born to low-, moderate-, and high-income parents. Coefficients that are significantly different across income models are indicated by different subscript letters. Results from additional models in which significant family structure changes were entered as lagged predictors on subsequent age slopes are reported next. Children of Low-Income Parents The results for children born to low-income parents are displayed in the first column of Table 2. The intercepts indicate that children born to lowincome parents have significantly higher initial levels of behavioral problems than their moderateand high-income peers. Among low-income children, the intercepts also indicate that those who experienced early change from a two biological to single-parent family had higher initial behavior problems (at age 3/4) than those who experienced no early change. No other early change was associated with higher behavior problems at age 3/4. As explained above, our primary interest is in how family structure changes predict the pattern of

growth in BPI during the four intervals; these estimates are less biased by child and family characteristics that select families into disruption than the intercepts. The nonsignificant coefficients for all family changes, in all age periods, indicate that no type of family change altered children’s long-term behavioral trajectories in low-income families. The one significant coefficient that emerged was a negative effect of an early move into a single-parent family on behavior changes during preadolescence. This effect reflects a recovery from initially higher levels of behavior problems, given the significant and positive intercept estimate for this group. Figure 1a displays children’s behavioral trajectories after preschool changes, as an exemplar, from two biological to single parents, into stepparent families and no changes in family structure. The means for the no-change group are calculated by taking the intercept from the low-income model as the age 3/4 BPI score and adding to that score each time slope to generate means for each subsequent age period. To calculate means for those who experienced a preschool change, the value of each family structure coefficient in the preschool period was added to the means for the no change group at preschool and all subsequent periods. As the coefficients suggest, no significant differences emerge between those who experienced any type of preschool change versus those who experienced none by age 11/12. Children of Moderate-Income Parents The results for children born to moderate-income parents are displayed in the second column of Table 2. Although none of the intercept coefficients for early family changes are significant, the coefficient for an early change from a two-biological-parent to single-parent family is not significantly different from same coefficient for children of lowincome parents. The slightly larger coefficient is nonsignificant because the standard error is larger, likely owing to the smaller sample size of the moderate- versus low-income subgroup. More importantly, an early change from a two-biological-parent to single-parent family is associated with a significant increase in behavior problems between ages 5/ 6 and 7/8 (middle childhood). This effect was significantly larger than the same effect for children of low-income families, although not significantly different from those in high-income families. No other type of change, at any age period, was associated with differences in children’s behavioral trajectories in moderate-income families.

Family Change and Child Behavior

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Table 2 Growth in Behavior Problems Predicted by Initial Level and Specific Changes in Family Structure Low income b Intercept Early changes Two bio to single Any two to step Single to step Other change Ages 3/4 to 5/6 Two bio to single Early change Concurrent change Any two to step Early change Concurrent change Single to step Early change Concurrent change Other change Early change Concurrent change Ages 5/6 to 7/8 Two bio to single Early change Concurrent change Any two to step Early change Concurrent change Single to step Early change Concurrent change Other change Early change Concurrent change Ages 7/8 to 9/10 Two bio to single Early change Concurrent change Any two to step Early change Concurrent change Single to step Early change Concurrent change Other change Early change Concurrent change Ages 9/10 to 11/12 Two bio to single Early change Concurrent change Any two to step Early change Concurrent change

8.81a

Medium income

High income

SE

b

SE

b

SE

0.36**

7.75b

0.34**

7.04c

0.23**

1.41a 0.04a 0.82a 0.11a 0.17a

0.68* 1.72 1.20 0.75 0.37

1.55a 2.22a 0.54a 0.21a 0.09a

1.34 1.95 1.89 1.27 0.33

0.91a 4.02a 0.79a 0.87a 0.15a

1.02 2.38† 2.75 1.78 0.23

0.05a 0.40a

0.70 0.49

1.26a 0.52a

1.28 0.75

0.99a 1.24a

1.04 0.63†

1.25a 0.72a

1.75 0.91

1.14a 1.49a

2.05 2.93

2.66a 0.43a

2.56 1.56

1.03a 0.13a

1.23 0.73

1.56a 0.92a

1.88 1.27

1.61a 0.66a

2.91 0.93

0.66a 0.78a 0.21a

0.76 0.59 0.23

0.70a 1.13a 0.14a

1.24 1.25 0.21

0.13a 2.00a 0.13a

1.83 1.65 0.14

0.03a 0.32a

0.48 0.49

1.65b 0.89a

0.81† 0.66

1.13ab 0.35a

0.66* 0.56

0.31a 0.99a

1.11 0.79

0.48a 0.75

1.67 1.36

0.31a 0.41

1.97 1.28

0.95a 0.32a

0.82 0.56

1.43a 0.18a

1.49 0.83

1.64a 0.69a

1.98 0.75

0.14a 0.12a 0.00a

0.46 0.51 0.23

0.48a 0.79a 0.14a

0.80 0.99 0.21

0.55a 1.40a 0.35a

0.87 0.93 0.14†

0.44a 0.02a

0.47 0.55

0.07a 0.95a

0.84 0.76

0.68a 0.72a

0.69 0.58

0.72a 0.78a

1.08 0.85

0.20a 0.29a

1.43 1.52

0.76a 1.02a

2.01 0.88

0.03a 0.14a

0.83 0.60

0.17a 0.29ab

1.61 1.04

0.06a 1.54b

1.64 0.82†

0.21a 0.07a 0.55a

0.48 0.49 0.23†

0.08a 0.21a 0.43a

0.85 0.92 0.21†

0.09a 0.38a 0.44a

0.90 0.82 0.13**

1.30a 0.25a

0.46* 0.66

0.81a 0.74a

0.77 0.81

1.15a 0.12a

0.68† 0.77

0.01a 0.83a

1.08 0.85

1.18a 0.36a

1.47 0.91

0.56a 0.82a

1.98 0.65

10

Ryan, Claessens, and Markowitz

Table 2 Continued Low income b Single to step Early change Concurrent change Other change Early change Concurrent change N

Medium income SE

b

High income SE

b

SE

1.26a 0.85a

0.82 0.71

0.03a 0.30a

1.78 1.06

1.51a 0.48a

1.65 0.93

0.01a 0.25a

0.46 0.54

0.70a 0.53a

0.80 1.09

0.30a 1.57a

0.88 1.00

1,733–1,778

804–861

1,330–1,381

Note. Data are drawn from the National Longitudinal Survey of Youth Maternal and Child Supplement. Coefficients with different subscript letters are significantly different at the 90% confidence level. † p < .10. *p < .05. **p < .01.

Figure 1b plots the trajectories for children who experienced changes during the preschool period versus those who experienced no changes. Means are calculated in the same way as those for the low-income sample. As the coefficients for preschool changes suggest, no significant differences emerged between those who experience any type of preschool change versus those who experienced none by age 11/12. The behavioral trajectories for children who experienced an early change from two biological parents to a single-parent family were also calculated, although not shown in the figure. The difference in behavior problems at age 11/ 12 between children who experienced an early change from a two biological to single-parent family and those who experience no change was .20 of a standard deviation (t = 1.42; p = .15). Although that difference was not significant, the effect approaches a moderate size. Children of High-Income Parents The results for children born to high-income parents are displayed in the third column of Table 2. No initial differences in behavior problems emerged between children who experienced early changes and those who did not, according to the nonsignificant intercept coefficients. Two significant differences in changes experienced later in childhood did emerge at the slopes, however. Children who experienced a preschool change from a two-biological-parent to single-parent family displayed a significant concurrent increase in behavior problems relative to those who experienced no change. When this preschool change was entered at each subsequent slope (not shown), it was not associated with significant increases or decreases in children’s

behavioral trajectories at any other period, suggesting the concurrent effect during the preschool period was sustained over time. The other significant difference to emerge was a negative effect of movement into a stepparent family during middle childhood (ages 7/8 to 9/10), suggesting children experience a reduction in behavior problems, relative to those who experience no middle childhood changes, when they move into a stepparent family. The coefficient was significant only at the trend level in the models displayed. However, when the two-parent to stepparent and single-parent to stepparent family groups were combined in supplementary models, the coefficient for movement into a stepparent family was significant at p < .01, despite being smaller in magnitude (b = 1.25, SE = .61, p < .01). This pattern suggests the negative effect is reserved for those who move from a single-parent to stepparent family, but that the association is estimated with weak statistical power. Figures 1c displays the trajectories for children who experienced changes during the preschool period versus those who experienced no changes. Because a preschool move into a single-parent family significantly predicted children’s behavior problems, to determine the ages 5/6, 7/8, 9/10, and 11/ 12 means for children who had experienced this change, supplementary models were run in which it was entered as a lagged predictor on subsequent age slopes (not shown). Results suggest children who move into single-parent families during the preschool period have behavior problem scores 0.31 SD higher by age 11/12 than children who experience no preschool change (t = 2.06, p < .05). Figure 1c also displays the negative effect of moving into a stepparent family during middle

Family Change and Child Behavior (a)

11

12.00 11.00 10.00 9.00 8.00 7.00 6.00 5.00 3 to 4

5 to 6

No Change

7 to 8

Two Bio to Single

9 to 10 Single to Step

11 to 12 Two Parents to Step

(b) 12.00 11.00 10.00 9.00

BPI Scores

8.00 7.00 6.00 5.00 3 to 4

5 to 6

No Change

(c)

7 to 8

Two Bio to Single

9 to 10 Single to Step

11 to 12 Two Parents to Step

12 11 10 9 8

d = .31

7 6 5 3 to 4

5 to 6

7 to 8

No Change

Two Bio to Single

Two Parents to Step

Single to Step 7/8 to 9/10

9 to 10

11 to 12

Single to Step

Child Age Figure 1. Behavior Problems Index (BPI) trajectories from ages 3/4 to 11/12 by preschool changes in family structure for (a) low-, (b) moderate-, and (c) high-income families.

childhood on behavior problems. To calculate means for this group at each age period, we assumed that children who moved into stepparent families in middle childhood experienced an earlier move into a single-parent family. Thus, the figure charts the hypothetical trajectory of a child who experienced a preschool divorce or separation and then moved into a stepparent family. In this scenario, the significant decrease in behavior problems associated with a move into a stepparent family follows a significant increase in behavior problems

during the preschool period, resulting in scores by age 11/12 nearly identical to those of children who experienced no changes. Covariates Although all covariates were entered at the intercept and slopes across models, their associated coefficients are not reported in Table 2 for the sake of parsimony. Interesting patterns emerged across income models, however. First, within income

12

Ryan, Claessens, and Markowitz

groups, mothers’ education level and marital status at birth did not significantly predict children’s behavioral trajectories, and among low-income families, African American and Hispanic children had significantly lower BPI scores at age 3/4 than White children. These findings suggest within relatively homogenous income groupings, other demographic characteristics have small effects on child behavior. The only consistent predictor of children’s BPI scores across income groups was scores on the HOME, which were negatively associated with BPI scores at age 3/4 for all groups (b = .79, SE = .13 for low income; b = .77, SE = .21 for middle income; and b = .46, SE = .17 for high income). Supplementary Analyses All analyses were run separately for boys and girls to test for differences across gender. Significant associations between family changes and children’s behavior problems—specifically the links between early and preschool moves from a two-biologicalparent to single-parent family and movement into a stepparent family during middle childhood for the high-income sample—were larger for girls than for boys. However, post hoc t tests did not reveal significant differences across boy and girl models. Analyses were also run in which moves from a two-biological-parent into a stepparent family and moves from a stepparent into a new stepparent family were separated. Too few children experienced the latter change, however, to estimate this independent effect reliability. Finally, models were run in which changes in household income over the early period, the preschool period, and between ages 7/8 and 9/10—the three ages at which significant family structure effects were detected—were entered at each time slope. These models tested whether reductions or increase in economic resources explained associations between family structure changes and children’s behavior. Significant family structure effects were substantively unchanged upon inclusion of these variables.

Discussion The foregoing investigation tested a central assumption that undergirds policies aimed at reducing incidence of family change or ameliorating its apparent effects: that associations are stronger—or at least as strong—in poor and near-poor families as they are in moderate- and high-income families. Results from piecewise HLMs do not necessarily support

this assumption: Significant effects of family structure changes only emerged for children born to moderate- and high-income parents, not to those born to low-income parents. Thus, results support the hypothesis that family structure changes impact children in higher income families more than those from lower income families and that they do so for better and worse depending on the type of family transition. Overall, these results suggest union dissolutions do influence children’s behavioral trajectories; however, they highlight the importance of family context to understanding the developmental implications of family instability. Because our approach examined how changes in family structure predict changes in child outcomes, it reduced the influence of time invariant characteristics that could confound family change—child behavior associations. Overall, few significant effects of family structure changes emerged using this conservative analytic approach. No family structure changes experienced during any age period were associated with children’s behavior problems in low-income families, and only one family structure change was associated with children’s behavior problems in moderate-income families. Two family structure changes were associated with changes in children’s behavior problems among high-income families, although these effects ran in opposite directions. These results suggest many factors other than family instability shape the course of children’s behavioral trajectories, particularly for children in low-income families. For this group, the quality of the home environment most strongly predicted children’s behavioral trajectories, suggesting programs to elevate the level of emotional support and cognitive stimulation provided in the home may enhance children’s socioemotional development more effectively than programs to promote marriage. The significant effects that did emerge were found among children born to moderate- and highincome families, indicating that the impact of family change varies across family contexts. Families with few economic resources at the outset may not suffer as much from the strain of family structure change as those with greater initial resources. Put simply, low-income families may have less to lose, in terms of both money and home environment quality, through separation or divorce than higher income families. A smaller change in terms of these potential mechanisms for low-income families would produce smaller changes in children’s behavior as a result. Moreover, given that single-parent and blended families are more common among

Family Change and Child Behavior

lower income families (McLanahan, 2004), both parents and children in them may perceive changes into these structures as more normative, more predictable, and, thus, less stressful (Elder & Shanahan, 2006; Maier & Seligman, 1976). Thus, these changes may do less to alter parenting behavior and child well-being. By contrast, among higher income families, the consequences of family dissolution might be more salient. Fathers in these families likely contribute a substantial proportion of the household income. The loss of their income, or a large part of it, could decrease the resources available for children in ways that alter behavior. Moreover, the stress mothers may experience as a result of this economic shock could negatively influence mother–child interactions and thus child well-being. Socioeconomically advantaged fathers also tend to spend more time with children (Guryan et al., 2008) and may interact with children in more sensitive and stimulating ways (Conger et al., 2002) than less socioeconomically advantaged fathers. If so, parental separation could meaningfully alter children’s daily lives and relationships with their fathers, distressing changes that could manifest in behavior problems. Finally, family structure change is less common among higher income families, as the frequencies displayed in Appendix S1 in the online Supporting Information indicate. A less normative change may be more unexpected, and, as a result, more stressful, particularly to the mothers. Mothers’ stress, in turn, could negatively impact children’s socioemotional well-being directly or through strained mother–child interactions. Our supplementary analyses suggest the loss of fathers’ time or the increase in mothers’ stress likely explains the association between family dissolution and child behavior in higher income families, rather than declines in income. Entering changes in family income did not reduce the effect of moves into a single-parent family during preschool on child behavior. Although these results suggest families’ socioemotional functioning explains the family structure–child behavior association, it is also possible that family income is measured with too much error in a single time period to reliably estimate the mediating effect of income. The present study also demonstrates the importance of type of family change to understanding the effects of family instability. Moving from a singleparent family into a stepparent family predicts decreases children’s BPI scores relative to those who experience no change during middle childhood. Because movements into stepparent families

13

typically follow divorces or separations, which are associated with increases in children’s behavior problems, this positive effect is more accurately described as a recovery rather than a benefit. Children’s behavior might improve when their mothers form more harmonious relationships after a period of marital conflict or single parenthood or when stepfathers bring additional economic resources into the home and alleviate maternal stress. This kind of transition may indeed benefit (or at least not harm) children’s behavioral development if it improves maternal parenting quality or introduces a higher quality father figure into the child’s home. The question remains, however, why the advantage emerges only for children in high-income families. Existing research suggests that low-income mothers tend to repartner with men who have greater economic resources than their child’s father (Bzostek, McLanahan, & Carlson, 2012), whereas married stepfathers partnered with divorced mothers tend to have lower incomes than married biological fathers (Hofferth, 2006). To the extent that low-income mothers are more likely to have a nonmarital birth than higher income mothers (McLanahan, 2004), one might thus expect repartnering to benefit children in low-income families equally or more than those in higher income families. But, even if low-income mothers are more likely to “trade up” upon repartnering, it is still true that higher income mothers repartner with men who contribute more economic and, possibly, parenting resources to the household than lower income mothers. If most mothers spend some period as single parents before repartnering, then the change for children in terms of money and parenting resources may simply be greater for those in higher income families in ways that uniquely benefit children’s socioeomotional well-being. The significant associations between family changes and child behavior problems emerged only for changes experienced during early childhood and preschool. We found weak or no associations for changes during subsequent periods. Our results thus replicate findings from prior work that family structure changes in the first 5 years of children’s lives are important for behavior throughout childhood (Cavanagh & Huston, 2008; Ryan & Claessens, 2013). These results suggest that family disruption during the first 5 years of life more strongly influences children’s development than changes later in childhood, a finding that supports both theory and research about the importance of early family environment on children’s trajectories (Kalil, Ryan, & Corey, 2012; Shonkoff & Phillips, 2000).

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Our investigation into the associations between family structure changes and changes in child behaviors is not without its limitations. Because we aimed to differentiate the effects of specific family structure changes at specific developmental periods, we did not examine the impact of multiple family changes on children’s trajectories. Doing so within this analytic framework would require computing combined effects of changes at different periods to create hypothetical family histories or estimating effects for extremely small numbers of children. As a result, our approach may obscure any nonlinear, compounding effects of multiple changes in family structure. Other limitations also reduce the scope and implications of our findings. First, the data used are observational; thus, our estimates may be influenced by omitted variables that covary with both family change and children’s development. Our analytic strategy adjusts for baseline confounds and timeinvariant child and family characteristics more rigorously than many similar studies of family instability; however, time-varying characteristics could have biased our findings. Second, because mothers reported family structure changes and children’s behavior, shared method variance could have biased our findings. This concern is tempered by our inclusion of a rich set of maternal characteristics, including self-esteem and locus of control, which could influence perceptions of children’s behavior. Third, family structure change may be measured imprecisely if mothers did not report the initiation or dissolution of a cohabiting relationship, or if mothers initiated and dissolved a relationship in between interviews. Finally, most effect sizes found here are modest. However, within the high-income group, moving from a two-biological-parent to single-parent family during the preschool period had a larger effect on children’s behavior problems than other key family characteristics, including mothers’ education and self-esteem. The only family characteristic more strongly associated with children’s behavior was early HOME scores, a measure of the quality of emotional support and cognitive stimulation in the home, suggesting that for children in higher income families, parental separation may have a practically significantly impact on behavioral trajectories. In sum, the present study was the first to ask whether associations between family changes and children’s development were consistent across family income levels while paying careful attention to the timing and type of family changes and the influence of selection bias. Our results suggest early family changes, particularly those from two-biologi-

cal-parent to single-parent families, may indeed elevate children’s behavior problems both concurrently and in the long term. Most significantly, our findings reveal the importance of considering family context when generalizing about the negative impacts of family instability. It is possible that children in disadvantaged families, although they experience more family instability on average, are not as impacted—for better or worse—by instability as their more advantaged counterparts. References Addy, S., & Wight, V. R. (2012). Basic facts about lowincome children: Children under age 18. Retrieved from http://www.nccp.org/publications/pdf/text_1049.pdf Ainsworth, M. S., Blehar, M. C., Waters, E., & Wall, S. (1978). Patterns of attachment: A psychological study of the strange situation. Hillsdale, NJ: Erlbaum. Amato, P. (2001). Children of divorce in the 1990s: An update of the Amato and Keith (1991) meta-analysis. Journal of Family Psychology, 15, 355–370. doi:10.1037/ 0893-3200.15.3.355 Amato, P. R., & Keith, B. (1991). Parental divorce and the well-being of children: A meta-analysis. Psychological Bulletin, 110, 26–46. doi:10.1037/0033-2909.110.1.26 Andersson, G. (2002). Children’s experience of family disruption and family formation: Evidence from 16 FFS countries. Demographic Research, 7, 343–363. doi:10. 4054/DemRes.2002.7.7 Aughinbaugh, A., Pierret, C. R., & Rothstein, D. S. (2005). The impact of family structure transitions on youth achievement: Evidence from the children of the NLSY79. Demography, 42, 447–468. doi:10.1353/dem. 2005.0023 Bowlby, J. (1982). Attachment and loss: Vol. 1. Attachment. New York, NY: Basic Books. Bumpass, L. L., & Lu, H. (2000). Trends in cohabitation and implications for children’s family contexts in the United States. Population Studies, 54, 29–41. Bzostek, S., McLanahan, S., & Carlson, M. (2012). Repartnering after a nonmarital birth. Social Forces, 90, 817– 841. doi:10.1093/sf/sos005 Caldwell, B. M., & Bradley, R. H. (1984). Home Observation for Measurement of the Environment. Little Rock: University of Arkansas. Cavanagh, S. E., & Huston, A. C. (2008). The timing of family instability and children’s social development. Journal of Marriage and Family, 70, 1258–1269. doi:10. 1111/j.1741-3737.2008.00564.x Center for Human Resource Research. (2009). NLSY79 child and youth adult data users guide. Columbus: Ohio State University. Cherlin, A. J. (2005). American marriage in the early twenty-first century. The Future of Children, 15, 33–55. doi:10.1353/foc.2005.0015 Coleman, M., Ganong, L., & Fine, M. (2000). Reinvestigating remarriage: Another decade of progress. Journal of

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Supporting Information Additional supporting information may be found in the online version of this article: Appendix S1. Frequencies of Changes at Each Age Interval for Full Sample and by Income Level at Birth

Associations between family structure change and child behavior problems: the moderating effect of family income.

This study investigated conditions under which family structure matters most for child well-being. Using data from the Children of the National Longit...
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