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Journal of Family Issues
DOI: 10.1177/0192513X07311232
2008;
2008; 29; 944 originally published online Jan 4,Journal of Family Issues
Shannon E. Cavanagh
Family Structure History and Adolescent Adjustment
http://jfi.sagepub.com/cgi/content/abstract/29/7/944
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944
Family Structure History
and Adolescent Adjustment
Shannon E. Cavanagh
The University of Texas at Austin
As patterns of union formation and dissolution in adult lives become complex,
the living arrangements of American children are becoming increasingly fluid.
With a sample (N = 12,843) drawn from the National Longitudinal Study of
Adolescent Health, this study attempted to capture this complexity by mapping
out children’s family structure histories across their early life course, investi-
gating the implications of these arrangements for their general adjustment, and
finally, identifying family processes that explained these associations. The find-
ings suggest that a sizable minority of young people experience dynamic
family structure arrangements. Moreover, family structure at adolescence best
predicted later emotional distress, and family structure at adolescence plus an
indicator of cumulative family instability across childhood best predicted cur-
rent marijuana use. More so than indicators tapping social control, levels of
family connectedness and parent–adolescent relationship quality were key con-
duits for these associations.
Keywords: family structure instability; adolescent emotional distress;
adolescent drug use; parenting processes
I
n the past half century, increases in divorce, nonmarital fertility, and
cohabitation, combined with declines in marriage and remarriage, have
dramatically altered the context in which American children and adoles-
cents are raised (Casper & Bianchi, 2002). For example, although point
estimates indicate that most children live with both biological parents, life
course estimates suggest that more than half of all children will spend some
time in an alternative family (i.e., single-parent, cohabiting-stepparent, or
Journal of Family Issues
Volume 29 Number 7
July 2008 944-980
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10.1177/0192513X07311232
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Author’s Note: Please address all correspondence to Shannon E. Cavanagh, Department of
Sociology, The University of Texas, 1 University Station A1700, Austin, TX 78712; e-mail:
scavanagh@mail.la.utexas.edu. This article is based on data from the Add Health project, a
program designed by J. Richard Udry (principal investigator) and Peter Bearman and funded by
Grant No. P01HD31921 from the National Institute of Child Health and Human Development
to the Carolina Population Center, University of North Carolina at Chapel Hill, with cooperative
funding participation from 17 other federal agencies. The author acknowledges the support of
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Cavanagh / Family Structure History and Adolescent Adjustment 945
married-stepparent family; Bumpass & Lu, 2000). These statistics paint a
portrait of a social institution and individual families in flux. In other
words, the growing diversity of family types in the population is matched
with an increase in family structure transitions in children’s life courses.
Importantly, this complexity in children’s living arrangements is linked to
their development. This study attempts to shed light on this link by map-
ping out children’s family structure histories across their early life courses
and then investigating the implications of these arrangements for their gen-
eral adjustment.
Two theoretical perspectives guide this study. First, the life course para-
digm, an orienting theoretical perspective, frames its basic aims. Specifically,
this perspective focuses attention on the linked lives of parents and their
children—the connections between parents’ romantic lives and their children’s
adjustment. Moreover, this perspective emphasizes the long view of life
course dynamics by moving beyond a focus of family structure at a point in
time to one that incorporates the whole of children’s family structure experi-
ences, including family structure statuses at key developmental phases, the
duration in these statuses, and the tempo and timing of family structure
change (Elder, 1998). To this end, this study investigates the link between the
marital histories of parents and the basic adjustment of their children during
adolescence. Coupling these foci demonstrates how the route to a particular
family structure helps to define its developmental impact.
Second, this study draws on the family process paradigm, an explanatory
theory, to answer the important why question: Why are the marital histories
of parents developmentally significant for their children? The family
process paradigm posits that parenting behaviors and family relationships
are key mechanisms underlying this form of linked lives (Amato, 2000;
McLanahan & Sandefur, 1994; Simons, 1996). Consequently, this study
attempts to explain how relationships and behaviors in the home serve as
conduits by which parents’ cumulative decisions about their romantic lives
affect their children’s well-being.
grants from the National Institute of Child Health and Human Development to Shannon E.
Cavanagh (HD046185-01), Kathleen Mullan Harris (HD07168-23), and Chandra Muller
(HD40428-02), as well as a Youth and Adolescent Dissertation Award from the Henry A. Murray
Research Center of the Radcliffe Institute for Advanced Studies at Harvard University, awarded
to Cavanagh. This research was also supported by a grant from the National Science Foundation
(REC-0126167) to the Population Research Center, The University of Texas at Austin, Chandra
Muller (principal investigator). Opinions reflect those of the author and not necessarily those of
the granting agencies. The author would like to thank Kelly Raley and Rob Crosnoe as well as
the editor and reviewers for their thoughtful comments.
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The National Longitudinal Study of Adolescent Health (or Add Health)
provides a valuable resource for this important line of family research
because it allows the construction of children’s perceptions of parents’ mar-
ital histories, the assessment of intrafamily dynamics, and the measurement
of a host of behaviors for a nationally representative sample of young
people in the United States. Utilizing this resource to unpack the link
between family composition and individual well-being will only become
more meaningful as American families continue to evolve and change.
A Closer Look at Family Structure
The structure of American families has undergone dramatic change over
the past 60 years. These changes are meaningful, in part, because of their
implications for child and adolescent well-being (Amato, 2000; McLanahan
& Sandefur, 1994). Although the magnitude and long-term implications of
family change continue to be contested (Booth & Amato, 2001; Cherlin, 1999;
Hetherington & Kelly, 2002; Wallerstein, Lewis, & Blakeslee, 2000), scholars
agree that young people in stable two–biological parent families generally
have higher levels of psychological well-being and engage in fewer problem
behaviors than do their peers in other family forms (Chase-Lansdale, Cherlin,
& Kiernan, 1995; Doherty & Needle, 1991; Johnson & Hoffman, 2000).
This body of work typically employs a static view of family structure,
capturing it at a point in parents’ marital histories (e.g., the family situation
at a given age or at the time of a survey). Young people, however, often live
in homes marked by multiple parental relationships, including spells of
marriage, divorce, single parenthood, cohabitation, remarriage, and all of
the above (Cherlin, Chase-Lansdale, & McRae, 1998; Hill, Yeung, &
Duncan, 2001; Osborne, 2005; Wu & Martinson, 1993).
An emerging literature has coupled retrospective reports of family struc-
ture histories with conventional point estimates to effectively capture the
complexity of children’s living arrangements (Bumpass & Lu, 2000; Hao
& Xie, 2002; Teachman, 2003; Wu & Martinson, 1993). For instance, more
than a third of all births occur outside of marriage (Hamilton, Martin, &
Ventura, 2006), with a third of these occurring within cohabiting relation-
ships (Bumpass & Lu, 2000; Osborne, 2005). At the same time, although
divorce rates have remained constant over the past 25 years, still about a half
of all marriages end in divorce, and half of these involve children (Amato,
2000). Remarriage rates have also declined, yet residing in a stepparent
family remains a common experience for American children (M. Coleman,
946 Journal of Family Issues
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Ganong, & Fine, 2000). An increasing proportion of these stepparent
families are formed through cohabitation, not marriage or remarriage,
making cohabiting-stepparent families an increasingly more prevalent family
structure status for young people (Bumpass, Raley, & Sweet, 1995). Last,
related to these changes in adult relationships, the likelihood that young
people will spend some time in single-parent family has increased signifi-
cantly (Bumpass & Lu, 2000). Although most young people reside in
mother-only families, father-only families are becoming more common
(Garasky & Meyer, 1996; Harris, Cavanagh, & Elder, 2000). Thus, time in
these different statuses suggests that family structure, from the perspective
of the child, is dynamic, evolving, and unstable.
Dimensions of Family Structure Histories
Guided by the life course perspective, this study seeks to extend this
emerging literature by exploring the complexity of children’s living arrange-
ments from birth through early adolescence. Viewing children’s living
arrangements longitudinally, assessing different family structure statuses (be
they families of two biological married parents, a single parent, or a married
or cohabiting stepparent), measuring the amount of time spent in a particular
family structure, and tracking transitions from one family status to another all
give family structure histories their form and meaning (Elder, 1998). The first
objective of this study, then, is to develop dimensions of family structure his-
tories that capture family structure across children’s early life courses.
Timing of family structure statuses is the first dimension considered. On
one hand, being born into an alternative family structure (e.g., single parent,
stepparent) may be most salient for adolescent adjustment because such a
status may undercut early socialization practices that can set the stage for
later, more risky development (Wu & Martinson, 1993). This status may
also serve as a proxy for the different economic contexts into which
children are born, as well as unmeasured selection factors (e.g., mother,
father, and child characteristics) that contribute to the likelihood of being
born into alternative families and being less well adjusted (Hao & Xie,
2002). On the other hand, residing in a single-parent family during adoles-
cence may undermine the ways that the resident parent monitors and super-
vises her or his children, which according to social control theory, increases
the likelihood that adolescents engage in inappropriate behaviors (Liska &
Reed, 1985; Wu & Martinson, 1993). Similarly, residing in a stepparent
family in adolescence may also increase the likelihood of emotional distress
Cavanagh / Family Structure History and Adolescent Adjustment 947
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and inappropriate behaviors because stepparents may lack the interest
and/or authority to monitor and supervise their stepchildren, especially in a
cohabiting-stepparent household (Teachman, 2003; Thornton, 1991). For
these reasons, I explore the associations between adolescent adjustment and
family structure at birth and during adolescence.
Another aspect of family structure timing is duration, or time spent in a
given family structure (Hao & Xie, 2002). Duration in a particular family
type can tap the extent to which household members are able to set and
negotiate practices, rules, and relationships within the family (Wu &
Martinson, 1993). Consider a 16-year-old adolescent born into a single-
mother family whose mother remarried when the child was 3 and stayed
married. This adolescent shares the same current family structure status as
another 16-year-old who was born into a two–biological married parent
family, resided in it through age 12, experienced a parental divorce, and then
the remarriage of his resident parent at age 14. The degree to which family
rules and relationships are established in each of these two situations is
likely different, thereby differentiating the implications of this particular
status for each adolescent (Hao & Xie, 2002). Overall, I expect that the
longer young people reside in a given family structure, the less likely they
will be to report behavior problems than those newer to the family structure.
The second dimension of family structure history that I consider is the
level of instability and change in family structure. This dimension is impor-
tant because changes in parents’ marital statuses constitute a major stressor
in a young person’s life. The loss or addition of a parental figure may dis-
rupt a child’s sense of security and create ambiguity in household rules,
family relationships, and parental expectations about behavior. This disrup-
tion is exacerbated by residential moves and dramatic changes in family
income and parents’ employment patterns (McLanahan & Sandefur, 1994;
Teachman, 2003; Wu & Martinson, 1993). Because family structure change
is repeatable across a child’s life, the stress associated with family instabil-
ity can accumulate over time. Thus, multiple family transitions are expected
to entail cumulative risk for adolescents (Cavanagh & Huston, 2006).
Just as the timing of different family structure statuses may have differ-
ential implications for young people, so too may the timing of family insta-
bility. Although less research has investigated the timing of family
instability, some researchers suggest that family structure change in early
childhood is most salient to child and adolescent development because dis-
ruption in this life stage undermines the quality of parent–child relationships,
which serve as the foundation for subsequent development (McLanahan,
1985; Parson & Bales, 1955; Yabiku, Axinn, & Thornton, 1999). Others
948 Journal of Family Issues
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point to the significance of parental monitoring and supervision during
adolescence for understanding adolescent adjustment (Liska & Reed, 1985;
McLanahan & Sandefur, 1994). Family instability during this stage might
be especially problematic because family rules and expectations designed
to control and support adolescents as they navigate this life stage may be
less well defined or enforced.
Together, these dimensions of family structure history provide a picture
of parents’ marital histories. The next section outlines how family structure
history, conceptualized in these ways, provides value to our understanding
of the role of family structure in the early life course.
Family Structure Histories and Adolescent Adjustment
Adolescent adjustment is certainly a broad construct, one that encom-
passes different behaviors and statuses. The two aspects of adolescent adjust-
ment considered in this study are emotional distress and current marijuana
use. These behaviors, although potentially serious to health and well-being in
adolescence and beyond, are part of the construction of adolescence in the
United States—10%–15% of the child and adolescent population has some
symptoms of depression at any one time, and nearly half of all high school
seniors have experimented with marijuana (Johnston, O’Malley, & Buchman,
2003; Smucker, Craighead, Craighead, & Green, 1986). Importantly, these
behaviors represent ways that young people cope with family change.
A large interdisciplinary literature has documented the associations
between family structure (measured at a point in time) and these indicators of
adolescent adjustment. First, parental divorce is associated with a rise in child
and adolescent psychological distress that typically dissipates within 2 years
following the event, although the negative effects of divorce can continue into
subsequent stages of life (Chase-Lansdale et al., 1995; Cherlin et al., 1998;
Cherlin & Furstenberg, 1994). Those in stepparent families also report higher
levels of emotional distress (Dawson, 1991). Second, a smaller body of
research suggests that parental divorce and remarriage are associated with
drug use, with young people in alternative families, especially those who
experience family disruption in adolescence, more likely to use marijuana
than others (Doherty & Needle, 1991; Hoffman & Johnson, 1998).
The second objective of this study, then, is to investigate the link
between multiple dimensions of family structure and adolescent emotional
distress and marijuana use. In doing so, this study identifies which of these
dimensions or which combination of dimensions best explains differences
in adolescent adjustment.
Cavanagh / Family Structure History and Adolescent Adjustment 949
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Explaining the Link Between Family Structure
Histories and Adolescent Adjustment
The life course perspective, as an orienting theory, focuses attention on
the link between parents’ marital histories and adolescent well-being and so
guides the conceptualization of family structure history. Yet this perspective
does less to address an equally compelling question: Why does this linked-
lives phenomenon exist? To answer this question, this study turns to an
explanatory theory: the family process paradigm.
This paradigm posits that alterations in family roles and functioning
brought on by changes in family status such as a divorce, remarriage, or
cohabitation explain the negative outcomes of these changes. Family
structure change can result in a home environment that is unstructured
and chaotic for young people. The entrance or exit of a parent’s romantic
partner can set in motion inconsistent parenting behaviors, brought on by
some combination of parental stress, economic insecurity, changes in a
resident parent’s work schedule, parental dating, or adjustment to new
household members (Amato, 2000). These changes introduce uncertainty
into adolescents’ lives, affecting how close they feel to their parents as well
as their connectedness with their families in general. In turn, parent–adolescent
closeness and family connectedness help to determine how effective
parents are at ensuring that their children engage in healthful behaviors
(Baurmrind, 1991).
Family transitions and their accompanying changes may also affect how
the resident parent or parents monitor and supervise their children’s everyday
lives, with lower levels of monitoring and supervision increasing adolescents’
opportunities to engage in problem behaviors (Amato, 2000; McLanahan &
Sandefur, 1994; Simons, 1996). This may be especially true if adolescents
are not close to their parents (J. S. Coleman, 1988). Taken together, the ero-
sion of family relationships and supervision in adolescents’ everyday lives
that can be triggered by family change may make young people more sus-
ceptible to emotional distress and drug use (Aseltine, 1996; Jessor, 1987).
The third objective of this study, then, is to investigate the degree to
which aspects of parenting behaviors and the parent–adolescent relation-
ship explain the links between family structure history and adolescent
adjustment. To pursue this objective, this study considers adolescent reports
of parental presence in the home, parental control over everyday aspects of
adolescents’ lives, warmth and closeness in the parent–adolescent relation-
ship, and adolescents’ feelings of connectedness with the family.
950 Journal of Family Issues
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Research Design and Methods
Source of Data and Sample
The data for this research come from Add Health, a nationally represen-
tative study of adolescents in Grades 7 through 12 in the United States in
1995. Add Health used a multistage, stratified school-based cluster sam-
pling design. For each study school, Add Health collected an in-school sur-
vey from every student (about 90,000) who attended on the day of
administration. About 1 year later, Add Health selected a nationally repre-
sentative sample from this pool of students to participate in the Wave 1 in-
home interview, conducted between April and December 1995. The Wave
2 in-home interview, nearly identical in content and form to Wave 1, was
conducted between April and September 1996 (see Bearman, Jones, &
Udry, 1997). Overall, about 14,700 adolescents completed both waves of
the in-home interview.
Two selection criteria were applied to create the analytical sample for
this study. First, because analyses required data from both in-home inter-
views, the sample was limited to adolescents who participated in Wave 1
and Wave 2 and who had valid sampling weights, which adjust for the dif-
ferential likelihood that certain social groups were included in Add Health
(Chantala & Tabor, 1999). Second, because of the focus on family structure
and relationships with parents, young people had to reside with at least one
biological or adoptive parent at birth to be included in the sample. Overall,
these criteria resulted in an analytical sample of 12,843.
This analytic sample is smaller than the full Wave 1 sample. The impor-
tant question is whether these exclusions introduced bias. The first criterion
is common to all longitudinal research using Add Health, and the second
excluded 534 adolescents. Although neither greatly biased the analyses
sample, these exclusions did result in a sample of adolescents who were
more advantaged and from more stable homes. These biases must be kept
in mind when interpreting results, although they are arguably offset by the
many benefits of Add Health, such as its large sample size and longitudinal
family data.
Measures
Table 1 presents descriptive statistics for all study variables, except those
dealing with family structure.
Cavanagh / Family Structure History and Adolescent Adjustment 951
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952 Journal of Family Issues
Adolescent emotional distress was based on adolescent responses to 20
items that capture symptoms of emotional distress. Most (n = 18) were
taken from the Center for Epidemiologic Studies Depression Scale
(Radloff, 1977) that ask about behaviors and feelings in the past week,
including feeling depressed, talking less than usual, and feeling unliked
(see appendix for complete list). The other 2 questions (frequent crying and
sleeplessness) refer to behaviors in the past year. Because responses for
some questions ranged from 0 to 3 and others from 0 to 4, all items were
Table 1
Descriptive Statistics for Study Variables (N
==
12,843)
Variable MSE% n
Outcomes
Psychological distress 0.00 0.00
Current marijuana use 26.0 4,450
Parenting behaviors
Parental presence 2.88 0.02
Parental control 2.01 0.05
Parent–adolescent closeness 3.47 0.01
Family connectedness 3.83 0.02
Individual characteristics
Male 50.1 6,266
Age 15.02 0.11
Race and ethnicity
White 67.5 6,926
African American 14.8 2,622
Asian American 4.2 980
Latino/Latina 12.1 2,144
Other 1.4 171
Children of immigrants 16.0 2,853
Family characteristics
Parents’ educational attainment
Less than high school 11.9 1,627
High school graduation 26.5 3,167
Some college education 29.6 3,642
At least college graduation 31.1 4,325
Missing education 0.6 82
Household income
Income less than $32,000 43.4 5,734
Missing income 8.5 1,151
Number of children in the home 1.48 0.03
Note: All values are weighted.
distribution.
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Cavanagh / Family Structure History and Adolescent Adjustment 953
standardized with a mean of 0 and a standard deviation of 1. The Emotional
Distress Index thus reflects the mean score across the 20 standardized items
(α=.88). Adolescent current marijuana use was based on adolescent
reports of marijuana use at Wave 2. Adolescents were asked about use since
the last in-home survey; current users were coded as 1, all others as 0.
Family structure history was indexed with a set of measures that reflect
family structure at birth; family structure at Wave 1; the duration of the
Wave 1 family structure; the number of parental divorces, remarriages, and
cohabitations that adolescents experienced from birth through early adoles-
cence; and the timing of these changes. Three sources of data from the
Wave 1 in-home interviews were used to construct these measures: youth
reports of current household composition (including resident parents, step-
parent, and parent’s partner), the duration of life spent with current house-
hold members, and reports of whether and for how long they lived with a
nonresident biological parent. These data were combined to create indica-
tors of resident parent composition for each year in an adolescent’s life
through Wave 1 (Heard & Harris, 2001).
For each year in an adolescent’s life, family structure was operational-
ized into eight mutually exclusive categories: two biological parents, bio-
logical mother and stepfather, biological father and stepmother, two adoptive
parents, surrogate parents (foster, relative, other parent type), single-mother
family, single-father family, and no parent or person acting in the parental
role. From this array, I created five indicators of family structure history:
family structure at birth; family structure at Wave 1; the number of years in
the Wave 1 family structure; a count of the cumulative changes in the resi-
dent parent’s marital trajectory from birth through Wave 1; and counts of
changes in the resident parent’s marital trajectory during early childhood,
middle childhood, and early adolescence.
The first measure, family structure at birth, was captured with two indi-
cators that reflect whether the adolescent resided with a single parent (n =
2,373), a biological parent and a stepparent (n = 179), or two biological or
adoptive parents (n = 10,291) in the first year of life. Adolescents who spent
their first year of life living in a mother-only (n = 2,132) or father-only (n =
241) family composed the single-parent family at birth category. The sec-
ond measure, family structure at Wave 1, was based on a set of six binary
variables that included two–biological parent or two–adoptive parent
families (n = 7,482), married-stepparent families (n = 1,696), cohabiting-
stepparent families (n = 327), single-mother families (n = 2,663), single-
father families (n = 355), and other families (n = 320). The other family
category was rather heterogeneous and included young people who lived
distribution.
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954 Journal of Family Issues
with a surrogate parent, such as a grandmother, aunt, or foster parent, or with
no parent or parent-like figure. These adolescents were more likely to be
born into single-parent families and to experience more family instability
than all others.
The third measure captured duration, or the number of years in which
adolescents resided in their Wave 1 family structure. This measure ranges
from 1 to an adolescent’s age at Wave 1, depending on the adolescent’s
family structure history. For instance, an adolescent who has always lived
with both biological parents or with only his single mother will have a dura-
tion value equal to his age, but an adolescent who has spent the past 3 years
living with her single father following a spell with her single mother will
have a duration value of 3. Because duration can vary among adolescents
in the same family type, these six independent measures were entered
together in regression models (Hao & Xie, 2002). Measures of family struc-
ture at Wave 1 and duration in this structure distinguished between married-
and cohabiting-stepparent unions.
The fourth measure, cumulative family instability, was a count that
increased by 1 for each transition from one family structure status to another,
from birth to Wave 1. For example, an adolescent born into a two–biologi-
cal parent family who then resided with her single mother at age 5 and who
then lived with her single father from age 8 through Wave 1 would have a
family instability score of 2. Likewise, an adolescent who resided in a
mother-only family or a two–biological parent family through Wave 1
would each have an instability score of 0 (Cavanagh, Schiller, & Riegle-
Crumb, 2006). The final measure divided up the number of cumulative family
transitions into three development stages: early childhood, or the number
of changes from birth through age 5; middle childhood, or the number of
changes from age 6 through 11; and early adolescence, or the number of
changes from age 12 through Wave 1.
These family structure measures have a few caveats. No distinction
between formal (marriage) and informal (cohabitation) unions was made
in the construction of measures reflecting two–biological parent families or
family instability. Consequently, those born into a cohabiting two–biological
parent family were included with married two–biological parent families
and married two–adoptive parent families, and a transition from cohabita-
tion to marriage for a given couple (biological parents or not) was not
included in measures of family instability. Recall, however, that distinc-
tions between cohabiting- and married-stepparent families at Wave 1 are
identified. Moreover, if sample members experienced multiple family struc-
tures in a given year, only one of the statuses is reflected in these measures.
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Finally, transitions that a child experienced that are separate from her or his
resident parent’s transitions were not included (e.g., a child of divorced
parents may have lived for a period with his aunt and experienced her
divorce and remarriage but at the time of Wave 1 was living with his bio-
logical mother). These limitations underestimate the level of family insta-
bility and the time spent in cohabiting families for this sample.
Four indicators that tapped parenting practices and family relationships, each
measured at Wave 1, were constructed. Beginning with family relationships,
family connectedness was the mean response to four items (0 = low, 4 = high)
that asked adolescents how much people in their family understood them, how
much they and their family had fun together, how much they wanted to leave
home (reversed), and how much their family paid attention to them (α=.74).
Second, a series of items asked adolescents about their level of closeness to their
resident mother and father, as well as their closeness to their nonresident bio-
logical mother and father, if applicable; their satisfaction in their relationship
with their resident mother and father; the extent to which they felt their resident
mother and father was warm and loving to them and cared for them; and about
their satisfaction with their communication with their resident mother and father.
Items were asked about each parent separately. To allow for the range of parental
figures to be considered in this measure, the maximum value for each item was
selected (i.e., the maximum score on satisfaction in relationships with resident
mother and father). Because the closeness-to-parent item was asked of resident
parents and nonresident biological parents, the maximum mother questions (res-
ident and nonresident) and the maximum father questions (resident and nonres-
ident) were calculated, and then the maximum of these measures was used to
calculate the final measure. Parent–adolescent closeness was the mean response
(0 = low, 4 = high) of these five items (α=.85).
These dimensions of family relationships are often combined into a sin-
gle scale (e.g., Crosnoe, 2004; Resnick et al., 1997), but researchers suggest
that global measures of family relationships obscure the mechanisms by
which family processes explain adolescent development (Dorius, Bahr,
Hoffman, & Harmon, 2004). Because of the focus on diverse family
forms—which included parents as well as full, step-, and half siblings—
uncoupling these aspects of family relationships was important. Although
these items were correlated, the r was less than .55.
Turning to parenting behaviors, parental control was based on seven items
that asked adolescents if they made all their own decisions about curfew on
weekends, friends, clothes, how much television to watch, which television
shows to watch, time to go to bed on weeknights, and what to eat. Each item
was reverse coded so that higher values indicated greater parental control.
Cavanagh / Family Structure History and Adolescent Adjustment 955
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956 Journal of Family Issues
The index averaged these items across adolescent reports (α=.63). Parental
presence, ranging from 0 to 4, indicated whether a parent was present in the
home most or all of the time when the adolescent went to school in the morn-
ing, came home from school in the afternoon, ate evening meals (five to seven
dinners a week), and went to bed at night (α=.32).
Four individual-level controls, all measured at Wave 1, were included in all
analyses. Age and gender account for adolescents’ differential opportunities to
experience family structure change (age) as well as emotional distress and
drug use (Hoffman & Johnson, 1998; Umberson, Williams, & Anderson,
2002). Adolescents’ race and ethnicity (dummy variables for non-Hispanic
White, non-Hispanic Black, Asian American, and other) and immigrant-
generation status were also controlled to account for unmeasured characteris-
tics of social location and culture associated with family structure history,
emotional distress, and drug use (Gil, Vega, & Biafora, 1998; Hoffman &
Johnson, 1998; Hogan & Kitagawa, 1985). First-generation immigrants were
foreign-born youth of parents who were not born in the United States, or they
were U.S. citizens born abroad who migrated to this country as children.
Second-generation immigrants were born in the United States but had at least
one foreign-born parent. Third-generation immigrants (and above) included
adolescents born in the United States to U.S.-born parents (Harris, 1999).
Three Wave 1 family-level controls were included. First, parents’ educa-
tional attainment, based on parent reports, was measured as the highest
number of years of schooling completed by the parent and her or his cur-
rent partner. Responses were coded into four dummy variables: college
education or more, some college education, high school graduation or GED
(reference category), or less than a high school education. Adolescent
reports were used when parent reports were missing (n = 1,441). For ado-
lescents who were missing both parent reports and self-reports of parent
education, parents’ educational attainment was coded as high school grad-
uate and a missing flag was included in the analyses (n = 82). Low family
income, again based on parent reports, was a binary measure that indicated
whether the family earned less than $32,000 per year. Those for whom
income information was missing were included in the reference category,
with a binary variable indicating missing data. Finally, the number of
children in the home was also controlled in these analyses.
Analytic Techniques
The first objective of this study was to develop measures of parents’ mar-
ital histories. Descriptive analyses explored these dimensions. The second
distribution.
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objective concerned the association between these histories and adolescent
adjustment. I explored these associations with a series of ordinary least
squares (OLS) regression models (when predicting emotional distress) and
logistic regressions models (when predicting marijuana use). Each dimen-
sion of family structure history was estimated separately to determine
which measure of family structure history resulted in the best-fitting model.
Given the large number of models estimated for each outcome, two mea-
sures of model fit are presented: the negative log likelihood (for logistic
regression models) or R
2
(for OLS regression models) and the BIC
(Bayesian information criterion) statistic. The BIC statistic is calculated as
–χ
2
+ df(ln N), where df is the degrees of freedom associated with the model
being estimated, N is the sample size (12,843), and χ
2
is the likelihood ratio
for testing the estimated model to a model with no covariates. The BIC sta-
tistic is negative when the estimated model is a better fit than the null
model; models with more negative BIC values fit the data better than do
those with less negative values.
All statistics are presented; however, following Teachman (2003), I
place greater emphasis on differences in the BIC statistics. Because con-
ventional test statistics are influenced by sample size and the number of
variables in the model, it is difficult to reject poor-fitting models using the
log likelihood statistic or R
2
. More important, substantively uninteresting
differences between models can be considered to be statistically significant
when sample sizes are large. Given these limitations and the large sample
in Add Health, the BIC statistic is a useful tool for model selection
(Rafferty, 1995).
The third objective concerned the role of family processes in explaining
or mediating the link between family structure histories and adolescent
adjustment. Each of the linkages among family structure history, dimen-
sions of family processes, and adolescent adjustment was tested in a multi-
variate context (Baron & Kenny, 1986), but for the sake of brevity, only
findings from models that include parenting behaviors (i.e., parental pres-
ence, parental supervision) and relationship factors (i.e., parent–child
closeness, family connectedness) are presented, estimated separately and
then jointly with family history measures from the best-fitting model iden-
tified in the second objective (full results available upon request).
All analyses used sampling weights to adjust for Add Health’s complex
sampling design. A robust cluster estimator was also used to adjust standard
errors for the clustered nature of the sampling frame (StataCorp, 2001).
Cavanagh / Family Structure History and Adolescent Adjustment 957
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Results
Family Structure Histories
The first objective of this study concerned the measurement of parents’
marital histories. The following section explores these family structure
history measures in depth (Table 2). The first dimension was the timing of
family structure statuses. Most children spent their first year of life in
two–biological parent or two–adoptive parent families (80%); about 19%
resided in single-parent families; and a handful resided in stepparent
families (1%). By early adolescence, more heterogeneity in family structure
was evident. Although two–biological parent families remained the modal
family type (58%), about a quarter of children resided in single-parent
families (21% with mother, 3% with father); another 16% resided in step-
parent families (13% with a parent and his or her married partner, 3% with
a parent and his or her cohabiting partner); and 3% resided in homes with
no parent figure present.
Duration in Wave 1 family structure was also considered. Not surprisingly,
those in two–biological parent families lived longest in their current family
structure (M = 14.96 years), followed by those in mother-only families
(M = 8.77), married-stepparent families (M = 8.09), father-only families
(M = 5.97), families with no parent figure present (M = 5.33), and finally,
cohabiting-stepparent families (M = 5.16).
Regarding family instability, most children experienced no family struc-
ture change from birth though Wave 1, but about a third of the sample expe-
rienced the exit or entry of a parent or a parent’s romantic partner at least
one time. About 19% experienced one family transition, 13% experienced
two, 4% experienced three, and 1% experienced four or more family struc-
ture transitions from birth through early adolescence. Examining family
instability at different developmental stages, about 13% of young people
experienced at least one transition during early childhood; about 17% expe-
rienced at least one family transition between ages 6 and 11 (about 5%
experiencing two or more transitions during this life stage); and about 15%
experienced at least one transition during adolescence (about 5% experi-
encing two or more transitions). Not surprisingly, in comparison with the
sample as a whole, adolescents who were older at Wave 1 and who there-
fore had more opportunity to experience family structure change during
adolescence reported more transitions during this development stage than
did others.
958 Journal of Family Issues
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Cavanagh / Family Structure History and Adolescent Adjustment 959
Family Structure History and Adolescent Emotional Distress
The second objective of this study was to investigate the association
between dimensions of family structure history and adolescent adjustment.
The first set of multivariate regression models sought to identify the aspects
of family structure history that best predicted adolescent reports of emo-
tional distress (Table 3).
Table 2
Descriptive Statistics for Family Structure History
Variables (N
==
12,843)
Variable MSE% n
Family structure at birth
Two–biological or adoptive parents 80.2 10,176
Stepparent family 1.3 179
Single parent family 18.5 2,488
Family structure at Wave 1
Two–biological parent family 58.3 7,310
Married stepparent family 13.3 1,749
Cohabiting stepparent family 2.5 307
Mother-only family 20.7 2,764
Father-only family 2.8 348
Other family form 2.5 365
Cumulative family instability 0.62 0.02
Number of transitions
0 63.3 8,130
1 18.5 2,379
2 13.0 1,671
3 3.6 467
4 1.1 146
5+ 0.4 51
Family transitions between
Ages 0 and 5 0.17 0.01
Ages 6 and 11 0.24 0.01
Ages 12 to Wave 1 0.19 0.01
Years in current family structure
Two–biological parent family 14.96 0.12 7,310
Married stepparent family 8.09 0.16 1,749
Cohabiting stepparent family 5.16 0.29 307
Mother-only family 8.77 0.18 2,764
Father-only family 5.98 0.38 348
Other family form 5.33 0.34 365
Note: All values are weighted.
distribution.
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Table 3
Ordinary Least Squares Regression Estimates Predicting Adolescent Depression (N
==
12,843)
Model 1 Model 2 Model 3 Model 4 Model 5 Model 6 Model 7 Model 8
bSEbSEbSEbSEbSEbSEbSE bSE
Family structure history
Family structure at birth
Single-parent family 0.022*** 0.00 0.013** 0.00
Stepparent family 0.021 0.02 0.016 0.02
Family structure at Wave 1
Married stepparent 0.020*** 0.01 0.016* 0.01 0.007 0.01
Cohabiting stepparent 0.041*** 0.01 0.036** 0.01 0.028* 0.01
Single-mother family 0.018*** 0.00 0.016** 0.00 0.009 0.00
Single-father family 0.046*** 0.01 0.043** 0.01 0.034** 0.01
Other family type 0.039*** 0.01 0.035** 0.01 0.027* 0.01
Number of years in Wave 1 family
structure
Two–biological parent family –0.001*** 0.00
Married stepparent 0.002 0.00
Cohabiting stepparent 0.000 0.00
Single-mother family 0.000 0.00
Single-father family 0.004
†
0.00
Other family type 0.003 0.00
Family instability
Cumulative change 0.009*** 0.00 0.002 0.00 0.004 0.00
Ages 0–5 0.009** 0.00
Ages 6–11 0.005 0.00
Ages 12–Wave 1 0.012*** 0.00
(continued)
960
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Table 3 (continued)
Model 1 Model 2 Model 3 Model 4 Model 5 Model 6 Model 7 Model 8
bSEbSEbSEbSEbSEbSEbSE bSE
Individual characteristics
Male –0.052*** 0.00 –0.052*** 0.00 –0.052*** 0.00 –0.052*** 0.00 –0.052*** 0.00 –0.052*** 0.00 –0.052*** 0.00 –0.052*** 0.00
Age 0.010*** 0.00 0.009*** 0.00 0.009*** 0.00 0.010*** 0.00 0.009*** 0.00 0.009*** 0.00 0.009*** 0.00 0.009*** 0.00
Race and ethnicity
African American 0.017*** 0.01 0.012* 0.01 0.012* 0.01 0.011* 0.01 0.016** 0.01 0.016** 0.01 0.012* 0.01 0.010* 0.01
Asian American 0.051*** 0.01 0.051*** 0.01 0.051*** 0.01 0.051*** 0.01 0.051*** 0.01 0.051*** 0.01 0.051*** 0.01 0.051*** 0.01
Latino/Latina 0.033*** 0.01 0.032*** 0.01 0.032*** 0.01 0.032*** 0.01 0.033*** 0.01 0.033*** 0.01 0.032*** 0.01 0.032*** 0.01
Other –0.007 0.01 –0.008 0.01 –0.009*** 0.01 –0.009 0.01 –0.009 0.01 –0.009 0.01 –0.009 0.01 –0.009 0.01
Children of immigrants –0.019** 0.01 –0.017** 0.01 –0.018** 0.01 –0.017** 0.01 –0.018** 0.01 –0.019** 0.01 –0.018** 0.01 –0.018** 0.01
Family characteristics
Parents’ educational attainment
Less than high school 0.026*** 0.01 0.030*** 0.01 0.033*** 0.01 0.033*** 0.01 0.035*** 0.01 0.034*** 0.01 0.033*** 0.01 0.032*** 0.01
Some college education –0.004 0.00 –0.004 0.00 –0.005 0.00 –0.005 0.00 –0.005 0.00 –0.005 0.00 –0.005 0.00 –0.005 0.00
At least college graduation –0.020*** 0.00 –0.022*** 0.00 –0.022*** 0.00 –0.022*** 0.00 –0.023*** 0.00 –0.022*** 0.00 –0.022*** 0.00 –0.022*** 0.00
Missing education 0.050 0.03 0.050 0.03 0.048 0.01 0.047 0.03 0.051 0.03 0.050 0.03 0.049 0.01 0.048 0.03
Household income
Income less than $32,000 0.020*** 0.00 0.012** 0.00 0.010* 0.00 0.009* 0.00 0.012** 0.00 0.012** 0.00 0.011** 0.00 0.010** 0.00
Missing income 0.005 0.01 0.004 0.01 0.005 0.01 0.005 0.01 0.006 0.01 0.006 0.01 0.006 0.01 0.005 0.01
Total number of siblings in home 0.003 0.00 0.003
†
0.00 0.004* 0.00 0.004* 0.00 0.003 0.00 0.003
†
0.00 0.004* 0.00 0.004* 0.00
Intercept –0.128 0.02 –0.129 0.02 –0.132 0.02 –0.128 0.02 –0.134 0.02 –0.128 0.02 –0.136 0.02 –0.134 0.02
R
2
.082 .082 .085 .085 .082 .081 .085 .086
BIC′ –909 –945 –958 –943 –950 –926 –951 –944
†
p < .10. *p < .05. **p < .01. ***p < .001.
961
distribution.
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The results shown in Model 1, the baseline model, indicate that most of
the individual and family control variables were significantly associated
with adolescent emotional distress. Older youth, girls, adolescents of color
(i.e., African American, Asian American, and Latino/Latina), and those in
lower-income households and with less-educated parents were significantly
more distressed than others. At the same time, adolescents whose parents
were immigrants were less distressed. Models 2, 3, and 4 added family struc-
ture statuses at different points in young people’s life courses. Results from
Model 2 suggest that when compared to those born into two–biological
parent or two–adoptive parent families, those born into single-parent
families were more depressed in adolescence. Those born into stepparent
families did not differ from those in two–biological parent or two–adoptive
parent families. Results from Model 3 indicate that adolescents living in
any type of alternative family at Wave 1 were significantly more depressed
at Wave 2 than were those in two–biological parent families. The family
structure effect appeared strongest among young people in cohabiting-step-
parent and single-father families. Additional analyses indicate that young
people in these family forms do not differ from one another on the measure
of emotional distress but are more depressed than those in mother-only
families and married-stepparent families (as well as those in two–biological
parent families).
Model 4 included indicators of family structure duration. These results
indicate that the longer that young people resided in married two–biological
parent or two–adoptive parent families, the less emotionally distressed they
were. Conversely, those who resided in father-only families for longer peri-
ods were somewhat more depressed. The BIC statistics associated with the
models indicate that all models fit the data better than does the baseline
model. Comparing the fit between Model 2 (BIC′=–945), Model 3 (BIC′=
–958), and Model 4 (BIC′=–943) suggests family structure at adolescence
best predicts adolescent depression.
Models 5 and 6 tested dimensions of family instability. Cumulative
family instability (Model 5) and instability in early childhood and adoles-
cence (Model 6) were significantly associated with adolescent emotional
distress. Again, the BIC statistics associated with these models indicate that
both fit the data better than the baseline model and that cumulative family
instability (BIC′=–950) fit the data better than instability measured at dif-
ferent stages across the early life course (BIC′=–926).
The last two models included dimensions of family structure timing and
family instability. Model 7 included indicators of timing and instability that
best fit the data—family structure at Wave 1 and cumulative family instability.
962 Journal of Family Issues
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Cavanagh / Family Structure History and Adolescent Adjustment 963
Model 8 added indicators of family structure at birth. Beginning with Model
7, the simultaneous consideration of family structure at Wave 1 and cumula-
tive instability altered the associations between these aspects of family struc-
ture history and emotional distress. For instance, cumulative family instability
was no longer associated with emotional distress once family structure at
adolescence was taken into account. Similarly, the strength of the associa-
tions between family structure statuses and depression were modestly
reduced, suggesting that part of the family structure effect is explained
by the instability that preceded it. Turning to Model 8, the addition of
family structure at birth also changed the strength of associations between
family structure history and emotional distress. Net of subsequent family
structure change, young people born into single-parent families remained
significantly more depressed in adolescence than those born into two–
biological parent or two–adoptive parent families. At the same time, the
associations between family structure at adolescence and emotional distress
was attenuated, with young people residing in married-stepparent families
and single-parent families during adolescence no longer significantly more
distressed than those in two–biological parent families.
Which specification of family structure history best explained the relation-
ship between family structure and adolescent emotional distress? The BIC sta-
tistics indicate that Model 3 (BIC′=–958), the more parsimonious model that
included only family structure at adolescence, better fit the data than did
Model 7 (BIC′=–951) and Model 8 (BIC′=–944). Therefore, current family
structure alone best predicted adolescent emotional distress in adolescence.
Family Structure History and
Adolescent Current Marijuana Use
A similar set of logistic regression models was estimated to determine
which aspects of family structure history best predicted current marijuana
use (see Table 4). Beginning with Model 1, older youth, Latino/Latinas,
those in the residual race and ethnicity category, and those whose resident
parents had some college education were more likely to recently use mari-
juana, whereas those who were children of immigrants and in families with
more siblings were less likely to engage in this behavior.
Models 2, 3, and 4 added family structure statuses at different points in
young people’s lives. According to Model 2, being born into single-parent
families increased the odds of current marijuana use by 48%. As with
emotional distress, those born into stepparent families did not differ from
those in two–biological parent or two–adoptive parent families. Living in
distribution.
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964
Table 4
Logistic Regression Estimates Predicting Current Marijuana Use in Adolescence (N
==
12,843)
Model 1 Model 2 Model 3 Model 4 Model 5 Model 6 Model 7 Model 8
be
b
be
b
be
b
be
b
be
b
be
b
be
b
be
b
Family structure history
Family structure at birth
Single-parent family 0.39*** 1.48 0.11 1.12
Stepparent family 0.37 1.45 0.37 1.45
Family structure at Wave 1
Married stepparent 0.39*** 1.47 0.17
†
1.18 0.07 1.07
Cohabiting stepparent 1.09*** 2.97 0.86*** 2.36 0.77*** 2.16
Single-mother family 0.56*** 1.75 0.43*** 1.53 0.36*** 1.43
Single-father family 0.77*** 2.15 0.60*** 1.83 0.52*** 1.68
Other family type 0.57*** 1.77 0.39* 1.47 0.30
†
1.35
Number of years in Wave 1
family structure
Two–biological parent family –0.04*** 0.96
Married stepparent –0.03** 0.97
Cohabiting stepparent 0.06* 1.06
Single-mother family –0.01 0.99
Single-father family 0.00 1.00
Other family type 0.02 1.02
Family instability
Cumulative change 0.23*** 1.26 0.12*** 1.13 0.14*** 1.15
Ages 0 –5 0.22*** 1.24
Ages 6–11 0.19*** 1.21
Ages 12–Wave 1 0.26*** 1.30
Individual characteristics
Male 0.00 1.00 0.00 1.00 0.00 1.00 0.00 1.00 0.00 1.00 –0.01 0.99 0.00 1.00 0.00 1.00
Age 0.17*** 1.19 0.17*** 1.19 0.17*** 1.19 0.20*** 1.22 0.17*** 1.19 0.17*** 1.19 0.17*** 1.19 0.17*** 1.19
(continued)
distribution.
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965
Table 4 (continued)
Model 1 Model 2 Model 3 Model 4 Model 5 Model 6 Model 7 Model 8
be
b
be
b
be
b
be
b
be
b
be
b
be
b
be
b
Race and ethnicity
African American –0.25 0.78 –0.34** 0.71 –0.39*** 0.68 –0.39*** 0.68 –0.29* 0.75 –0.29* 0.75 –0.40*** 0.67 –0.39*** 0.68
Asian American –0.08 0.92 –0.07 0.93 –0.06 0.94 –0.06 0.94 –0.06 0.94 –0.06 0.94 –0.06 0.94 –0.06 0.94
Latino/Latina 0.41** 1.51 0.39** 1.47 0.39** 1.47 0.39** 1.47 0.41*** 1.51 0.42*** 1.52 0.40*** 1.49 0.39*** 1.48
Other 0.63** 1.88 0.62* 1.85 0.59* 1.80 0.58* 1.78 0.59* 1.80 0.59* 1.80 0.58** 1.79 0.58* 1.79
Children of immigrants –0.49*** 0.61 –0.46*** 0.63 –0.45** 0.64 –0.45** 0.64 –0.48*** 0.62 –0.48*** 0.62 –0.46*** 0.63 –0.46*** 0.63
Family characteristics
Parents’ educational attainment
Less than high school 0.02 1.02 –0.01 0.99 –0.01 0.99 –0.01 0.99 0.03 1.03 0.03 1.03 –0.01 0.99 –0.02 0.98
Some college education 0.26** 1.30 0.25** 1.29 0.25** 1.28 0.25** 1.28 0.25** 1.28 0.25** 1.28 0.24** 1.27 0.24** 1.27
At least college graduation 0.11 1.12 0.14 1.15 0.16 1.17 0.16 1.17 0.14 1.15 0.14 1.15 0.16
†
1.17 0.17
†
1.19
Missing education 0.24 1.27 0.22 1.24 0.25 1.28 –0.22 0.80 0.31 1.36 0.31 1.36 0.28 1.32 0.28 1.32
Household income
Income less than 32,000 0.09 1.09 0.04 1.04 –0.08 0.93 –0.05 0.95 0.02 1.02 0.02 1.02 –0.07 0.93 –0.07 0.93
Missing income –0.13 0.88 –0.15 0.86 –0.16 0.85 –0.16 0.85 –0.13 0.88 –0.13 0.88 –0.16 0.85 –0.17 0.84
Number of children –0.09** 0.91 –0.08** 0.92 –0.06* 0.94 –0.06* 0.94 –0.08** 0.92 –0.08** 0.92 –0.06* 0.94 –0.06* 0.94
in the home
–2 log likelihood –7,137 –7,110 –7,045 –7,049 –7,074 –7,078 –7,036 –7,033
BIC′ –205 –241 –342 –325 –321 –294 –350 –337
†
p < .10. *p < .05. **p < .01. ***p < .001.
distribution.
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any alternative family structure during adolescence increased the odds of
marijuana use by at least 47% (Model 3). The observed effect was strongest
for young people in cohabiting-stepparent families. The likelihood that
these young people recently used marijuana was significantly greater than
all others, except those in father-only families.
Duration in current family structure was also important (Model 4).
Specifically, the longer that young people resided in married-parent
families—either with two biological/adoptive parents or with a biological
parent and his or her married partner—the less likely they were to use mar-
ijuana. Conversely, young people who resided in cohabiting families for
longer periods, net of other factors, were more likely to use marijuana.
Although each model highlighted interesting links between family structure
history and adolescent marijuana use and the BIC statistics indicate that all
models fit the data better than the baseline model does, the model fit statis-
tics favored Model 3 (BIC′=–241 for Model 2, BIC′=–342 for Model 3,
BIC′=–325 for Model 4). Again, measures of family structure in adolescence
best predicted lifetime marijuana use.
Both cumulative family instability (Model 5) and instability at each
stage of the early life course (Model 6) were significantly associated with
marijuana use. As illustrated in Model 5, each transition in childhood liv-
ing arrangement was associated with a 26% increase in the odds of using
marijuana. The BIC statistics associated with these models indicate that
both fit the data better than the baseline model does, but cumulative family
instability (BIC′=–321 in Model 5) better fit the data than did family insta-
bility at different stages (BIC′=–294 in Model 6).
As with emotional distress, Models 7 and 8 included dimensions of family
structure timing and family instability when predicting current marijuana use.
Model 7 included indicators of family structure at adolescence and cumula-
tive family instability—dimensions of family structure and family instability
that best fit the data. Model 8 added family structure at birth. Beginning with
Model 7, once cumulative family instability was taken into account, the coef-
ficients associated with each family structure status during adolescence were
reduced by at least 20%, with more than half the negative association
between marijuana use and living in a married-stepparent family, for
instance, explained by the instability that preceded it (b = 0.39 in Model 3,
b = 0.17 in Model 7). With the exception of married-stepparent families, all
family structure measures remained significantly associated with the likeli-
hood of adolescent marijuana use. The coefficient associated with cumulative
family instability was also reduced by nearly half (b = 0.23 in Model 5, b =
0.12 in Model 7). Family structure at birth was added to Model 8. Neither
966 Journal of Family Issues
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indicator of family structure at birth was associated with marijuana use in
adolescence. The effects of family structure at Wave 1 were modestly atten-
uated, and the coefficient associated with cumulative instability changed little
once family structure at birth was controlled. The BIC statistics associated
with Models 7 and 8 indicate that both fit the data better than the baseline
model does, but indicators of family structure at adolescence and cumulative
instability (BIC′=–350 in Model 7) better fit the data than did indicators in
Model 8 (BIC′=–337) and all other models.
Worth noting is that the associations between the control variables in
models predicting adolescent depression and current marijuana use changed
little once aspects of family structure history were added to the model. The
effect of being African American on both outcomes, however, did change
once family structure was taken into account. Beginning with emotional
distress, African American youth were significantly more depressed than
Whites, net of individual and family factors (Model 1 in Table 3). This
effect was attenuated but still significant once family structure was taken
into account, suggesting that part of the African American disadvantage in
mental health is explained by differences in family composition. Turning to
current marijuana use, African American youth did not differ from other
youth in the baseline model (Model 1 in Table 4). Differences between
African American and Whites emerged once family structure status was
taken into account; that is, African American youth were significantly less
likely to use once family structure was specified. Together, changes in the
associations among race, family structure, and adolescent adjustment sug-
gest that African American youth look about the same or better than Whites
once family structure is held constant.
The Role of the Family Environment
The final objective of this study was to investigate the degree to which
aspects of the family environment explained the link between family struc-
ture history and adolescent adjustment. Building on the best-fitting model
identified in the prior section, this set of analyses explored whether parent-
ing behaviors (i.e., parental supervision and control) and relationships
within the family (i.e., parent–adolescent closeness and family connected-
ness) mediated the associations between family structure history and the
two indicators of adolescent adjustment.
The baseline for analyses of emotional distress included only measures
of family structure at Wave 1 (Model 1; see Table 5). Parental control and
parental presence were added in Model 2. Each was significantly associated
Cavanagh / Family Structure History and Adolescent Adjustment 967
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with emotional distress, positively for parental control and negatively for
parental presence. With the exception of the residual family category, the
coefficients associated with each family structure status were attenuated
once these behaviors were taken into account, but all remained statistically
significant. The reduction was strongest for youth in father-only families,
where the effect was reduced by more than half (b = 0.046 in Model 1, b =
0.022 in Model 2).
Indicators of parent–adolescent closeness and family connectedness were
added in Model 3. These factors were both negatively associated with emo-
tional distress. Once taken into account, the effects were attenuated across
all current family structure statuses, with the married- and cohabiting-step-
parent family and mother-only family effects no longer significant at con-
ventional levels. These reductions suggest that the family structure effects
associated with these kinds of families were explained by differences in the
quality of family relationships. Model 4 considered the joint role of all
family environment measures. Each measure remained significantly associ-
ated with emotional distress. And with the exception of the residual family
category, the coefficients for all current family structure statuses were
reduced and no longer significant at conventional levels. The BIC statistic
(BIC′=–2,332) indicates that Model 4 is by far the best-fitting model for
adolescent emotional distress.
The best-fitting model previously identified served as the baseline for
the current marijuana use analyses (Table 6). Parental control and presence
were added, and both predicted lower lifetime marijuana use (Model 2).
The magnitude of the family structure and family instability coefficients
was reduced once these factors were taken into account. The greatest reduc-
tion in the family structure effect again involved father-only families, with
differences in indicators of social control reducing this coefficient by about
half (b = 0.60 in Model 1, b = 0.27 in Model 2). These family factors had
little effect on the cumulative family instability coefficient.
Parent–adolescent closeness was modestly associated with current mar-
ijuana use. Family connectedness, however, was significantly related to this
outcome, with a one-unit increase in family connectedness associated with
a 45% decrease in the odds of current marijuana use (Model 3). Once taken
into account, the family structure and cumulative family instability effects
were attenuated but remained statistically significant.
When considering the joint role of parenting behaviors and family relation-
ships (Model 4), only coefficients associated with cohabiting-stepparent
families and single mother families remained significantly associated with mar-
ijuana use. Cumulative family instability remained significantly associated
968 Journal of Family Issues
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Table 5
Ordinary Least Squares Regression Estimates Predicting Adolescent Depression,
Net of Parenting Behaviors (N
==
12,843)
Model 1 Model 2 Model 3 Model 4
bSEbSEbSEbSE
Family structure history
Family structure at Wave 1
Married stepparent 0.020*** 0.01** 0.016 0.01 0.009
†
0.01 0.008 0.01
Cohabiting stepparent 0.041*** 0.01** 0.034 0.01 0.019
†
0.01 0.018 0.01
Single-mother family 0.018*** 0.00** 0.012 0.00 0.009
†
0.00 0.008
†
0.00
Single-father family 0.046*** 0.01* 0.022 0.01 0.022* 0.01 0.015 0.01
Other family type 0.039*** 0.01*** 0.038 0.01 0.030** 0.01 0.030** 0.01
Parenting behaviors
Parental presence –0.015*** 0.00 –0.006*** 0.00
Parental control 0.005*** 0.00 0.006*** 0.00
Parent–adolescent closeness –0.018*** 0.00 –0.017*** 0.00
Family connectedness –0.050*** 0.00 –0.049*** 0.00
Individual characteristics
Male –0.052*** 0.00 –0.052*** 0.00 –0.045*** 0.00 –0.046*** 0.00
Age 0.009*** 0.00 0.009*** 0.00 0.003* 0.00 0.004*** 0.00
Race and ethnicity
African American 0.012* 0.01 0.009
†
0.01 0.016** 0.00 0.014** 0.01
Asian American 0.051*** 0.01 0.048*** 0.01 0.046*** 0.01 0.045*** 0.01
Latino/Latina 0.032*** 0.01 0.030*** 0.01 0.034*** 0.01 0.033*** 0.01
Other –0.009 0.01 –0.009 0.01 –0.013 0.02 –0.012 0.02
Children of immigrants –0.018** 0.01 –0.019** 0.01 –0.015* 0.01 –0.016** 0.01
(continued)
969
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Family characteristics
Parents’ educational attainment
Less than high school 0.033*** 0.01 0.034*** 0.01 0.034*** 0.01 0.033*** 0.01
Some college education –0.005 0.00 –0.006 0.00 –0.007 0.00 –0.028 0.00
At least college graduation –0.022*** 0.00 –0.024*** 0.00 –0.023*** 0.00 –0.023*** 0.00
Missing education 0.048 0.01 0.033 0.01 0.031 0.01 0.032** 0.01
Household income
Income less than 32,000 0.010* 0.00 0.010* 0.00 –0.008* 0.00 0.008* 0.00
Missing income 0.005 0.01 0.007 0.01 0.009 0.01 0.009 0.01
Number of children in the home 0.004* 0.00 0.004* 0.00 0.003
†
0.00 0.002
†
0.00
Intercept –0.132 0.02 –0.102 0.02 0.218 0.03 0.199 0.03
R
2
.085 .096 .176 .181
BIC′ –958 –1,096 –2,281 –2,332
†
p < .10. *p < .05. **p < .01. ***p < .001.
970
Table 5 (continued)
Model 1 Model 2 Model 3 Model 4
bSEbSEbSEbSE
distribution.
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at UNIV OF TEXAS AUSTIN on August 10, 2008 http://jfi.sagepub.comDownloaded from
Table 6
Logistic Regression Estimates Predicting Adolescent Lifetime Marijuana Use,
Net of Parenting Behaviors (N
==
12,843)
Model 1 Model 2 Model 3 Model 4
be
b
be
b
be
b
be
b
Family structure history
Family structure at Wave 1
Married stepparent 0.17
†
1.18 0.16 1.17 0.10 1.11 0.10 1.11
Cohabiting stepparent 0.86*** 2.36 0.75*** 2.12 0.71*** 2.03 0.65*** 1.92
Single-mother family 0.43*** 1.53 0.33*** 1.39 0.36*** 1.43 0.31*** 1.36
Single-father family 0.60*** 1.83 0.27 1.31 0.41** 1.51 0.24 1.27
Other family type 0.39* 1.47 0.39* 1.48 0.33
†
1.39 0.34
†
1.40
Cumulative family instability 0.12*** 1.13 0.10** 1.11 0.10* 1.10 0.10* 1.11
Parenting behaviors
Parental presence –0.20*** 0.82 –0.11** 0.90
Parental control –0.08*** 0.92 –0.07*** 0.93
Parent–adolescent closeness –0.14
†
0.87 –0.13 0.88
Family connectedness –0.60*** 0.55 –0.58*** 0.56
Individual characteristics
Male 0.00 1.00 0.13 1.01 0.08 1.08 0.09* 1.09
Age 0.17*** 1.19 0.01*** 1.14 0.11*** 1.12 0.08*** 1.08
(continued)
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Race and ethnicity
African American –0.40*** 0.67 –0.40*** 0.67 –0.34** 0.71 –0.36** 0.70
Asian American –0.06 0.94 –0.11 0.90 –0.12 0.89 –0.14 0.87
Latino/Latina 0.40*** 1.49 0.39*** 1.48 0.43*** 1.53 0.43*** 1.53
Other 0.58** 1.79 0.55* 1.74 0.58* 1.79 0.56
†
1.75
Children of immigrants –0.46*** 0.63 –0.45*** 0.64 –0.43** 0.65 –0.43** 0.65
Family characteristics
Parents’ educational attainment
Less than high school –0.01 0.99 0.05 1.05 0.02 1.02 0.06 1.06
Some college education 0.24** 1.27 0.22** 1.25 0.23** 1.26 0.22** 1.25
At least college graduation 0.16
†
1.17 0.13 1.14 0.17
†
1.19 0.16
†
1.17
Missing education 0.28 1.32 –0.19 0.83 0.09 1.09 –0.17 0.84
Household income
Income less than 32,000 –0.07 0.93 –0.05 0.95 –0.09 0.91 –0.08 0.92
Missing income –0.16 0.85 –0.13 0.88 –0.15 0.86 –0.13 0.88
Number of children in the home –0.06* 0.94 –0.05
†
0.95 –0.08* 0.92 –0.07* 0.93
–2 log likelihood –7,036 –6,977 –6,732 –6,708
BIC′ –350 –450 –917 –946
†
p < .10. *p < .05. **p < .01. ***p < .001.
Table 6 (continued)
Model 1 Model 2 Model 3 Model 4
be
b
be
b
be
b
be
b
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with current use. The BIC statistic for this model (BIC′=–946) indicates
that it was by far the best-fitting model estimated for this outcome.
Conclusion
The “ideal” structure of families continues to be a great source of debate
among the American public as well as among social scientists and policy
makers. This study sought to contribute to this debate by integrating the life
course perspective and the family process paradigm with an important data
source that combines detailed information about adolescents and their
families. Three key themes emerged from this research.
The first concerns the complexity of adolescent living arrangements. The
Add Health study design allowed for the construction of retrospective family
structure histories covering birth through adolescence. It is important to note
that these histories likely underreported family change. Adolescents may not
know their family context during their earliest years, or they may not know
enough about the relationships that their mothers and fathers had with their
respective partners to decide whether they were cohabiting or not (Raley &
Wildsmith, 2004), notwithstanding other data limitations already discussed.
Still, these measures reflect adolescent perceptions of their families (White,
1998) and highlight how change and instability within the family define the
experience of a growing proportion of American youth.
Although most adolescents in this sample were born into two–biological
parent families, the family experiences of these young people became much
more complex by adolescence. At Wave 1, more than 40% of young people
resided in a home without both biological parents. Moving beyond single-
parent or stepparent designations, the household roster format revealed that
although most youth in alternative families were residing in mother-only or
married-stepparent families, a nontrivial proportion of young people lived
in father-only families, cohabiting-stepparent families, and other family
forms. Reflecting these shifts in family structure, about a third of the sam-
ple experienced a family transition by Wave 1, with those who experienced
at least one change reporting an average of nearly two family structure
changes. Consequently, these young people spent time adjusting to house-
holds with varying family compositions as well as differing rules and
expectations about behaviors as they underwent the myriad social, psycho-
logical, and biological changes marking childhood and adolescence.
The second theme relates to the linked lives of parents and children, how
timing and stability in family structure shape adolescent adjustment. This
Cavanagh / Family Structure History and Adolescent Adjustment 973
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study operationalized family instability in two ways: instability at different
developmental stages and cumulative family instability. Family structure
timing was operationalized in three ways: family structure at birth, family
structure during adolescence, and time or duration in a family structure
status during adolescence. Beginning with instability, cumulative family
instability was the better predictor of adolescent adjustment and in the case
of current drug use, remained significantly associated with this behavior, net
of a host of family and individual characteristics, including family structure
at adolescence. Turning to the timing of family structure, family structure at
adolescence was most strongly related to both dimensions of adolescent
adjustment, net of cumulative instability plus a host of individual and family
characteristics. For both outcomes, young people in cohabiting-stepparent
families and father-only families were among the most likely to be
emotionally distressed and currently using marijuana.
The other indicators of family structure history also inform our under-
standing of family structure and adolescent behavior. For instance, family
structure at birth was significantly related to adolescent emotional distress.
The persistence of this effect, even after family structure during adolescence
was taken into account, suggests that selection factors (e.g., the social and/or
genetic factors that predict psychological well-being of the mother and father)
play an important role in explaining differences in adolescent emotional distress.
At the same time, the effects of duration in two-parent families—two–biological
parent families as well as married- and cohabiting-stepparent families—on
current marijuana use highlight important aspects of family life. On one hand,
once duration in married-stepparent families was taken into account, adoles-
cents in these families were significantly less likely to use than others, a pat-
tern similar to the one observed in two–biological parent families. On the
other, longer durations in cohabiting-stepparent families were associated with
a higher likelihood of current use when compared to those newer to this
family form. This pattern suggests that duration alone is not what distin-
guishes stepparent and cohabiting-parent families.
A third theme relates to role of family processes in understanding the
link between family structure history and adolescent adjustment. Again,
family structure at adolescence was strongly related to both indicators of
young people’s adjustment. Theory suggests that family structure at this
development stage taps differences in social control processes in the home.
Results from the mediation analyses provide modest support for this explanation,
with, for example, the observed father-only effect on both outcomes reduced once
differences in parental presence and control were taken into account. This
suggests that one reason why young people in this family form do less well
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Cavanagh / Family Structure History and Adolescent Adjustment 975
may be a result of single fathers being less effective or experienced in mon-
itoring their children’s behavior and well-being (Hogan & Kitagawa, 1985;
Maccoby & Martin, 1983; Steinberg, 1990). But overall, the effects of these
social control measures were relative modest.
A more compelling explanation for why family structure matters during
adolescence relates to differences in the quality of relationships and level of
connectedness within the family. The BIC statistics in Model 3 in Tables 5
and 6 are nearly 2.5 times the size the BIC in the best-fitting family struc-
ture history model, suggesting that indicators of family relationships go a
long way in predicting adolescent emotional distress and marijuana use.
Importantly, the associations between family structure and emotional distress
were largely explained by these indicators. For instance, the coefficients asso-
ciated with married- and cohabiting-stepparent families, single-mother
families, and single-father families were reduced by at least half once these
indicators were taken into account. Moreover, the coefficients associated
with stepparent families, married or cohabiting, and single-mother families
were no longer significant at conventional levels. This finding suggests that
family relationships buffer young people in important ways as they navigate
adolescence but that the quality of these ties differs across family structure
statuses in ways that increase the emotional distress of young people.
Although these findings provide strong evidence for the linkage between
parents’ relationship histories and adolescent adjustment, a few caveats are
in order. First, although Add Health’s household roster format provided
leverage in identifying and distinguishing different family structure statuses,
it is still unclear, for instance, whether young people in two–biological
parent households are living with cohabiting or married parents and, more
generally, how issues of recall bias these reports. Another option is to use
longitudinal prospective reports of family structure, collected in a similar
way from both parent and child. Such an approach captures family change
as it is happening and can improve operationalizations of family structure
history. A second caveat relates to omitted variable bias—that is, parents’
mental health history and drug use may be driving the associations among
family structure history, parenting practices and family relationships, and
adolescent emotional distress and drug use. Unfortunately, few measures of
adult well-being were collected in Add Health. Future research using data
sets that include comprehensive measures of both parent and child well-
being (e.g., The NICHD Study of Early Child Care and Youth Development;
National Institute of Child Health and Human Development, 2006) as well
as family structure history can be used to disentangle this effect.
distribution.
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Also worth noting is that the parenting measures in Add Health are
nowhere near as rich as those commonly used in family research. For
instance, indicators that tap parental discipline—or better yet, parents’ own
reports of monitoring, presence in the home, or discipline—may better cap-
ture the social control processes by which family structure is related to ado-
lescent adjustment. Even with better measures, however, the mediational
framework tested here could be reinterpreted. Specifically, the more delete-
rious family processes may not be a consequence of family structure history
but are endogenous with family structure or simply co-occur in families at
greater risk for divorce, remarriage, or cohabitation (Hetherington, Bridges,
& Insabella, 1998). If this is the case, the causal sequences outlined above
are spurious. To disentangle the associations among family structure,
family processes, and adolescent adjustment further, future research should
use prospective longitudinal data that consider pre- and postdisruption mea-
sures of the family context. Such research will better illuminate how the
family shapes adolescent adjustment.
This study attempted to look at an old problem—the significance of
family structure for adolescent adjustment—in a new way by threading
together different theories with a unique data set that allows for detailed
measurement of the family. The findings help to explain the evolution of
families and its implications for the development of children.
Appendix
Emotional Distress Index
How often was each of the following true during the last week?
1. You were bothered by things that usually don’t bother you.
2. You didn’t feel like eating, your appetite was poor.
3. You felt that you couldn’t shake off the blues, even with help from your family
and friends.
4. You felt that you were just as good as other people.
5. You had trouble keeping your mind on what you were doing.
6. You felt depressed.
7. You felt that you were too tired to do things.
8. You felt hopeful about the future.
9. You thought your life had been a failure.
10. You felt fearful.
11. You were happy.
12. You talked less than usual.
13. You felt lonely.
14. People were unfriendly to you.
(continued)
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Cavanagh / Family Structure History and Adolescent Adjustment 977
Appendix (continued)
15. You enjoyed life.
16. You felt sad.
17. You felt that people disliked you.
18. It was hard to get started doing things.
Scale: 0 = never or rarely,1 = sometimes,2 = a lot of the time,3 = most of the time
or all the time.
Please tell me how often you have had each of the following conditions in the past
12 months.
1. Cried frequently?
2. Felt very tired, for no reason?
Scale: 0 = never,1 = just a few times,2 = about once a week,3 = almost every day,
4 = every day.
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