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Does Amount of Time Spent in Child Care Predict Socioemotional Adjustment

During the Transition to Kindergarten?

National Institute of Child Health and Human Development Early Child Care Research Network

To examine relations between time in nonmaternal care through the first 4.5 years of life and children’ssocioemotional adjustment, data on social competence and problem behavior were examined when childrenparticipating in the National Institute of Child Health and Human Development (NICHD) Study of Early ChildCare were 4.5 years of age and when in kindergarten. The more time children spent in any of a variety ofnonmaternal care arrangements across the first 4.5 years of life, the more externalizing problems and conflictwith adults they manifested at 54 months of age and in kindergarten, as reported by mothers, caregivers, andteachers. These effects remained, for the most part, even when quality, type, and instability of child care werecontrolled, and when maternal sensitivity and other family background factors were taken into account. Themagnitude of quantity of care effects were modest and smaller than those of maternal sensitivity and indicatorsof family socioeconomic status, though typically greater than those of other features of child care, maternaldepression, and infant temperament. There was no apparent threshold for quantity effects. More time in carenot only predicted problem behavior measured on a continuous scale in a dose-response pattern but alsopredicted at-risk (though not clinical) levels of problem behavior, as well as assertiveness, disobedience, andaggression.

Over the past 25 years, the United States hasexperienced major changes in childrearing arrange-ments for young children. This transformationstems, in part, from increased maternal employmentassociated with changes in the role women play insociety. In 1975, 34% of mothers with children under6 years of age were in the workforce. In 1999, thecorresponding figure was 61% (National ResearchCouncil and Institute for Medicine, 2000). Even moresignificant have been the changes with regard tomothers of infants. Today, the majority of mothers inthe United States who return to work after having achild do so before their child’s first birthday. Recent

figures (for 1998–1999) indicate that 58% of allwomen with infants under 1 year of age are in thelabor force (Bureau of Labor Statistics, 2000);comparable rates in 1970 and 1985 were 27% and46%, respectively (Kamerman, 2000). In the NationalInstitute of Child Health and Human Development(NICHD) Study of Early Child Care, the over-whelming majority of mothers who were employedin their infants’ first year returned to work andplaced their child in some kind of routine non-maternal care arrangement before the child was 6months of age (NICHD Early Child Care ResearchNetwork, 1997a; see also Hofferth, 1996). Moreover,

r 2003 by the Society for Research in Child Development, Inc.All rights reserved. 0009-3920/2003/7404-0002

This study is directed by a Steering Committee and supportedby National Institute of Child Health and Human Development(NICHD) through a cooperative agreement (U10), which calls forscientific collaboration between the grantees and the NICHD staff.Participating investigators, listed in alphabetical order, are:Virginia Allhusen, University of California, Irvine; Jay Belsky,Birkbeck University of London; Cathryn L. Booth, University ofWashington; Robert Bradley, University of Arkansas, Little Rock;Celia A. Brownell, University of Pittsburgh; Margaret Burchinal,University of North Carolina, Chapel Hill; Susan B. Campbell,University of Pittsburgh; K. Alison Clarke-Stewart, University ofCalifornia, Irvine; Martha Cox, University of North Carolina,Chapel Hill; Sarah L. Friedman, National Institute of Child Healthand Human Development, Bethesda, MD; Kathyrn Hirsh-Pasek,Temple University; Aletha Huston, University of Texas, Austin;Elizabeth Jaeger, St. Josephs University; Deborah J. Johnson,Michigan State University; Jean F. Kelly, University of Washing-ton; Bonnie Knoke, Research Triangle Institute, Research Triangle,NC; Kathleen McCartney, Harvard University; Marion O’Brien,University of Kansas; Margaret Tresch Owen, University of Texas,

Dallas; Chris Payne, University of North Carolina, Greensboro;Deborah Phillips, Georgetown University; Robert Pianta, Uni-versity of Virginia; Suzanne M. Randolph, University of Mary-land, College Park; David Redden, Research Triangle Institute,Chapel Hill, NC; Wendy Robeson, Wellesley College; SusanSpieker, University of Washington; Deborah Lowe Vandell,University of Wisconsin, Madison; Marsha Weinraub, TempleUniversity. We express our appreciation to the study coordinatorsat each site who supervised the data collection, the researchassistants who collected the data, and especially the families andchild care providers who welcomed us into their homes andworkplaces and cooperated willingly with our repeated requestsfor information.Correspondence concerning this article should be addressed to

NICHD Early Child Care Research Network, OEP, Office of theDirector, NICHD, 6100 Executive Blvd. Room 4A01, Rockville, MD20852.

Child Development, July/August 2003, Volume 74, Number 4, Pages 976–1005

the amount of time that children enrolled in theNICHD study spent in nonmaternal care on aweekly basis once it was initiated remained morerather than less stable, with those experiencing 30 ormore hours of care per week beginning in the firstyear of life (i.e., the majority) likely to maintain thishigh level of care throughout their preschool years.Considered together, these data suggest that non-maternal care initiated in infancy for an extensiveperiod each week is now a routine experience formany families in the United States.

How does such early and extensive nonmaternalcare experience affect children’s development, espe-cially their socioemotional adjustment? Despite long-standing debate regarding the effects of nonmaternalcare, reviews of the relevant literature published inthe late 1970s and early 1980s revealed few discern-ible negative associations between early child careand psychosocial adjustment (Belsky & Steinberg,1978; Clarke-Stewart & Fein, 1983). As more evidencebecame available, one series of papers argued thatearly and extensive nonmaternal careFthat is, careinitiated in the first year of life for more than 20 to 30hr per weekFwas associated with elevated levels ofaggression and noncompliance when children were 3to 8 years of age (Belsky, 1986, 1988, 1990, 1994; Belsky& Rovine, 1988). There followed considerable discus-sion of this interpretation of the available data by anumber of scholars (Clarke-Stewart, 1988; Phillips,McCartney, Scarr, & Howes, 1987; Richters & Zahn-Waxler, 1990; Thompson, 1988). Moreover, the emer-gence of additional evidence (e.g., Bates et al., 1994;Belsky & Eggebeen, 1991; Vandell & Corasaniti, 1990)led to the suggestion that it was lots of time in spentin care across the infancy, toddler, and preschoolyearsFthat is, early, extensive, and continuouscareFthat was associated with poorer socioemo-tional adjustment (Belsky, 1994, 2001).

Weaknesses of the data available in the 1980swere numerous and included (a) failure to take intoaccount pre-existing family background factors thatcould account for the association between care useand child outcomes (Richters & Zahn-Waxler, 1990;Thompson, 1988), (b) failure to evaluate the qualityof care (e.g., Belsky, 1984; Goelman & Pence, 1987;Howes, 1990; Howes & Olenick, 1986; McCartney,1984; Phillips, 1987; Phillips et al., 1987; Thompson,1988), and (c) questions as to whether elevated levelsof aggression and noncompliance associated withthe timing and amount of nonmaternal care mightreflect assertiveness and independence from adultsrather than problem behavior (Clarke-Stewart, 1989).

The NICHD Study of Early Child Care waslaunched in the early 1990s as a response to the

need for well-designed research to address questionsconcerning whether experiences in nonmaternalchild care settings are associated with children’sdevelopment. Designed to remedy a number of theproblems inherent in previously collected data, theNICHD study measured children’s experiences withnonmaternal care and their developmental outcomesfrom birth in a diverse sample of families in 10locations across the United States. The NICHD dataare thus well suited to address the followingquestions concerning relations between the use ofnonmaternal care and children’s socioemotionaldevelopment that have been much discussed overthe past 15 years: First, does cumulative amount oftime spent in nonmaternal child care across theinfancy, toddler, and preschool years predict chil-dren’s socioemotional adjustment after consideringpotentially confounding family background factors?Second, do aspects of child care other than cumu-lative amount of time in nonmaternal care, especiallyquality of care (but also type and stability), accountfor detected associations? Third, might the quality ofparenting explain, at least in part, the process bywhich cumulative amount of time in nonmaternalcare is related to children’s social functioning?Fourth, is there a threshold at which the effects oftime in care become particularly pronounced (e.g.,20 hr per week, 30 hr per week?)? Fifth, to whatextent does timing of child care experience, in termsof average hours of care per week across the first 4.5years, account for the detected effects of cumulativequantity of care? And finally, do detected linkagesbetween quantity of care and socioemotional adjust-ment reflect high levels of problem behavior andchildren’s cognitive-linguistic and social-emotionaldevelopment at 54 months of age (NICHD EarlyChild Care Research Network, in press-b).

In addressing each of these questions, theresearch reported herein extends in new anddifferent ways work reported to date by the NICHDEarly Child Care Research Network. By focusingupon the socioemotional adjustment of children at54 months of age and in kindergarten, the currentinquiry provides a longitudinal update of relatedfindings pertaining to children’s self control andadjustment at 24 and 36 months of age previouslyreported in this journal (NICHD Early Child CareResearch Network, 1998; see below). By focussingprimarily, though not exclusively, on issues ofquantity of care, this work complements two otherfocused reports dealing principally with quality ofcare (NICHD Early Child Care Research Network, inpress-a) and with type of care (NICHD Early ChildCare Research Network, submitted) that together

Amount of Time Spent in Child Care 977

form a set of three in-depth, follow-up investigationsof a more general analysis of multiple features ofchild care and children’s cognitive-linguistic andsocial-emotional development at 54 months of age(NICHD Early Child Care Research Network, inpress-b). What is thus new and different about thisstudy is the focus on children before and after entryinto school, along with consideration of issues ofthresholds of quantity of care, timing of care,elevated levels of problem behavior, and distinctionsbetween aspects of socioemotional adjustment (i.e.,aggression, disobedience, assertiveness).

Before detailing the methods and results of thisinquiry, we review evidence linking quantity of timechildren spend in child care, operationalized interms of the early initiation of care, the averageweekly hours spent in care, or both, with children’ssocioemotional adjustment. First, we consider litera-ture related to quantity of care and adjustment inpreschool-aged children and then early-school-agedchildren in the United States, before briefly review-ing findings related to children living outside theUnited States. Next, we consider research examiningthe cumulative quantity of care experienced acrossthe infant, toddler, and preschool years (rather thanduring the first or second year only). Finally, weconsider alternative explanations of research linkingtime spent in child care and socioemotional adjust-ment. The review shows that even though findingsfrom reported studies are by no means perfectlyconsistent, there are grounds for anticipating asso-ciations between cumulative time in child careacross the infancy, toddler, and preschool yearsand socioemotional adjustment, while highlightinga variety of issues that merit further consideration.

Early Child Care and Adjustment in the Preschool Years

Findings from small-sample studies linking in-creased levels of aggression or noncompliance withmore experience in a variety of child care arrange-ments (Crockenberg & Litman, 1991; Rabinovich,Zaslow, Berman, & Hyman, 1987) or center-basedcare (Rubenstein, Howes, & Boyle, 1981; Schwarz,Strickland, & Krolick, 1974) were confirmed in twomuch larger scale survey studies. Baydar andBrooks-Gunn (1991) found, using data on 572(White) children included in the 1986 child supple-ment of the National Longitudinal Study of Youth(NLSY) and controlling for maternal education andintelligence, that maternal employment in the firstyear, and thus reliance on a variety of nonmaternalcare arrangements, predicted increased behaviorproblems in 3- and 4-year-old children. In a second

study, Hofferth (1999) found, using data from thenationally representative Panel Study of IncomeDynamics (n5 519) and controlling for a host ofbackground variables (i.e., child age, gender, age ofhousehold head, family structure, parental employ-ment, parental education, number of children infamily, urban vs. rural residence, race), that enteringany of a variety of child care arrangements duringthe first year of life predicted higher scores on anindex of behavior problems at ages 3 to 4, especiallyaggressive behavior. Furthermore, entry in thesecond year also increased the child’s behaviorproblems, with the association increasing from Year1 to Year 2. Findings held across socioeconomicstrata in both of these large-scale surveys.

Even though the research just cited indicates thatthe early initiation of nonmaternal care is a predictorof problem behavior at preschool age, not allresearch replicates such results. Indeed, some in-vestigations of center-based care have documentedpositive effects of early care experiences (Field, Masi,Goldstein, Perry, & Park, 1988; Howes, 1988;MacCrae & Herbert-Jackson, 1975; McCartney &Rosenthal, 1991; Prodromidis, Lamb, Sternberg,Hwang, & Broberg, 1995). Crockenberg and Litman’s(1991) study of working mothers and childrenexperiencing a variety of nonmaternal care arrange-ments found longer work hours by mothers (i.e.,full-time employment) to be associated with greaterchild compliance at home and in the lab in a small-sample study of 2-year olds (see also Greenstein,1993).

The Early School Years

Although it is clear that associations betweentiming and amount of care and socioemotionaladjustment are variedFand that type of carearrangement does not easily account for the incon-sistency in the databaseFfurther evidence that earlyor extensive child care is associated with problembehavior comes from research on children of schoolage. When Haskins (1985) prospectively followed asample of high-risk, African American children whobegan center-based care in infancy as part of an earlyintervention program, he found them to be morephysically aggressive (i.e., hit, kick, push) askindergarteners than children whose center-basedcare was initiated after the first year of life. Similarresults emerged in Egeland and Heister’s (1995)prospective investigation of children from high-risk,predominantly Caucasian families who experienceda variety of child care arrangements, in that thosewho began child care in their first year were more

978 NICHD Early Child Care Research Network

aggressive and had more externalizing problems inkindergarten, even after controlling for family back-ground factors. Moreover, in first grade, childrenwith early child care experience had higher behaviorproblem scores than agemates without early childcare experience, much like home-reared childrenwith histories of insecure attachment. In bothstudies, child care effects dissipated as children gotolder.

Further evidence of a link between early childcare and problem behavior in the early school yearscan be found in follow-up studies of the effects ofearly maternal employment. Such a literature isrelevant to a discussion of child care effects because,especially in the first years of life, the major reasonnonmaternal care is used is because of mothers’ (andfathers’) employment. Although Harvey (1999)reported that early maternal employment was notrelated to children’s social adjustment in a study ofchildren associated with NLSY, when Han, Waldfo-gel, and Brooks-Gunn (2001) followed up a sub-sample of 138 White children from the NLSYoriginally studied by Baydar and Brooks-Gunn(1991), they observed that resumption of employ-ment by mother before the end of the child’s firstyear, and thus reliance on early nonmaternal care (ofa variety of types), predicted higher levels ofmother-reported externalizing behavior in the earlyschool years. Similarly, Youngblade, Kovacs, andHoffman’s (1999) recent research on 171 eight- tonine-year-olds from maritally intact Caucasian fam-ilies indicated that children whose mothers wereemployed during the first year of life and were invaried routine nonmaternal care arrangements‘‘evinced more acting out, less frustration tolerance,less skill with peers and were nominated more oftenby peers for ‘hitting’ and ‘being mean’ than childrenwhose mothers were not employed’’ (Youngblade etal., 1999, p. 2), though effects on girls becamenonsignificant when concurrent maternal employ-ment was controlled (Hoffman & Youngblade, 1999).When considered collectively, the investigations justreviewed indicate that the timing or amount of earlychild care, or both, have repeatedly, though notalways, been related to problem behavior in theearly school years.

International Studies

Research conducted outside the United States alsohas documented associations consistent with thosealready reviewed between time in, and timing of,early child care and problem-type behavior (e.g.,Bermuda: McCartney, Scarr, Phillips, Grajik, &

Schwarz, 1982; Italy: Varin, Crugnola, Ripamonti, &Molina, 1994; Sweden: Sternberg et al., 1991; but seeAndersson, 1989, 1992, for contrasting evidence).Perhaps most noteworthy given how quantity ofchild care will be examined in the research reportedherein is Borge and Melhuish’s (1995) investigationof a complete cohort of 10-year-olds in one Norwe-gian community. In this work, ‘‘children were ratedby teachers (but not parents) to show higher levels ofproblems when there had been a higher degree ofmaternal employment (i.e., years and hours), henceof nonparental care, in the first four years’’ (p. 37),even after controlling for family social class andchild IQ.

Multiple Years of Care

The Borge and Melhuish (1995) strategy ofexamining care not simply in a single year butacross multiple years was first implemented in theUnited States by Vandell and Corasaniti (1990)studying 236 White third graders from suburbanelementary schools in Dallas, Texas. It is also thestrategy adopted in this investigation because, aswhat follows suggests, it may be lots of time spent innonmaternal care across the entire infancy, toddler,and preschool periods that most consistently pre-dicts noncompliance, aggression, and problem be-havior at school age (Belsky, 1994, 2001). Vandell andCorasaniti found that children whose full-time carewas initiated in their first year and continued untilschool entry were rated by mothers and by teachersas more noncompliant and as getting along morepoorly with peers than other children. These 8-year-olds were also less liked by classmates, according topeer reports, and received the poorest conductevaluations on their report cards. Children whosecontinuous full-time care began in their second yearof life functioned almost as poorly. This may explainwhy Belsky and Eggebeen (1991), who modeledtheir own analysis of the problem behavior of morethan 1,200 four- to six-year-olds included in the 1986NLYS child supplement on those of Vandell andCorasaniti, found that children whose mothers wereemployed full-time (more than 30 hr per week)beginning in their first or second year and continu-ing through their first 3 years were indistinguishablefrom one another. Like Vandell and Corasaniti,however, both sets of children evinced more behaviorproblems than agemates whose mothers worked lessextensively and who thus experienced less nonma-ternal care. This was so even after controlling for birthorder, maternal education, family poverty status,maternal intelligence, and race.


These findings are consistent with those from aninvestigation of 589 kindergartners from three publicschools in Tennessee and Indiana. After controllingfor family background factors, Bates et al. (1994)observed that children who spent more time in anykind of child care during their first 5 years of lifeshowed more negative adjustment (i.e., teacher-rated behavior problems, peer-rated aggression, peerdislike, observed aggression) than children with lesschild care experience. Likewise, Park and Honig(1991) found in their investigation of 105 preschoo-lers in center-based child care that children whobegan full-time nonparental care in the first 9months of life and continued in full-time carethereafter were rated by teachers as, and observedto be, more hostile-aggressive (i.e., fights, destruc-tive, kicks, hits) and noncompliant than agemateswith less intensive and extensive care experience. Ina study of 120 boys from working- and middle-classfamilies, Belsky (1999) reported that more time innonmaternal care across the first 3 and 5 years of lifepredicted more mother-reported externalizing pro-blems when children were, respectively, 3 and 5years old.

It is not the case, however, that all investigationsof multiple years of early child care or maternalemployment detect links with problem behavior.Harvey’s (1999) NLYS investigation of maternalemployment through the first 3 years of life andsocial functioning up to the late elementary schoolyears provided little evidence of risk associated withearly, extensive, and continuous maternal employ-ment. And although the NICHD Study of EarlyChild Care found that more time in nonmaternalcare through the first 2 years of life predicted lesssocial competence and more problem behavior whenchildren were 24 months of age, by the time childrenwere 3 years of age, this was no longer the case(when the predictor was quantity of care throughage 36 months). Indeed, one reason for the currentinquiry is to determine whether the findings thatemerged at age 2 but were gone by age 3 are presentaround the time of the transition to kindergarten.

Explanations of Quantity-of-Care Effects

Should it turn out that quantity effects remerge,even after taking into consideration family back-ground factors (e.g., maternal education, familyincome-to-needs ratio, maternal depression), it willbe necessary to discount or entertain other potentialconfounding or explanatory factors and processes. Itmight be the case, for example, that features of childcare other than quantityFsuch as its quality, type,

and stabilityFcould be responsible for apparenteffects of time spent in nonmaternal care. Alterna-tively, or additionally, it could be that effects ofquantity of care on socioemotional adjustment them-selves derive from effects of child care on parenting.

Other features of child care. Although much of themore recent work published on quantity of care andchildren’s social development has controlled forfamily background factors, almost no researchreporting negative associations between quantity ofearly child care and socioemotional adjustment hassimultaneously examined child care quality. Thus,many of the findings under consideration could be afunction of low-quality care rather than of a highamount of care. Because quality of care is perhapsthe single most consistent predictor of child func-tioning of all features of care examined in theliterature (e.g., Burchinal, Roberts, Nabors, & Bryant,1996; Clarke-Stewart, 1989; Howes, Phillips, &Whitebrook, 1992; Howes, Smith, & Galinsky, 1995;Lamb, 1998), it is imperative to determine whethereffects of quantity of care remain when quality ofcare is taken into consideration.

The same holds for (in)stability of care, as it isconceivable that changes in child care arrangementscould engender uncertainty about the continuingavailability of any nonmaternal care provider,undermine the child’s emotion regulation, or both.Consistent with this possibility is evidence thatfamiliar and stable caregivers or care arrangementsmay reduce the stress experienced by infants(Cummings, 1980, 1986) and the likelihood of thechild developing an insecure attachment to his orher mother (Suwalsky, Zaslow, Klein, & Rabinovitch,1986; see also NICHD Early Child Care ResearchNetwork, 1997b). Therefore, we include as an indexof instability of care the number of different childcare arrangements experienced through the first 3years of life (i.e., the operational definition ofinstability) when testing whether effects of quantityremain after taking into consideration other featuresof child care.

Because type of child care might also be respon-sible for quantity of care effects, two ways ofmeasuring type of care are used when determiningwhether quantity of care effects remain significantafter taking into consideration other features of childcare. Findings on aggression from experimentalinvestigations of high-quality center-based programs(Haskins, 1985; Schwarz et al., 1974) seem to lendsupport to the hypothesis that it is specifically timespent in large-group settings that is responsible for


associations between quantity of care and problemoutcomes. Alternatively, it may not be center-basedcare per se, but peer-group experience beginning atan early age that is responsible for the quantity ofcare effects under consideration. In light of thesepossibilities, in this research we evaluate quantity ofcare effects after taking into consideration experi-ence in center-based care and in any nonmaternalcare arrangement in which at least three children arepresent. By operationally defining peer-group ex-posure in this way, we identify children whoexperienced something akin to a ‘‘peer group’’ evenwhen not in a child care center.

Parenting quality. What transpires in the familymay also account for why time in child care could beassociated with noncompliance, aggression, andproblem behavior. Both theory and evidence suggestthat parenting or, more specifically, the sensitivity ofmothers in interacting with their children couldcontribute to the association between quantity ofcare and socioemotional adjustment. Brazelton(1986), Sroufe (1988), and Belsky (1999, 2001) havesuggested that routine nonmaternal care, particu-larly when initiated very early in life, could under-mine a mother’s sensitivity toward her infant byreducing the amount of time available for the motherto learn the baby’s signaling patterns and behavioralrhythms. Consistent with this argument is evidencefrom the NICHD study showing that more time incare predicts less harmonious mother–infant inter-action and less sensitive mothering at 6, 15, 24, and36 months of age, even when quality of child careand family selection variables are controlled(NICHD Early Child Care Research Network,1999). Also consistent with this reasoning arefindings from Belsky’s (1999) study showing thatnot only did more time in care predict more negativemothering during the child’s second and third yearsof life but that controlling for these effects in amediational analysis eliminated significant linkagesbetween lots of time in care and elevated levels ofmother-reported externalizing problems at 3 and 5years of age. The fact, however, that other studiesfailed to find effects of amount of care on parenting(Braungart-Reiker, Courtney, & Garwood, 1999;Burchinal, Bryant, Lee, & Ramey, 1992; Goldberg &Easterbrooks, 1988; Zaslow, Pedersen, Suwalsky, &Rabinovich, 1989) raises questions about the poten-tial role of parenting with respect to the negativeeffects of lots of time in care on socioemotionaladjustment. In this inquiry we address this issue byexamining maternal sensitivity and hypothesize thatcontrolling for this factor will attenuate any detectedeffects of quantity of child care.

Research Questions

On the basis of the literature reviewed, theresearch presented herein was designed to deter-mine the following: (a) whether amount of timespent in nonmaternal child care through the first 54months of life is related to socioemotional adjust-ment after controlling for family selection effects; (b)whether any such findings pertaining to quantity ofcare can be accounted for by other aspects of childcare experience (i.e., quality, stability, type, peer-group exposure); (c) whether maternal sensitivitymight be a mechanism by which quantity of carecomes to predict socioemotional adjustment; (d)whether quantity-of-care effects are especially ap-parent when some threshold of time in care isreached; (e) whether timing of child care experience,in terms of average hours of care per week across thefirst 4.5 years, accounts for detected effects ofcumulative quantity of care; and (f) whetherdetected linkages between quantity of care andsocioemotional adjustment pertain to high levels ofproblem behavior or actual aggressive behaviorrather than assertiveness.

Method

Overview of Study Design

Children at 10 geographic sites were followedfrom birth to kindergarten. Mothers were inter-viewed in person when infants were 1 month old.When children were 6, 15, 24, 36, and 54 months old,we assessed the home and family environment and,for those in nonmaternal child care, we observed thechild care setting. Mothers were telephoned every 3or 4 months in the intervening periods to updatereports on child care use. Children’s cognitive skillsand social behavior were assessed at 15, 24, 36, and54 months of age and in the fall of the kindergartenyear. Parent, caregiver, and kindergarten teacherreports of children’s social competence and behaviorproblems at 4.5 years and in kindergarten are thefocus of this report.

Participants

Families were recruited through hospital visits tomothers shortly after the birth of a child in 1991 in 10locations in the United States. During selected 24-hrintervals, all women giving birth were screened foreligibility and willingness to be contacted again. Ofthe 8,986 mothers who gave birth during thesampling period, 5,416 (60%) agreed to be tele-phoned in 2 weeks and met the eligibility require-


ments (mother over 18 and spoke English, motherhealthy, baby not multiple birth or released foradoption, live within an hour of research site,neighborhood not too unsafe for teams of research-ers to visit). Of that group, a conditionally randomsample of 3,015 was selected (56%) for a 2-weekphone call; the conditioning assured adequaterepresentation (at least 10%) of mothers withoutpartners, mothers without a high school degree, andethnic mothers. At these calls, families were ex-cluded if the infant had been in the hospital morethan 7 days, the family expected to move in the next3 years, or the family could not be reached with atleast three attempts at contact. A total of 1,526 whowere selected for the call were eligible and agreed toan interview; of these, 1,364 completed a homeinterview when the infant was 1 month old andbecame the study participants. These 1,364 familieswere similar on years of maternal education,percentage in different ethnic groups, and presenceof partner in home to the eligible hospital sample.The resulting sample was diverse, including 24%ethnic minority children, 11% mothers not complet-ing high school, and 14% single-parent mothers.

At 4.5 years, 1,081 children and their parentscontinued to be enrolled in the study. Mothers hadan average of 14.4 years of education, and 15% werewithout a partner; average family income was 3.6times the poverty threshold; and 79% of the childrenwere European American, non-Hispanic. The parti-cipants differed from the 283 children who wererecruited but were lost to follow-up. Mothers ofparticipants had significantly (po.05) more educa-tion (M5 14.4 years vs. 13.6 years) and were morelikely to have a husband or partner in the household(85% vs. 77%); The children were less likely to beAfrican American, non-Hispanic (11% vs. 17%); andthe families had higher family incomes as deter-mined by their income-to-needs ratio (M5 3.6 vs.3.4). (The income-to-needs ratio is an annuallyadjusted, per capita index comparing householdincome with federal estimates of minimally requiredexpenditures for food and shelter. An income-to-needs ratio of 1.0 is the U.S. government definitionof poverty; therefore, a ratio of 3.0 represents a percapita income 3 times the poverty level.)

In kindergarten, 1,058 children and their parentscontinued to be enrolled in the study. Mothers hadan average of 14.4 years of education, and 15% weresingle; average family income was 3.7 times thepoverty threshold; and 79% of the infants wereEuropean American, non-Hispanic. The participantsdiffered from the 306 children who were recruitedbut were lost to follow-up. Mothers of participants

had significantly (po.05) more education (M5 14.4years vs. 13.5 years) and higher family incomes(income-to-needs ratio: M5 3.7 vs. 3.3), and weremore likely to have a husband or partner in thehousehold (85% vs. 77%). The children weresignificantly less likely to be African American,non-Hispanic (11% vs. 19%).

Despite the selective loss of more minority and at-risk families over time (i.e., lower income, lesseducation), the sample was by no means a simpleWhite, middle-class one. With respect to the retainedfamilies, 27% of mothers had no more than a highschool education (at time of enrollment), 25% hadincomes no greater than 200% of the poverty level,and 20% were minority (i.e., not non-HispanicEuropean American).

Longitudinal Modeling of Repeated Measures

Before detailing the methods and measures usedin this study, note should be taken of the fact that asmany, but not all, measures were obtained more thanonce, hierarchical linear modeling (HLM) techniqueswere often used to generate two scores to describepatterns of change over time from measurements ofthe same construct obtained by the same method onmultiple occasions. HLM is useful in describingpatterns of change over time for each individual, andit provides more precise and interpretable indexes ofdevelopmental patterns than more traditional meth-ods (Bryk & Raudenbush, 1992). The HLM analysesestimated individual growth curves, and we selectedtwo measures of individual trajectories from theindividual growth curves. One measure, the max-imum-likelihood estimate of the intercept, reflectsthe predicted mean of the repeated measurements ofa particular construct at the mean of the includedassessment ages (i.e., 6, 15, 24, 36, and 54 months).Age was mean centered; therefore, intercepts reflectthe predicted mean for the individual’s growthcurve at the average age (27 months). The secondmeasure, the individual’s slope, estimates the linearchange in the construct across the multiple measure-ment occasions. Positive slope values reflect degreeof increase in the variable over time, whereasnegative slope values reflect degree of decrease inthe variable over time. Thus, a child whose averagenumber of hours spent in nonmaternal care in-creased over time would have a positive slope forthe variable quantity of care, whereas one whosehours per week in care decreased over time wouldhave a negative slope. The use of HLM to createsummary scores on repeatedly measured constructsreduces the autocorrelation and error misestimation


associated with simple averages. A full descriptionof the method can be found in Bryk and Raudenbush(1992).

Child Care Characteristics

Consistent with all other reports from the NICHDStudy of Early Child Care and many other studies ofchild care (e.g., Casper, 1996; Kisker & Maynard,1991; Riley & Glass, 2002; Veum & Gleason, 1991),nonmaternal child care was defined as regular careby anyone other than the mother, including care byfathers, relatives, and nannies (whether in home orout of the home), family day care providers, andcenters. Several features of individual children’s careexperiences measured from birth through 54 monthsfigure importantly in this report.

Quantity. Cumulative amount of time in nonma-ternal care through the first 4.5 years of life wasdetermined from telephone interviews with mothersat 3- or 4-month intervals about the number of hoursand the types of care used during the prior 3 to 4months. Two indexes were used in the analyses: theaverage number of hours per week that nonmaternalcare was used from ages at the mean age between 3months through 4.5 years (i.e., HLM intercept) andthe linear rate of change (i.e., HLM slope) of hoursper week over time.

Quality. Quality was defined by the caregiver–child interaction and stimulation experienced by thetarget child in the child care setting. Observationalassessments were obtained at ages 6, 15, 24, and 36months of age for children who were in nonmaternalcare for 10 or more hours per week. At 54 months,the criterion for observation was being in care for 7.5or more hours per week (because many childrenwere enrolled in preschool programs 3 days perweek for 2.5 hours per day). Quality was assessedduring two half-day visits scheduled within a 2-week interval at 6 to 36 months and one half-dayvisit at 54 months.

At each visit, observers completed two 44-mincycles of the Observational Record of the CaregivingEnvironment (ORCE), during which they first codedthe frequency of specific caregiver behaviors andthen rated the quality of the caregiving. Positivecaregiving composites were calculated for each agelevel observed by averaging these ratings. At 6, 15,and 24 months, the positive caregiving compositewas based on the mean of five 4-point ratings:sensitivity to child’s nondistress signals, stimulationof child’s development, positive regard towardchild, detachment (reflected), and flatness ofaffect (reflected). Cronbach’s alphas for the compo-

site were .89 at 6 months, .88 at 15 months, and .87at 24 months. At 36 months, these same fiveratings plus two additional ratingsFfosters child’sexploration and intrusive (reflected)Fwere in-cluded in the composite. Cronbach’s alpha was .83.At 54 months, the positive caregiving composite wasthe mean of 4-point ratings of caregivers’ sensitivityand responsivity, stimulation of cognitive develop-ment, intrusiveness (reflected), and detachment(reflected). Cronbach’s alpha was .72. For this report,two cumulative indexes of positive caregiving wereformed by HLM analyses using scores for all periodsin which a child’s care settings were rated: averagequality of nonmaternal care at the mean age (i.e.,HLM intercept) and linear rate of change over timein quality of nonmaternal care (i.e., HLM slope).

To ensure that observers at the 10 sites weremaking comparable ratings, all observers werecertified before beginning data collection. Thecertification test at each age consisted of six 44-minvideotapes that had been master-coded by experts.Exact agreement with the master codes at 60% orbetter was required. To prevent observer drift, allobservers took two additional coding tests duringthe 10 months of data collection at each ageassessment; a criterion of 60% exact agreement wasused to allow continued data collection. In addition,observer agreement was assessed during live,on-site observations. At each site, all possible pairsof observers were required to visit both home-basedand center-based child care.

Reliability estimates for the positive caregivingcomposite score were computed for both the master-coded videotapes and live observations using Pear-son correlations and the repeated-measures ANOVAformulation described in Winer (1971,p. 287). Relia-bility exceeded .90 at 6 months, .86 at 15 months, .81at 24 months, .80 at 36 months, and .90 at 54 months.

At least one child care quality assessment wasobtained for 91% of the sample of children studied at4.5 years of age (985 of 1,081); at least twoassessments were obtained for 779 children. Of the1,064 children, 697 were observed for all eligibleperiods; 232 children were not observed in aneligible setting at one age; 70 children were notobserved at two ages; 34 children were not observedat three ages; and 16 children were not observed atfour ages. Thirty-four children were never in regularnonmaternal care for 10 or more hours a week.

Type. For each of 16 epochs (3-month intervalsfrom birth to 36 months and 4-month intervals after36 months), the child’s primary care arrangementwas classified as a center, a child care home (anyhome-based care outside the child’s own home


except grandparent care), in-home care (by anycaregiver in the child’s own home except father orgrandparent), grandparent care, or father care.Information was available on each setting withrespect to the number of children present other thanthe target child. Epochs in which children were innonmaternal care less than 10 hours per week werecoded as exclusive maternal care. Two indicators oftype of care are used in this report. Center carereflects the proportion of epochs in which the childreceived care in a center, and peer-group exposurereflects the proportion of epochs in which a childwas in any child care setting in which there were atleast two other children in addition to himself orherself.

Instability. For each of twelve 3-month epochsfrom birth to 36 months, a count was made of thenumber of times the mother reported that the childstarted a new child care arrangement, or one thathad previously stopped and then started again.Because of a few extreme scores, this variable wastruncated at the 95th percentile. Children whoreceived no nonmaternal care received scores of 0.Thus, low scores reflected more stable arrangementsand high scores reflected less stable arrangements.

Maternal, Child, and Family Characteristics

The following maternal, child, and family char-acteristics were included in the analyses as controlsfor selection effects: maternal education in years, themean family’s income-to-needs ratio (family incomedivided by the poverty threshold for its householdsize), partner status (the proportion of 3- to 4-monthepochs during which mother reported living with apartner or husband), child sex, infant temperament(see the following discussion), ethnic group (non-Hispanic African American, non-Hispanic EuropeanAmerican, Hispanic, or other), and maternal depres-sive symptoms (the intercept or predicted mean andlinear slope over 4 years) as measured by the Centerfor Epidemiological Studies Depression Scale (CES–D; Radloff, 1977). The mother reported the familyincome and household size at 6-, 15-, 24-, 36-, and 54-month assessments. The family income was dividedby the income determined by the U.S. Census as thepoverty threshold for a familly of that size. Forexample, an income of $15,455 in 1995 for a family offour would have an income-to-needs ratio of 1.

Infant temperament was measured by means offifty-five 6-point items from the Infant TemperamentQuestionnaire completed by mothers at 6 months(Medoff-Cooper, Carey, & McDevitt, 1993). Thecomposite measure reflected difficult temperament

and was created by calculating the mean of thenonmissing items with appropriate reversal ofscales, so that numerically large scores consistentlyreflected a more ‘‘difficult’’ temperament. Thus,higher scores reflected a more negative disposition,that is, a child seen as more intense, less positive inmood, and less adaptable to daily routines. Cron-bach’s alpha was .81.

In addition, a composite measure of maternalsensitivity, based on evaluations of observed mater-nal behavior at 6, 15, 24, 36, and 54 months (NICHDEarly Child Care Research Network, 1999) served asa potential explanatory factor, rather than controlvariable, in the analyses to be reported. Mother–child interaction was videotaped in semistructured15-min observations at each age. The observationtask at 6 months had two components. In the first 7min, mothers were asked to play using any toy orobject available in the home or none at all; for theremaining 8 min, mothers were given a standard setof toys they could use in play with their infants. At15, 24, 36, and 54 months, the observation proce-dures followed a three-boxes task in which motherswere asked to show their children age-appropriatetoys in three containers in a set order (see Vandell,1979). The mother was instructed to have her childplay with the toys in each of the three containers andto do so in the order specified.

At each age a maternal sensitivity composite wasconstructed based on three ratings. At 6, 15, and 24months it comprised the sum of three 4-pointratings: sensitivity to nondistress, positive regard,and intrusiveness (reversed). At 36 and 54 months,three 7-point ratings were composited: supportivepresence, respect for autonomy, and hostility (re-versed). Two cross-age composite scores werecreated, after standardizing scores at each age, usingHLM procedures, predicted mean maternal sensi-tivity at the average age (i.e., intercept), and linearrate of change over time (i.e., slope).

Tapes from all research sites were shipped to asingle site for coding. Coders were blind as to otherinformation about the families. Intercoder reliabilitywas determined by assigning two coders to 20% ofthe tapes randomly drawn at each assessmentperiod. Intercoder reliability was calculated as theintraclass correlation (Winer, 1971), which rangedfrom .83 to .87 for the maternal sensitivity compo-sites over time.

Child Adjustment at 4.5 Years and Kindergarten

The child outcomes examined in this study weremother-, caregiver-, and teacher-reported measures


of social competence and problem behavior, andchild–teacher relationship conflict obtained whenchildren were 54 months of age or in kindergarten.In addition, at 54 months, children were observed ina dyadic interaction with a friend, and their peerinteractions in child care settings were observed.

Social competence. Mothers completed the SocialSkills Questionnaire from the Social Skills RatingSystem (SSRS; Gresham & Elliott, 1990) for theirchildren at both times of measurement. Teacherscompleted the instrument when children were inkindergarten. This instrument is composed of 38items describing child behavior, each rated on a 3-point scale reflecting how often the child exhibitedeach behavior. Items are grouped into four areas:cooperation (e.g., keeps room neat and clean withoutbeing reminded), assertion (e.g., makes friendseasily), responsibility (e.g., asks permission beforeusing someone else’s property), and self-control(controls temper when arguing with other children).The total score used in this report represents the sumof all 38 items, with higher scores reflecting higherlevels of perceived social competence (alpha5 .93for kindergarten teacher; .86 and .88 for mothers at54 months and kindergarten, respectively). The SSRSwas normed on a diverse, national sample ofchildren in the 3- to 5-year age range and showedhigh levels of internal consistency (median5 .90 )and test–retest reliability (.75 to .88) and moderateconcurrent and predictive validity to other indexesof social competence.

For children who were in child care at least 7.5 hrper week at age 54 months (n5 833), caregiverscompleted a slightly modified version of theCalifornia Preschool Social Competency Scale(CPSC; Levine, Elzey, & Lewis, 1969). The CPSC isa 30-item instrument assessing a range of socialcompetencies especially relevant in settings wherethere are other children (e.g., safe use of equipment,using names of others, greeting new child, initiatinggroup activities). Four items were added to indexspecific features of peer play (cooperation, followingrules in games, empathy, and aggression). All itemswere rated on 4-point scales. Items scored as notapplicable were set as missing. The total score forsocial competence was the sum of the 34 items, withhigher scores denoting greater social competence(alpha5 .88).

Behavior problems. Mothers, caregivers, and tea-chers completed appropriate versions of the ChildBehavior Checklist (CBCL; Achenbach, 1991), awidely used measure of behavior problems. TheCBCL 4–18y, completed by mothers at 54 monthsand in kindergarten, lists 113 problem behaviors; the

parent rates each as not true (0), somewhat true (1),or very true (2) of her child. At 54 months, the TRF2–5y was completed by caregivers, and at kinder-garten age, the 100-item TRF 5–18y was completedby teachers. Reliability and validity of these instru-ments are well established (Achenbach, 1991). Thetotal problem behavior score served as the depen-dent variable in the primary analysis. Raw scoreswere converted into standard T scores, based onnormative data for children 4 to 11 years of age. Insecondary analyses designed to address the questionof whether quantity of child care was related toaggression and disobedience or just assertiveness,three composite measures were created by summingselect items from the CBCL. An assertivenesssubscale was based on items such as bragging andboasting, talks too much, argues a lot, and demandsattention. A disobedience subscale was based onitems such as defiantFtalks back to staff, disruptsclassroom discipline, temper tantrums, lying andcheating, and fails to carry out assigned tasks. Theaggression subscale was based on items such ascruelty to others, destroys own things, gets in manyfights, attacks others, hits others, and explosive andunpredictable. The decision was made not to use theformal aggression subscale score from the CBCLbecause it includes items from all the subscales justdescribed (i.e., items reflecting disobedience andassertiveness, as well as aggression).

Teacher–child conflict. At 54 months and at kinder-garten, child care providers and teachers completedthe StudentFTeacher Relationship Scale (STRS;Pianta, 2001). The STRS is a widely used indicatorof teachers perceptions of the quality of theirrelationship with a specific child, and one specificscale, Conflict, appears to be a particularly informa-tive indicator (e.g., Birch & Ladd, 1997; Hamre &Pianta, 2001). Teacher-reported conflict in kinder-garten is predictive of later disciplinary problems inschool, controlling for pre-existing problem behavior(Hamre & Pianta, 2001) and is related to poorerschool achievement and problems in peer relations(Birch & Ladd, 1997; Hamre & Pianta, 2001). Assuch, teacher–child conflict reflects a unique featureof the child’s adjustment to the school (or preschool)setting.

In the present sample, caregivers completed theSTRS at 54 months for the children in child care, andkindergarten teachers completed the STRS in the latefall of kindergarten. Coefficient alphas for theConflict scale were .86 at 54 months and .90 atkindergarten.

Dyadic peer interaction. At 54 months, studychildren were videotaped while they interacted with


a peer during three structured play episodes, andtrained observers coded 10 ratings of social behaviorfor each episode. The first episode involved acartoon mouse pop-up game, the second used asingle hand-held picture-viewing apparatus, and thethird used a doctor kit. Ratings were made using 5-point Likert scales for behavior in each episode, thenaveraged across episodes for each scale. Scalesindicating negative behaviors were recoded into 3-point scales because of the low frequency ofoccurrence. Two composite indicators of the qualityof interactions with the peer were computed for eachchild. Negative interaction was computed as thesum of ratings on ‘‘contributes to negative interac-tion,’’ ‘‘aggression,’’ and ‘‘negative mood.’’ Positiveinteraction was computed as the sum of ‘‘contributesto positive interaction,’’ ‘‘prosocial behavior,’’ and‘‘positive mood.’’ Coefficient alpha was .75 for thepositive interaction composite and .67 for thenegative interaction composite. The negative dyadicinteraction composite was highly skewed and hadlimited nonzero values; therefore, it was dichoto-mized as 0 if below the median (–.94) and 1otherwise.

Behavior in child care. As part of the ORCE childcare observation at 54 months, the interactions of thestudy child with peers in the child care setting wererecorded during each of the observation periods (seeprevious description). Two composites were derivedfrom time-sampled codes of the child’s behavior.Children were observed across six 10-min segments,with each 10 min consisting of ten 30-s observe and30-s record intervals. Observers were trained toreliability based on agreement with precoded video-tapes. Observers were required to reach exactagreement on 60% of intervals coded across 18 tapedobservations before being certified to collect data inthe field. Once in the field, observers were alsorequired to conduct two live reliability observationswith a second observer. Estimates of live agreementfor individual time-sampled codes (correlations)ranged from .58 to .99.

The child care positive composite was computed asthe sum of three codes: cooperative play, boisterousplay, and other positive or neutral interaction. Thechild care negative composite was computed as thesum of four codes: physical aggression, verbalaggression, negative behavior toward peer, and othernegative nonaggressive acts. The negative interactioncomposite was highly skewed and had limitednonzero values; therefore, it was dichotomized as 0if below the median (2) and 1 otherwise.

Correlations were obtained for concurrent associa-tions within the sets of 54-month and kindergarten

outcomes and for cross-time associations betweenanalogous measures at both periods (all correlationsreported are significant at po.05). For the 54-monthoutcomes, the highest associations were betweencaregiver reports of social competence and externa-lizing problems (r5 –.63), social competence andconflict (r5 –.50), and externalizing problems andconflict (r5 .68). Mother-reported social competenceand externalizing problems were moderately asso-ciated (r5 –.32). The remaining concurrent associa-tions among dependent measures at 54 months werelow. At kindergarten, the highest associations wereagain between teacher reports of social competenceand externalizing problems (r5 –.56), social compe-tence and conflict (r5 –.55), and externalizingproblems and conflict (r5 .75). Mother-reportedsocial competence and externalizing problems weremoderately associated (r5 –.42). Over time, correla-tions were as follows: mother-reported social com-petence (r5 67), mother-reported problems (r5 .72),teacher-reported social competence (r5 .28), teacher-reported externalizing problems (r5 .40), and tea-cher-reported conflict (r5 .34). Thus, for severaloutcome measures at both periods there wasmoderate overlap, and for similar measures acrossperiods there was also moderate overlap. Because ofthese associations, the findings reported are dis-cussed in relation to the overall pattern of results,recognizing that these multiple indicators of socialfunctioning share variance.

The mean, standard deviation, and ranges of allvariables are presented in Table 1. Overall, mothers,caregivers, and teachers rated the sample wellwithin the normal range on all standardizedmeasures. The mean CBCL externalizing T scoreswere approximately 50 at both time points accordingto mothers and teachers, with 50 being the mean forthe norming population. Few children scored in theclinical range, defined as a T score of 65 or above,and about 16% of the same scored in the at-riskrange, defined as a T score of 60 or above (1 SDabove the mean). The SSRS social competence meanscores ranged from 99 to 103, again falling close tothe expected population mean of 100. Further detailsabout all data collection procedures are documentedin Manuals of Operation of the study (http://public.rti.org/secc/).

Results

Table 1 presents descriptive statistics on all variablesused in the analyses. Most of the sample entered intoat least part-time nonmaternal care early in theirlives. By 3 months of age, almost half of the sample


Table 1

Descriptive Statistics on All Analytic Variables

Variable M SD Minimum Maximum N

Quantity of child care (hours)

Mean hours per week (3–54 months) 27.0 15.4 � 0.5 61.6 982

Linear change/slope (3–54 months) 0.2 0.3 � 1.1 1.2 982

Mean child care hours 3–6 months 21.0 18.0 0 62.5 982





Child and family characteristics

Gender (15male) 50% 982

Ethnicity:

Black 11% 982

Hispanic 6% 982

Other 4% 982

Mother’s education 14.5 2.4 7.0 21.0 982

Income-to-needs ratio (6–54 months) 3.8 2.8 0.2 27.4 982

Child temperament at 6 months 3.2 0.4 1.5 4.4 982

Depression mean (i.e., HLM intercept: 6–54 months) 9.2 5.4 1.4 31.7 982

Depression linear slope (6–54 months) 0.2 0.7 � 2.8 4.1 982

Parenting mean (i.e., HLM intercept: 6–54 months) 0.1 0.6 � 2.9 1.3 982

Parenting linear slope (6–54 months) � 0.0 0.1 � 0.3 0.2 982

Other child care experiences

Center care: proportion time 0.2 0.3 0 1.0 982

Quality mean (HLM intercept: 6–54 months) 2.8 0.2 2.1 3.4 982

Peer group exposure: proportion time 0.6 0.3 0 1.0 982

Instability (3–34 months): number of changes 10.8 4.2 0 18.0 982

Child Social Functioning

Mother ratings: 54 months

Social competence 98.7 13.3 53.0 130.0 935

Externalizing problems 51.6 9.4 30.0 82.0 940

Caregiver ratings: 54 months



Conflict 18.8 6.6 12.0 49.0 691

Observations: 54 months

Dyadic play positive 0.1 2.5 � 8.4 7.9 698

Dyadic play negative (4 median) 0.4 0.5 0 1.0 694

Child care positive (log)a 2.3 1.0 0 4.0 838

Child care negative (4 median) 0.5 0.5 0 1.0 838

Mother ratings: kindergarten



Teacher ratings



Conflict 10.7 5.4 7.0 34.0 905

Mother ratings: 54 months

Aggression items 1.3 1.9 0 12.0 931

Disobedience items 2.9 2.1 0 14.0 934

Assertive items 3.7 2.1 0 10.0 936

Caregiver ratings: 54 months

Aggression items 1.3 2.4 0 14.0 593

Disobedience items 1.9 2.6 0 14.0 676

Assertive items 0.4 0.6 0 2.0 698


was being cared for by someone other than themother for at least 10 hr per week. Almost threefourths of the children were in at least part-time careby their first birthday. On average, children wereexperiencing 21 hr of child care between 3 and 6months, with increases in average hours of care perweek across early childhood. Between 37 and 54months, children were experiencing an average of32.7 hr of routine nonmaternal care per week. Notethat the average hours per week that children werein nonmaternal child care from 3 to 54 monthsranged from 0 to almost 62 hr, with a mean of 27 hrper week and a standard deviation of almost 15 hr.However, more than half the children experiencedchanges in hours of care per week of at least 15 hrper week across adjacent reporting periods. At thesame time, cross-time stability coefficients show thatchildren who experienced lots of time in nonmater-nal care at one point also experienced lots of time incare at other periods. Indeed, the cross-age correla-tion in average hours per week in care ranged from ahigh of .81 (po.001) between the third and fourthyear to a low of .46 (po.001) between mean hours ofcare in the first and fourth years. Overall, childrenshowed modest increases in hours of care across the51-month period. We chose indexes from individualgrowth curve analyses to describe individual differ-ences in child care experience and they serve as theprimary quantity-of-care predictors of child out-comes in this investigation. The hours intercept(M5 27.0, SD5 15.4) represents the predicted meanhours of care experienced at the mean age between 3and 54 months (i.e., HLM intercept). The hoursslope (M5 .17 hr per month, SD5 18) describesthe extent to which hours increased (i.e., positivevalue) or decreased (i.e., negative value) linearlyover time.

Statistical analyses were undertaken to answer aseries of questions posed in the introduction andsome additional questions that emerged during thedata analysis. Fourteen outcomes were evaluated,

and we used the consistency of findings across themultiple indicators of social functioning as thecriterion for determining when an effect wasdetected. Rather than relying on a single regressionanalysis that would include in a single predictionmodel all the explanatory variables considered inthis inquiry, we analyzed the data in a series ofinterrelated steps (i.e., nested regression models) sothat we could address distinctive and logicallyordered questions in a lawful sequence. Thus, wefirst asked whether quantity of care predictedmultiple indexes of socioemotional adjustment whencontrolling for a substantial number of familybackground factors that past work has shown to beconfounded with use of child care. Next, we askedwhether initially detected quantity effects remainedsignificant or were attenuated when other features ofchild care that might account for quantity effectswere taken into consideration (i.e., quality, type,instability) and, thereafter, whether the same wastrue when maternal sensitivity was added to theprediction model. Results from these analysesstimulated the search for thresholds at which hoursof care per week might have especially pronouncedeffects and for evidence that effects of time in caremight be more or less pronounced during particularperiods such as the first 6 months or the first 2 yearsof life. Finally, we sought to determine whetherquantity of care predicted high levels of problembehavior and, in particular, aggression and disobe-dience rather than just assertiveness.

Effects of Quantity Controlling for Background Factors

To evaluate the relation between time in non-maternal child care and socioemotional adjustment,an initial multiple regression model (i.e., Model 1)was estimated. This base model was applied to eachof nine dependent variables measured at 54 monthsand five measured at kindergarten to determine the

Mother ratings: kindergarten

Aggression items 1.2 1.8 0 13 924

Disobedience items 2.7 2.2 0 13 925

Assertive items 3.4 2.1 0 10 926

Teacher ratings: 54 months

Aggression items 0.7 1.8 0 16 891

Disobedience items 1.8 2.9 0 20 880

Assertive items 1.4 2.12 0 10 894

aLog transformation.

Table 1 Continued

Variable M SD Minimum Maximum N


effects of two indicators of quantity of child careafter controlling for family background factors. Thetwo indicators of child care quantity represented thepredicted mean number of hours per week of careexperienced by that child at the mean age between 3and 54 months of age (i.e., HLM intercept) and linearchange over time for that child (i.e., HLM slope) inhours of care per week across the same develop-mental period. The covariates included in the basemodel to control for background factors were site(represented by nine dummy-coded variables), childgender (male), child ethnicity (African Americannon-Hispanic, Hispanic, other), 6-month difficulttemperament, maternal education at enrollment,average income-to-needs ratio from 6 to 54 months,and average maternal depression (i.e., HLM inter-cept) and linear change over time in maternaldepression (i.e., HLM slope) across the same period.As analyses revealed that significant interactionsbetween family and child care factors, and amongchild care factors, were detected at a rate no greaterthan chance, no interaction effects are presented,consistent with standards of reporting adopted bythe NICHD Study of Early Child Care.

The effects of the two quantity of care predictorsin the base model, mean hours of care per week, andlinear change in hours per week are detailed underthree columns in Table 2, each labeled ‘‘Model 1:Base.’’ The first such column (i.e., fourth column oftable) indicates whether the two quantity variables,when considered together, significantly predicted aparticular outcome (i.e., ‘‘pooled’’). Inspection of therelevant data shows that the two quantity variablessignificantly predicted three of the nine 54-monthoutcomesFcaregiver-reported social competence,externalizing problems, and conflictFand three ofthe five kindergarten outcomesFmother-reportedexternalizing problems and teacher-reported exter-nalizing problems and conflict. The second columnlabeled ‘‘Model 1: Base’’ (i.e., seventh column in thetable) presents the effect of the individual quantitypredictor, mean hours per week of care, net of allother factors in the base model; the third columnlabeled ‘‘Model 1: Base’’ (i.e., tenth column) presentsthe same information for the second quantitypredictor, linear change over time in hours per weekin nonmaternal care.

Inspection of the relevant data indicates thatchildren who averaged more time in nonmaternalchild care across their first 54 months of life scoredlower on caregiver-rated social competence andhigher on caregiver-rated externalizing problemsand caregiver–child conflict at 54 months, and wereobserved to engage in more negative dyadic play at

this age. More time spent in child care alsopredicted, at kindergarten age, more mother- andteacher-reported externalizing problems and moreteacher–child conflict. Furthermore, when hoursspent in child care increased (vs. decreased) overtime, children were rated by caregivers at age 4.5years as being less socially competent and asshowing more externalizing problems, but theywere observed to engage in more positive behaviorduring the child care observations. Change in timespent in child care proved unrelated to kindergartenoutcomes, however.

Alternative Explanation: Other Features of Child Care

As part of a series of nested regression analyses, asecond model (i.e., Model 2) was estimated todetermine whether effects of quantity of care justdescribed would remain significantFor evenemerge for the first-time (i.e., suppression effects)Fwhen four other features of child care were added tothe base model: average quality of child care,proportion of time spent in center care, proportionof peer-group exposure, and instability of care. Tothe extent that quantity of care effects remainsignificant with these other variables included, thiswould indicate that quantity effects were notexclusively a function of other features of child carepurposefully excluded from the base model. Inspec-tion of the fifth column in Table 2, which reports thesignificance of the two quantity of care predictorscombined when additional child care factors areadded to the base model, indicates that in only oneof six cases did a previously significant pooled effectof quantity prove insignificant in Model 2 (i.e.,maternal report of externalizing problems in kinder-garten: from po.05 to p4.05) and that two pre-viously insignificant pooled quantity of care effectsproved significant once other features of child careexperience were taken into consideration (i.e.,mother-reported social competence at 54 monthsand in kindergarten: from p4.05 to po.01 and .05,respectively).

Inspection of the second column labeled ‘‘Model2’’ in Table 2 (i.e., eighth column of table) shows thatall significant effects of hours of care per week (i.e.,HLM intercept) detected in Model 1 remainedsignificant when the four additional child carevariables were added to the prediction model, andthat in the case of mother-rated social competence, apreviously insignificant effect of hours per week incare proved significant, indicating that more time incare predicted less mother-reported social compe-


Table 2

Effects of Quantity of Care (Mean Hours per Week, Linear Change Over Time) in Model 1 (Base: Covariates Only), Model 2 (Plus Additional Child Care Predictors), and Model 3 (Plus Maternal

Sensitivity)

Pooled testsa Hours of care per weekb Linear change over time in hours/week

54 monthsc N Statd Model 1:

base

Model 2:

adds child care

Model 3:

adds parenting

Model 1:

Base

Model 2:

adds child care

Model 3:

adds parenting

Model 1:

base

Model 2:

adds child care

Model 3:

adds parenting

M social competence 935 B n n � 0.05 � 0.09 nn � 0.07 n � 0.04 � 0.04 � 0.03

CG social competence 725 B nnn nnn nnn � 0.13 nnn � 0.15 nnn � 0.13 nn � 0.12 nn � 0.12 nn � 0.10nn

M.externalizing prob. 940 B 0.06 0.05 0.04 � 0.01 � 0.01 � 0.01

CG externalizing prob. 689 B nnn nnn nnn 0.25 nnn 0.23 nnn 0.22 nnn 0.08 n 0.07 n 0.07

CG conflict 691 B nnn nnn nnn 0.22 nnn 0.20 nnn 0.18 nnn 0.07 0.06 0.05

Dyad. play (positive) 698 B 0.01 � 0.04 � 0.03 � 0.02 � 0.03 � 0.02

Child care (positive) 838 B 0.02 � 0.01 0.01 0.07 n 0.06 0.07

Dyad. play (negative) 694 OR 1.01 n 1.01n 1.01 1.42 1.43 1.41

Child care (negative) 838 OR 1.01 1.01 1.01 0.92 0.86 0.85

Kindergartenc

M social competence 938 B n � 0.03 � 0.06 � 0.04 � 0.04 � 0.04 � 0.03

T social competence 893 B � 0.05 � 0.04 � 0.02 0.02 0.02 0.03

M externalizing prob. 940 B n 0.09 nn 0.08 n 0.08 n 0.01 0.00 � 0.00

T externalizing prob. 903 B nnn nn nn 0.14 nnn 0.11 nn 0.10 nn � 0.03 � 0.05 � 0.05

T conflict 905 B nnn n n 0.13 nnn 0.09 n 0.09 n � 0.02 � 0.03 � 0.04

aTwo degrees of freedom tests of no effect on intercept (mean) or slope (linear change).bBase model with factors for site, gender, ethnicity, mother’s education, maternal depression (intercept, slope) 6–54 months average income-to-needs ratio, 6-month temperament.Model 2 adds child care factors: average quality of care, proportion of center care, proportion of peer group exposure, instability of care. Model 3 adds parenting factors: mean (i.e.,HLM intercept) and linear change (i.e., HLM slope) from 6 to 54 months.cM5mother; CG5 caregiver; T5 teacher.dB5 beta; OR5 odds ratio.npo.05. nnpo.01. nnnpo.001.

990

NIC

HD

Early

Child

Care

Research

Netw

ork

tence at 54 months. Nevertheless, comparison ofparameter estimates from Model 1 and Model 2shows that effects of hours per week were modestlysmaller in the model including additional child carepredictors than in the base model for five of theoriginal seven significant effects (e.g., T externaliz-ing) and in one case larger (i.e., caregiver socialcompetence). The largest changes in regressioncoefficients for hours per week emerged for kinder-garten teacher’s report of externalizing problemsand conflict.

Inspection of the third column labeled ‘‘Model 2’’in Table 2 (i.e., 11th column of table) shows that inone of three cases a previously significant effect (inModel 1) of linear change over time in hours perweek of child care became insignificant whenadditional features of child care were taken intoconsideration (i.e., positive child care). In sum, eventhough inclusion of four additional features of childcare in the prediction model attenuated some of theoriginally detected effects of quantity of child care,in the main, effects of quantity remained significanteven with indicators of child care quality, type, andinstability taken into consideration.

Alternative Explanation: Maternal Sensitivity

As the third step in the series of nested regressionanalyses, Model 2 was modified by adding two

additional predictors reflecting average maternalsensitivity (i.e., HLM intercept) and linear changein maternal sensitivity (i.e., HLM slope) so thatquantity of care effects could be re-examined aftercontrolling for patterns of parenting that prior workon this sample showed were predicted by quantityof care (NICHD Early Child Care Network, 1999).Although small reductions of the hours coefficientswere observed for selected outcomes, all but one ofthe pooled effects of the two quantity of carepredictors in Model 3 (i.e., six of seven) remainedsignificant when the two maternal sensitivity pre-dictors were added to Model 2. Consideration ofspecific parameter estimates from Models 2 and 3indicates that seven of eight significant effects ofhours of care per week (i.e., HLM intercept) and oneof two significant effects of linear change over timein hours per week (i.e., HLM slope) remainedsignificant when maternal sensitivity predictorswere added to the prediction model. By and large,then, the previously detected effects for quantity ofcare remained significant when the alternativepredictor of maternal sensitivity was added to themodel.

To illuminate further the effects of average hoursper week of care across the first 54 months detectedin Model 3, Table 3 presents the adjusted meanscores for each outcome as a function of quantity ofcare grouped in terms of 0–9, 10–29, 30–45, and445hr

Table 3

Adjusted Means (SD) for Developmental Outcomes by Quantity of Care Groups (Average Hours per Week)

0–9 10–29 30–45 445

54 months

M social competence 101.1 (1.09) 98.9 (0.66) 97.5 (0.70) 98.7 (1.28)

CG social competence 108.2 (1.30) 104.7 (0.80) 103.8 (0.84) 103.2 (1.47)

M externalizing 50.4 (0.79) 51.8 (0.47) 51.6 (0.50) 52.9 (0.91)

CG externalizing 47.8 (0.94) 49.0 (0.60) 51.3 (0.60) 53.1 (1.02)

CG conflict 16.8 (0.67) 18.4 (0.43) 19.7 (0.43) 19.9 (0.73)

Dyadic play positive 0.4 (0.25) � 0.1 (0.15) 0.3 (0.16) � 0.4 (0.29)

Child care positive 2.35 (0.09) 2.3 (0.06) 2.4 (0.06) 2.3 (0.10)

Kindergarten

M social competence 104.6 (1.20) 102.2 (0.72) 102.6 (0.76) 103.2 (1.44)

CG social competence 104.6 (1.23) 103.4 (0.74) 103.4 (0.77) 102.3 (1.45)

M externalizing 48.2 (0.80) 50.2 (0.48) 50.3 (0.51) 51.1 (0.96)

T externalizing 48.4 (0.77) 49.3 (0.47) 50.6 (0.49) 52.1 (0.91)

T conflict 10.2 (0.47) 10.3 (0.29) 11.0 (0.30) 11.7 (0.56)

Note: Quantity effects controlling for sex, gender, ethnicity, mother’s education, maternal depression (intercept and slope), 6–54 monthsaverage income-to-needs ratio, 6-month temperament, average quality of care, proportion of peer group exposure, instability of care,parenting (intercept and slope). M5mother; CG5 caregiver; T5 teacher.


of nonmaternal care per week. Given the variousgroupings of hours per week of care found in theliterature, these groupings were selected to reflectlimited, moderate, high, and very high quantities ofnonmaternal care. Inspection of the relevant ad-justed means shows that where significant effects ofaverage quantity of care emerged in Model 3, smallbut steady increases in negative outcomes (e.g.,externalizing problems) and small but steady de-creases in positive outcomes (e.g., social compe-tence) are evident as one progresses from limited tomoderate to high to very high quantities ofnonmaternal care. At the same time, it should benoted that even children experiencing the greatestamount of nonmaternal care do not score, onaverage, in (or even near) the at-risk range onexternalizing problems (i.e.,X60). Furthermore, theabsolute differences between the adjusted means forthe groups with the lowest and highest amounts ofcare are in all cases less than one half of the samplestandard deviation (see Table 1).

Effect sizes. Through this point the focus of resultshas been on the nature and significance of quantityof care effects. To evaluate the explanatory power ofeach predictor included in Model 3 and to illuminatethe absolute and relative power of quantity of care topredict socioemotional adjustment, effect size esti-mates were calculated for all predictors in Model 3for the 7 (of 14) developmental outcomes found to berelated to quantity of care. Two effect size estimateswere calculated for each predictor for each outcome,one conservative and one liberal. The former tookthe form of a partial correlation representing therelation between each predictor in Model 3 and eachoutcome after controlling for all other predictors inthe model. The latter took the form of a structuralcoefficient, reflecting the ratio of the correlationbetween the intercept from hours of care growthcurve and the outcome divided by the multiplecorrelation from Model 3 (Courville & Thompson,2001); the multiple correlation is the square root ofthe R2 for Model 3 as shown in Table 3. Thus, thestructural coefficient reflects the relative predictivepower of each predictor included in the analysismodel without adjusting for shared variance amongthe predictors. Structural coefficients reflect theattenuated zero-order correlations under the as-sumption that all unmeasured variance would showthe same linear association as the measured var-iance. These coefficients are interpreted within thecontext of a given model (i.e., within each column inTable 3) by identifying the coefficients that arelargest as the best unconditional predictors if theoverall model provides significant prediction of the

outcome. Examination of both the structural coeffi-cients and partial correlations provides informationabout both the degree to which the predictor isassociated with the outcome and the degress towhich it provides unique prediction. Table 4presents these effect size statistics for the fourrelevant 54-month outcomes, and Table 5 presentsthe corresponding statistics for the three relevantkindergarten outcomes.

Considering first the 54-month data, it can be seenin Table 4 that significant quantity of care effectsrange from small, as evident in the partial correla-tion of .08 (po.05) linking average hours of care perweek with mother-reported social competence, tosubstantial, as evident in the structural coefficient of.64 (po.001) linking the same quantity variable withcaregiver-reported conflict. Noteworthy, though, isthat quantity of care, especially average hours perweek, is typically a stronger predictor of the childoutcomes displayed in Table 4 than are the otherfour features of child care included in Model 3. Ofimportance, nevertheless, is that higher quality ofcare significantly predicts greater mother- andcaregiver-rated social competence and lower levelsof caregiver-rated behavior problems and conflict,though only when the more liberal effect sizeestimates are considered (i.e., no significant partialcorrelations); that a greater proportion of time spentin center-based care predicts more caregiver-re-ported behavior problems and conflict; that greaterpeer-group exposure predicts less social competenceand greater conflict reported by caregivers; and thatinstability of care is generally unrelated to theoutcomes under consideration.

Not to be missed in this description of child careeffect sizes is the fact that several background factorstreated as covariates in the nested regressionanalyses proved to be stronger predictors of someor several outcomes than any child care factor. Thisis especially true of the social class indicators ofmaternal education and family income-to-needsratio (averaged across 6–54 months) and to somelimited extent of maternal depression (averagedacross 6–54 months; see mother-reported socialcompetence) and child gender. With the exceptionof maternal report of social competence, the mater-nal report measure of difficult temperament provedto be an insignificant predictor. It is maternalsensitivity, especially average sensitivity over time,that proves to be the most consistent and generallystrongest predictor of all outcomes displayed, withgreater sensitivity predicting greater caregiver- andmother-reported social competence and less care-giver-reported behavior problems and conflict.


Table 4

Summary of Predictors of Select Month 54 Outcomes: Conservative (rp ) and Liberal (r/R) Effect Sizes

Mother-reported

social competence

Caregiver-reported

social competence

Caregiver-reported

problems

Caregiver-reported

conflict

Covariate predictors rp r/R rp r/R rp r/R rp r/R

Child gender5male .16nnn .36nnn � .15nnn � .36nnn � .03 � .05 .01 .06

Child ethnicity5Black .02 � .27nnn .04 � .26nnn � .02 .22n .02 .22

Child ethnicity5Hispanic � .02 � .14 .01 � .11 � .03 .04 .02 .17

Child ethnicity5Other � .01 � .09 � .01 � .01 .02 .03 � .00 � .05

Mother’s education � .02 .39nnn .06 .49nnn � .06 � .42nnn � .00 � .18

Average income-to-needs ratio (6–54 months) .00 .33nnn � .02 .36nnn .02 � .24nn .03 � .09

Difficult temperament (6 months) � .17nnn � .52nnn � .04 � .15 .04 .13 .05 .09

Mother’s depression (average) � .14nnn � .63nnn � .03 � .33nnn .03 .32 .03 .28n

Mother’s depression (slope) .01 � .00 � .04 � .08 .03 .06 .07 .20

Child care predictors

Quantity: mean hours/week � .08n � .05 � .12nn � .24nnn .20nnn .56nnn .16nnn .64nnn

Quantity: linear change/slope � .04 � .14 � .11nn -.37nnn .08n .25nn .05 .21

Quality mean � .02 .16n .00 .30nnn � .01 � .32nnn -.02 � .37nn

Center proportion .04 .04 .03 � .08 .06 .40nnn .07 .49nnn

Instability (3–34 months) .08n .15 .00 .01 � .05 .01 .01 .10

Peer group exposure: proportion � .06 � .15 � .08n � .21nnn .01 .14 .07 .28n

Parenting

Maternal sensitivity: mean .16nnn .67nnn .16nnn .72nnn � .13nnn � .61nnn � .09n � .46nnn

Maternal sensitivity: linear change over time .04 .21nn .05 .21n � .13nnn � .32nnn � .07 � .26n

R2 .18nnn .16nnn .17nnn .11nnn

npo.05. nnpo.01. nnnpo.001.

Table 5

Summary of Predictors of Select Kindergarten Outcomes: Conservative (rp ) and Liberal (r/R) Effect Sizes

Mother-reported problems Teacher-reported problems Teacher-reported conflict

Covariate predictor rp r/R rp r/R rp r/R

Child gender5male � .05 � .11 .00 .05 .12nnn .37nnn

Child ethnicity5Black � .06 .15 .06 .39nnn .09n .43nnn

Child ethnicity5Hispanic � .04 .02 � .00 .03 � .02 � .06

Child ethnicity5Other � .05 � .08 � .02 � .04 � .01 � .00

Mother’s education � .00 � .41nnn .02 � .29nnn .01 � .24n

Average income-to-needs ratio (6-54 months) � .01 � .36nnn .00 � .22n .00 � .17

Difficult temperament (6 months) .03 .24nn � .04 .06 � .05 � .02

Mother’s depression (mean) .24nnn .80nnn .04 .35nnn .00 .22n

Mother’s depression (linear change) .08n .23n � .06 � .10 � .04 � .03

Child care predictors

Quantity: mean hours/week .07n .16n .08n .36nnn .08n .37nnn

Quantity: linear change over time .00 .09 � .06 � .00 � .03 .03

Quality .01 � .16 � .04 � .39nnn .01 � .26nn

Center proportion .02 .11 .08n .40nnn .10nn .45nnn

Instability (3-34 months) � .01 .01 .02 .08 � .03 � .00

Peer group exposure: proportion � .00 .06 .02 .21 � .01 .12

Parenting

Maternal sensitivity: mean � .09nn � .56nnn � .12nnn � .63nnn � .09nn � .55nnn

Maternal sensitivity: linear change over time � .13nnn � .43nnn � .16nnn � .39nnn � .14nnn � .39nnn

R2 .17nnn .14nnn .12nnn

npo.05. nnpo.01. nnnpo.001.


Similar trends emerge when kindergarten out-comes are examined in Table 5. Effect sizes ofquantity of care tend to be smaller relative to thoseconsidered in Table 4, which is generally true of theexplanatory power of other child care and non–childcare predictors as well. In the case of averagequantity of care across the first 54 months of life,significant associations range from a small partialcorrelation of .07 (po.05), linking this quantityindicator with mother-reported problems, to amoderate structure coefficient of .37 (po.001), link-ing the same predictor with teacher-reportedconflict. Even though this average hours of careper week is more consistently related to thekindergarten outcomes in question than any otherchild care predictor, in several cases other child carevariables are stronger predictors of particular out-comes. Specifically, more time spent in centerspredicts more teacher-reported problems and con-flict, and lower quality of child care predicts moreteacher-reported problems and conflict (but onlystronger than quantity in the former case). Neitherinstability of care nor peer-group exposure signifi-cantly predicts the three kindergarten outcomesunder consideration.

Once again, though, maternal education, familyincome-to-needs ratio, and maternal depressionconsistently predict the kindergarten outcomes,though only sometimes more strongly than childcare variables. Finally, maternal sensitivity oncemore most consistently and, often, most stronglypredicts the kindergarten outcomes, with change inmaternal sensitivity over time (i.e., slope) becominga more consistent predictor than was evident inTable 4. Not only is greater sensitivity predictive ofless negative socioemotional adjustment in kinder-garten, but increases in sensitivity over time alsopredict lower levels of mother- and teacher-reportedbehavior problems and less teacher–child conflict.

Identifying Threshold Effects

There have been repeated suggestions in theliterature that potentially negative effects of timespent in child care emerge after some threshold ofhours is crossed (e.g., 20 hr per week: Belsky &Rovine, 1988; 30 hr per week: Vandell & Corasaniti,1990). Results presented through this point showingthat more time in care predicts less socioemotionaladjustment (i.e., lower social competence, greaterproblem behavior, more conflict) when quantity ofcare is treated as a continuous variable thusstimulated interest in identifying thresholds forthese effects. But because there is no consensus in

the literature regarding specific a priori cutoffs forhours per week in care, we conducted piecewiseregression in hopes of identifying thresholds. Thisanalytic technique is designed to identify points atwhich the relation between independent variable(i.e., quantity of care) and dependent variable (i.e.,child outcomes) changes over the scale of theindependent variable (Willett, Singer, & Martin,1997). As it turned out, no threshold effects proveddetectable, indicating that the linear relation be-tween more time in care and poorer socioemotionaladjustment was just thatFunchanging in the sensethat as quantity of care increased, so did problembehavior scores in a constant dose–response fashion.

Timing of Child Care

Some studies in the literature have focused onamount of child care or maternal employment in thefirst year (e.g., Bates et al., 1994; Belsky & Rovine,1988), other studies have focused on the first 2 and 3years (e.g., Belsky & Eggebeen, 1991; Harvey, 1999),and still other studies have focused the first 4 or 5years (e.g., Borge & Melhuish, 1995; Vandell &Corasaniti, 1990). In addressing the issue of quantityof care as a function of timing or child age, thisnaturalistic study confronted an ecological chal-lenge, in that hours per week in care was morerather than less stable across the first 4.5 years of life.Indeed, the cross-age correlation in average hoursper week in care ranged from a high of .77 (po.001)across Years 1 and 2 to a low of .50 (po.001) for Years1 through 4. In fact, the part–whole correlationbetween average hours of care in any one year andaverage hours of care across the first 54 monthsranged from .95 (po.001) for Year 2 (as well as Year3) to .66 (po.001) for Year 5. Such realities maketeasing apart timing effects difficult, as the varianceshared by predictors reflecting amount of care in anyyear is so great that it is likely that the varianceshared between quantity of care in any year and anyoutcome will likely be shared across years (i.e.,limited unique effects).

In view of this circumstance, two strategies wereadopted in attempt to illuminate timing effects inpredicting each of the outcomes found to be relatedto average hours per week of nonmaternal careacross the first 54 months of life (i.e., Model 3). First,Model 3, which included all covariates, multiplefeatures of child care, and maternal sensitivity, wasestimated several times to predict each dependentvariable, replacing the variable mean hours of careper week across the first 54 months with a quantityof care variable reflecting different periods (i.e., an


alternative prediction approach). Thus, in onemodel, average hours of care across 3 to 54 monthswas replaced with a variable reflecting averagehours of care across 3 to 6 months. In a secondFandseparateFmodel, the quantity of care predictor wasaverage hours of care across 3 to 12 months; in athird model, average hours of care across 3 to 24months; in a fourth model, average hours of careacross 3 to 36 months; and in a fifth model, averagehours of care across 3 to 54 months. This strategywas adopted because time in care accumulates frombirth through age 4.5 years; therefore, it seemedappropriate to examine the effect of accumulatingcare history.

The second strategy adopted in hopes of illumi-nating timing effects involved a separate estimationof Model 3, again predicting each of the outcomesthat were related to average hours of care in the first54 months. In this second set of regression equations,however, four quantity of care variables wereincluded as predictors in the same regressionequation (along with covariates, other child carevariables, and maternal sensitivity), each reflecting adistinct and nonoverlapping developmental period:3 to 6 months, 7 to 12 months, 13 to 24 months, 25to 36 months, and 37 to 54 months (i.e., a uniqueprediction approach). In this model, the uniquepredictive power of amount of care experience ineach developmental period (i.e., not shared with anyother developmental period) is tested. It is importantto note that if one (or more) periods in such ananalysis appears to be a stronger predictor of theoutcome in question than another period, this does

not necessarily mean that a period of greatersusceptiblity to quantity of care effects (i.e., asensitive period) has been detected. All it means isthat quantity of care during the period in questionshares more unique variance with the outcome.Other periods may actually be stronger predictors, ifconsidered alone, but because they share theirpredictive power with other periods, that sharedpredictive power is accorded to no particular periodin the analyses.

The results of the two sets of regression analysesare presented in Table 6 for the four relevant 54-month outcomes and in Table 7 for the three relevantkindergarten outcomes. Under each dependentvariable listed in each table are two sets ofstandardized regression weights. The weights inthe first set reflect the power of quantity of care,across different periods, to predict each outcome asdetermined in a series of separate regressionequations focused on amount of care in five periods(i.e., alternative predictors): 3 to 6 months, 3 to 12months, 3 to 24 months, 3 to 36 months, and 3 to 54months. Below this first set of regression weights arestandardized regression coefficients emerging froma single regression equation estimated for eachdependent variable, each weight reflecting theunique predictive power of quantity of care in fiveperiods (i.e., unique predictors): 3 to 6 months, 7 to12 months, 13 to 24 months, 25 to 36 months, and 37to 54 months. Several trends are evident in the data.

First, as quantity of care across longer periods isconsidered, the predictive power of average hoursper week of nonmaternal care tends to increase,

Table 6

Cumulative and Unique Effectsa of Quantity of Care (Average Hours per Week) on Select 54-Month Outcomes as a Function of Periodb

Alternative

predictors

Mother-reported

social competence (n5 933)

Caregiver-reported

social competence (n5 724)

Caregiver-reported

problems (n5 688)

Caregiver-reported

conflict (n5 690)

3–6 months � 0.01 � 0.02 0.11nn 0.09n

3–12 months � 0.03 � 0.04 0.11nn 0.10n

3–24 months � 0.05 � 0.06 0.15nnn 0.13nn

3–36 months � 0.06 � 0.09n 0.19nnn 0.16nnn

3–54 months � 0.06 � 0.12nn 0.20nnn 0.17nnn

Unique predictors

3–6 months 0.05 0.06 0.04 0.03

7–12 months � 0.00 0.01 � 0.08 � 0.07

13� 24 months � 0.06 0.03 � 0.02 0.02

25–36 months � 0.02 � 0.11 0.24nnn 0.19n

37–54 months � 0.04 � 0.11n 0.06 0.04

aStandardized regression coefficients.bQuantity effects controlling for sex, gender, ethnicity, mother’s education, maternal depression (intercept and slope), 6-54 monthsaverage income-to-needs ratio, 6-month temperament, average quality of care, proportion of peer group exposure, instability of care,parenting (intercept and slope).npo.05. nnpo.01. nnnpo.001.


usually slightly, except for teacher-reported externa-lizing problems and conflict, where predictionremains virtually unchanged from the period 3 to 6months to the period 3 to 54 months. Thus, averagehours per week in care across the first 4.5 yearsproves generally more predictive than average hoursper week in care across the first 3 years, whichproves generally more predictive than average hoursper week in care across the first 2 years, whichproves more predictive than average hours per weekin care across the first year only. At the same time,the data in the top halves of Tables 6 and 7 indicatethat the more time children spend in nonmaternalcare during the earliest period (i.e., 3 to 6 months),the more externalizing problems and conflict withcaregiver and teacher are evident at 54 months andin kindergarten.

Inspection of the data in the bottom halves ofTables 6 and 7 reveals, however, that with theexception of predicting teacher-reported externaliz-ing problems in kindergarten, neither amount ofnonmaternal care in the period 3 to 6 months or inthe first year makes a significant unique predictiononce time in care during other age periods is takeninto consideration. The same is true across alloutcomes for the second year of life and for mostoutcomes for the third year of life and for the period37 to 54 months. Note, however, that more time innonmaternal care during the third year of lifeuniquely predicts greater caregiver-reported exter-

nalizing problems and conflict at 54 months (but notin kindergarten), and that more time in care for theperiod 37 to 54 months uniquely predicts lesscaregiver-reported social competence at 54 months.These results may reflect the results found for theeffect of hours slope reported in Table 2.

Predicting Higher Levels of Problem Behavior

Having found in the first set of analyses presentedin Table 2 that more time in nonmaternal child carethrough 54 months of age predicted more problembehavior, as reported consistently by teachers at both54 months and kindergarten, but by mothers only atkindergarten age, we sought to determine whetherchildren who experienced more hours of child carewere more likely to evince levels of problemsconsidered high by certain a priori standards. It isimportant to note that we were not able to examinethe criteria for identifying clinical levels of problems(i.e., scores above 2 SD the mean at any time orabove 1.5 SD according to two informants) becausetoo few children in our sample met those criteria(no50 in the largest group and no20 in othergroups). Instead, we used a recommended ‘‘risk’’cutoff score of 1 SD above the mean (e.g., tX60). Thiscorresponds to scoring at or above the 84thpercentile. Achenbach (1991) recommended thatchildren in this range could be, but not necessarilyshould be, referred for further evaluation for clinical-level problems.

Logistic regression analyses examined whetherchildren classified in the four quantity-of-caregroups reflecting limited, moderate, high, and veryhigh levels of child across their first 4.5 years of lifecare differed in the proportions of children scoringin the at-risk range of tX60. The full modelcovariates were included in these analyses, alongwith the four child care hours groups. Descriptiveresults shown in Table 8 indicate, consistent with theregression analyses reported using the continuousscaling of child care hours and externalizingbehavior, that the proportion of children scoring inthe at-risk range tended to increase as amount oftime in care increased, relative to the proportion ofchildren in the at-risk range who experienced loweramounts of nonmaternal care. At the same time, theoverwhelming majority of children did not score inthe at-risk range, even when considering thoseexperiencing the most child care.

Aggression and Disobedience or Assertiveness?

When data such as those emerging from thisinquiry have been reported in the past linking

Table 7

Cumulative and Unique Effectsa of Quantity of Care (Average Hours per

Week) on Select Kindergarten Outcomes as a Function of Periodb

Alternative

predictors

Mother-reported

problems

(n5 938)

Teacher-reported

problems

(n5 901)

Teacher-reported

conflict

(n5 903)

3–6 months 0.05 0.12nnn 0.10nn

3–12 months 0.06 0.12nnn 0.09nn

3–24 months 0.07n 0.12nnn 0.10nn

3–36 months 0.07n 0.11nn 0.10nn

3–54 months 0.08n 0.12nn 0.10nn

Unique predictors

3–6 months 0.01 0.11n 0.09

7–12 months 0.02 � 0.01 � 0.05

13–24 months 0.05 0.02 0.07

25–36 months � 0.05 � 0.01 � 0.01

37–54 months 0.07 0.03 0.01

aStandardized regression coefficients.bQuantity effects controlling for sex, gender, ethnicity, mother’seducation, maternal depression (intercept and slope), 6-54 monthsaverage income-to-needs ratio, 6-month temperament, averagequality of care, proportion of peer group exposure, instability ofcare, parenting (intercept and slope).npo.05. nnpo.01. nnnpo.001.


amount and timing of child care with problembehavior, the suggestion has been advanced thatassertiveness may be confused with aggressive anddisobedient behavior by raters (Clarke-Stewart,1989). To address this issue, items on the externaliz-ing problem behavior scale were sorted to createthree subscales: one reflecting aggression (e.g.,cruelty to others, destroys own things, gets in manyfights, threatens others, hits others), one reflectingdisobedience or noncompliance (e.g., defiant,uncooperative, fails to carry out assigned tasks,temper tantrums, disrupts class discipline), and athird reflecting assertiveness (e.g., bragging orboasting, talks too much, demands or wants atten-tion, argues a lot). Internal consistency reliability(i.e., coefficient alpha) for these three subscales formothers (54 months, kindergarten), caregivers (54months), and teachers (kindergarten) ranged from.67 (54 months, mother-rated assertiveness) to .85 (54months, kindergarten-teacher-rated aggression). Theresulting scale scores were then subject to a Poissonregression that included all the predictors fromModel 3 (i.e., covariates, multiple child care pre-dictors, maternal sensitivity). (The Poisson regres-sion is appropriate for analyzing data in which theoutcome is a count of relatively rare events. TheCBCL ratings on the selected items were all highlyskewed toward 0; therefore, use of mulitple regres-sion methods was inappropriate.)

The resultant effects of the two quantity of carepredictorsFmean hours per week of care across theperiod 3 to 54 months (i.e., HLM intercept) andlinear change in hours per week across the sameperiod (i.e., HLM slope)Fare displayed in Table 9.Inspection of the table reveals that the more timechildren spent in nonmaternal care across their first4.5 years, the more mothers, caregivers, and teacherscharacterized them as assertive at 54 months and inkindergarten, and the more caregivers at 54 monthsand teachers in kindergarten characterized them asaggressive and disobedient. It would seem, then,that more time in care across the first 4.5 years,though not change in hours of care per week overtime, is predictive of more aggression and disobe-dience, according to caregiver and teacher reports,and assertiveness, according to mother, caregiver,and teacher reports.

Discussion

The primary purpose of the current study was toexamine in more detail than has been possible inprior research issues related to the socioemotionaladjustment of children who vary in the amount ofroutine nonmaternal care they experience across

Table 8

Adjusted Proportion of Children Scoring 1 or More SDs Above or Equal

to Mean (60) on Externalizing Problems as Rated by Mothers (M) and

Caregivers (CG) at 54 Months and Mothers (M) and Teachers (T) in

Kindergartena

54 months CBCL externalizing scores

Hours/week N Caregiver N Mother

0-9 117 2% 153 10%

10-29 241 12% 358 17%

30-45 249 18% 331 14%

445 82 24% 98 26%

Kindergarten CBCL externalizing scores

Hours/week N Teacher N Mother

0-9 147 9% 152 8%

10-29 341 12% 359 12%

30-45 324 15% 336 12%

445 91 19% 93 21%

aProportions are adjusted for site, child gender, child ethnicity,maternal education, average income-to-needs ratio (6–54 months),6-month temperament, maternal depression (intercept and slope),parenting (intercept and slope), child care quality (intercept),proportion of center care, proportion of peer group exposure,instability of care. CBCL stands for Child Behavior Checklist(Achenbach, 1991).

Table 9

Effects of Quantity of Care (Mean and Slope) in Model 3 on Aggression,

Disobedience, and Assertiveness: Using Poisson Regressiona

54-month subscales N Pooledb Hours of

care per

week

Linear

change: hours

per week

Mother-rated aggression 931 ns ns

Mother-rated disobedience 934 ns ns

Mother-rated assertiveness 936 (1) n ns

Teacher-rated aggression 593 nnn (1) nnn ns

Teacher-rated disobedience 676 nnn (1) nnn ns

Teacher-rated assertiveness 698 nnn (1) nnn ns

Kindergarten subscales

Mother-rated aggression 924 ns ns

Mother-rated disobedience 925 ns ns

Mother-rated assertiveness 926 n (1) nn ns

Teacher-rated aggression 891 n (1) nn ns

Teacher-rated disobedience 880 nn (1) nnn ns

Teacher-rated assertiveness 894 nnn (1) nnn ns

aPositive coefficients (1) indicate an increase in mean outcome;negative coefficients (-) indicate a decrease in mean outcome.bTwo degrees of freedom tests of no effect of hours intercept andslope controlling for sex, gender, ethnicity, mother’s education,maternal depression (intercept and slope), 6-54 months averageincome-to-needs ratio, 6-month temperament, average quality ofcare, proportion of peer group exposure, instability of care,parenting (intercept and slope).npo.05. nnpo.01. nnnpo001.


their first 4.5 years of life. Perhaps most significant interms of extending past work was our ability toexamine effects of time in child care after taking intoconsideration a variety of other factors, includingfamily background factors and other attributes ofchild care, especially its quality. Much past worksummarized in the introduction suggesting thatrisks are associated with lots of time spent innonmaternal care beginning in the opening year(s)of life has been challenged because it failed to takeinto account factors such as these that could beresponsible for previously detected associationsbetween timing of and time spent in nonmaternalcare and indexes of socioemotional adjustment(e.g., McGurk, Caplan, Hennessy, & Moss, 1993;Phillips et al., 1987; Richters & Zahn-Waxler, 1990;Thompson, 1988).

Effects of Quantity and Other Features of Child Care

Consistent with findings from many, but not all,studies cited in the introduction, results from thecurrent research show, even after controlling formultiple family background factors, multiple fea-tures of child care assessed longitudinally, andrepeated assessments of maternal sensitivity, thatmore time in nonmaternal care across the first 4.5years of life (i.e., quantity intercept) predictedseveral interrelated indicators of social functioning:(a) less social competence reported by mothers andcaregivers at 54 months (but not at kindergartenage), (b) more externalizing problems reported bycaregivers at 54 months and by mothers and teachersat kindergarten age, and (c) more adult–childconflict reported by caregivers at 54 months and bykindergarten teachers. Thus, quantity of nonmater-nal care was significantly, even if modestly, asso-ciated with less positive adjustment as reported bythree sets of respondents, each of whom hadextensive experience with children in one of threesettings (child care, home, kindergarten).

The fact that quantity of care proved unrelated toobservational measures of child functioning once thefull set of control variables were included in theprediction model (i.e., Model 3) could be a result ofthe limited amount of time that behavior wasobserved and the relatively low frequencies of thebehaviors measured. The fact that change over timein quantity of care (i.e., quantity slope) significantlypredicted only a single 54-month outcome and nokindergarten outcomes in the full prediction modelindicated that it was likely the cumulative amount oftime that children spent in nonmaternal care acrossthe infancy, toddler, and preschool years that was

related to socioemotional adjustment (in the primaryanalyses: Models 1–3) rather than increases ordecreases in nonmaternal care experience. Further-more, the fact that significant effects of averagehours of care per year detected at 54 months oncaregiver reports of externalizing problems andadult–child conflict were replicated at kindergartenage using different raters (i.e., kindergarten tea-chers)Fraters with roughly equivalent experiencewith and knowledge of the children they evalua-tedFsuggests that the results reported are unlikelyto be an artifact of raters of children in full-time caresimply having more opportunity to witness negativebehavior than raters of children with more limitedcare experience.

It must be noted, of course, that effect sizeestimates of time spent in nonmaternal care werenever large and certainly modest, though whenconsidered in terms of the more liberal, structure-coefficient effect size estimates, several qualified asmoderate in magnitude, especially at 54 months.Recall as well that effect sizes of all predictors,including maternal sensitivity and indexes of socio-economic status, were generally smaller when itcame to predicting kindergarten outcomes. Suchresults that show that as time between themeasurement of predictor and outcome increasespredictive power tends to decrease, are con-sistent with the notion that development is anopen-ended process and that early developmentalexperiences should not be presumed to haveenduring effects.

Even though detected effects of quantity of childcare were by no means strong, it must be regarded asnoteworthy that they emerged after controlling for avariety of family background factors; remainedsignificant for the most part even when multipleaspects of child care, as well as maternal sensitivity,were taken into consideration; and were not re-stricted to adjustment outcomes based on a singlereporter. This is not to say, however, that they wereunaffected by inclusion in prediction models ofother features of child care (i.e., Model 2) andmaternal sensitivity (i.e., Model 3). Recall thatregressions coefficients representing quantity effectsoften (but not always) became smaller when theseother predictors were taken into consideration, evenas they remained significant. Clearly, child carefactors other than quantity of care and maternalsensitivity accounted for some, though by no meansall or even most, of the initially detected (in Model 1)quantity-of-care effects.

In fact, findings pertaining to the effects of otherfeatures of child care should not be overlooked in


this investigation, the primary focus of which hasbeen on quantity of care. Recall that effect sizeestimates showed, at least when more liberalstructure coefficients were considered, that higherquality of care predicted more mother- and care-giver-reported social competence, lower levels ofexternalizing problems, and less caregiver-reportedconflict at 54 months and fewer externalizingproblems and less conflict according to kindergartenteachers. Recall as well that more experience in childcare centers predicted more externalizing problemsand adult–child conflict at both 54 months and atkindergarten age, with kindergarten effects beingindependent of the parallel effects of more timespent in nonmaternal care of any kind. What thismeans is that when children spent more time innonmaternal care of any kind, and more of that timein center-based care in particular, they were espe-cially likely to be rated highly on externalizingproblems and teacher–child conflict in kindergarten.Experience with two or more agemates in a childcare arrangement and instability of care provedrelated to some 54-month outcomes, but none ofthese were replicated at kindergarten age (and so arenot discussed further). In sum, even though averagenumber of hours of child care proved to be the mostconsistent and usually the strongest predictor ofsocioemotional adjustment among the multiplefeatures of child care considered in this inquiry, itwas by no means the only aspect of child care thatwas significantly related to children’s socioemo-tional adjustment.

Thresholds of Quantity

In addition to examining effects of time spent innonmaternal care, net of a variety of other factors,we sought to extend research on child care in severalways. In some cases we proved successful, whereasin others we did not. Evidence did not emerge, forexample, that it was experiencing more than 20 or 30hr per week of care on average that was especiallyimportant, as regression analyses designed to detectpoints at which the relation between quantity of careand development outcome changed failed to identi-fy any such thresholds. Thus, it appears that a lineardose–response relation most accurately characterizesthe association between amount of child careexperience and socioemotional adjustment in theNICHD data.

Timing and Quantity

When it came to illuminating issues of timing, thefact that individual differences in amount of care

were stable over time, even though change wasdiscerned, severely constrained our capacity to drawstrong conclusions. Consider in this regard the factthat with respect to 54-month outcomes, there was aconsistent tendency for predictive power to increaseFever so modestlyFas the period examined withrespect to quantity of care became progressivelylonger, from 3 to 6 months, to 3 to 12 months, to 3 to54 months. Such data seem consistent with Belsky’s(1999, 2001) conclusion that it is early, extensive, andcontinuous careFand thus cumulative quantity ofcareFthat is most likely to be predictive of lowerlevels of socioemotional adjustment. Yet the samepattern of increasing predictive power of quantity ofcare as the period considered increased was notevident when teacher-reported kindergarten out-comes were the focus of analysis. In fact, whenconsidering externalizing problems and adult–childconflict, the predictive power of average quantity ofcare across 3 to 54 months was exactly the same asthat across 3 to 6 months. Such findings raise theprospect that it is the initiation of lots of time in carein the first 6 months of life that is most developmen-tally influential.

Yet when issues of timing were examined in adifferent way, not by considering increasinglylengthy periods but by testing unique effects ofparticular periods, the story that emerged wasdifferent. This is because controlling for amount ofcare at ages older than 3 to 6 months resulted in onlya single instance of quantity of care in this periodevincing any unique predictive power (i.e., teacher-reported externalizing problems). Results such asthese suggest that the first 6 months of life is not asensitive period per se, but rather that the effects ofearlier experiences (i.e., the first 6 months) arecontingent on later experiences. Thus, it may be thatreductions in amount of time spent in child care atolder ages would, for the most part, attenuate themodest effect of lots of time in care in the first 6months of life. But this conclusion must be embracedcautiouslyFand contingentlyFbecause the data onunique effects also revealed some unique predictivepower, though only for 54-month outcomes, ofquantity of care experienced in the third year (i.e.,caregiver-reported problems and conflict) and in theperiod 37 to 54 months (i.e., caregiver-reportedsocial competence).

Perhaps the best summary of timing findingswould be that (a) some evidence indicates that it isthe cumulative quantity of nonmaternal care, typi-cally initiated in infancy and experienced across theinfancy, toddler, and preschool years, that is mostpredictive of socioemotional adjustment rather than


the amount of time spent in nonmaternal care duringany particular period; (b) some evidence indicatesthat later social adjustment varies as a function ofexposure to more hours in care during infancy itself;and (c) some evidence indicates that amount of careexperienced in the third year (i.e., caregiver-reportedproblems and conflict at 54 months) and fourth year(i.e., caregiver-reported social competence at 54months) affects socioemotional adjustment indepen-dent of hours in care experienced in other develop-mental periods. However, qualifying theseobservations must be appreciation of the fact thatthe nonexperimental nature of the NICHD studygreatly limits our capacity to tease apart timingeffects, given that children who spent lots (or little)time in nonmaternal care in their first 6 months orfirst year or two of life tended to do so at later agesas well.

High Levels of Problem Behavior and Aggression,Disobedience, and Assertiveness

Given the fact that the children participating inthe NICHD Study of Early Child Care are notprincipally from high-risk families, it would besurprising to find a substantial portion of the sampleevincing clinical levels of problem behavior. And, infact, they do not, so much so, in fact, that even astudy like this of hundreds of children provides aninsufficient number scoring above the clinical cutoff(T470) to permit confident, multivariate evaluationof relations between very high levels of externalizingproblems and quantity of child care. Nevertheless,examination of relations between at-risk levels ofexternalizing problems, defined as scoring at orabove 1 SD above the mean of the standarizationsample (i.e., T460), did prove possible. It showed,consistent with results considered through thispoint, that as average amount of child care per weekacross the first 54 months of age increased, so did thelikelihood that childrenFthough only a modestproportionFwould score in the at-risk range onexternalizing problems.

It is important to point out that although, bydefinition, approximately 17% of children scored ator above 1 SD above the mean in the normingsample of the CBCL, this is a questionable bench-mark to use when considering the proportions ofchildren presented in Table 8 scoring high onproblem behavior as a function of time in nonma-ternal care, for two reasons: (a) the NICHD studysample cannot be assumed to be identical to theCBCL norming sample and (b) the data presentedare based on adjusted problem behavior scores. In

fact, the proportion of the analysis sample scoring ator above 1 SD above the mean on externalizingproblems (after adjusting for multiple covariates)was always less than 17%: 13.9% and 15.7%, respec-tively, in the case of caregiver and mother ratings at 54months, and 13.3% and 12.2%, respectively, in the caseof teacher and mother ratings in kindergarten. These,perhaps, are more appropriate benchmarks.

Ever since findings linking aggression and pro-blem behavior with lots of time in nonmaternal careor its initiation very early in life emerged in theresearch literature, questions have arisen about theinterpretation of such data. Clarke-Stewart (1988,1989) and others (McGurk et al., 1993) havesuggested that assertiveness may be confused withaggression or disobedience and thus discussion of‘‘risks’’ associated with early child care could bemisguided. To illuminate this issue and advanceresearch in this area, items from the problembehavior inventory administered to mothers, care-givers, and teachers were combed for items thatcould reasonably be categorized as reflecting asser-tiveness, disobedience, and aggression. The final setof analyses reported in this article showed that theeffects detected in the case of caregiver- and teacher-rated externalizing problems were not simply afunction of more time in child care predicting greaterassertiveness (though this was true with respect tomaternal reports); rather, more time in child carepredicted more aggressive, disobedient, and asser-tive behavior according to caregiver reports at 54months and teachers in kindergarten.

Importance of the Family

One of the most significant conclusions of thisinquiry is that even though quantity of child care, aswell as other features of child care, significantlypredicted multiple indexes of socioemotional adjust-ment, it was other facets of the ecology of childdevelopment that proved most predictive of socialcompetence, problem behavior, and adult–childconflict. Recall that the most consistent and strongestpredictor of all developmental outcomes to whichquantity of child care proved to be related wasmaternal sensitivity: When mothers provided moresensitive care (i.e., sensitivity intercept), their chil-dren evinced greater social competence, fewerproblem behaviors, and less conflict with adults atboth 54 months and in kindergarten; this was alsotrue of child functioning at kindergarten age whenmaternal sensitivity increased across the infant,toddler, and preschool years (i.e., sensitivity slope).Important as well is that more positive and less


negative adjustment was related to growing up in ahousehold in which mothers were more highlyeducated and less depressed, and in which therewere greater economic resources. It is critical to keepin mind such apparent influences of the familyrearing environment when considering relationsdetected in this inquiry between any feature of childcare and child development.

Conclusion

The past 15 to 25 years have witnessed a greatdeal of debate among developmentalists, parents,and policy makers about how child care may affectchild development. The NICHD study was plannedto address this issue (and many more), withinacknowledged limits of a nonexperimental researchdesign. In planning this work, the investigators werecognizant of multiple problems that plagued pastresearch and the confusion it has caused in inter-preting research findings in this area of inquiry. TheNICHD study has not overcome all these limits byany means, though it has been able to overcomemany of them. In so doing, it has been able toaddress a variety of issues about which there hasbeen open and healthy intellectual disagreement.

The results of the present study advance under-standing of potential effects of amount of time spentin early child care in showing, most significantly, (a)that the more time children spend in any of a varietyof nonmaternal care arrangements across the first 4.5years of life, the more externalizing problems andconflict with adults they manifest at 54 months ofage and in kindergarten, as reported by mothers,caregivers, and teachers; (b) that these effectsremain, for the most part, even when quality, type,and instability of child care are controlled, as well asmaternal sensitivity and other family backgroundfactors; (c) that the magnitude of quantity of careeffects are limited, though typically greater thanthose of other features of child care, though not ofmaternal sensitivity and family socioeconomic sta-tus; (d) that there is no apparent threshold forquantity effects; (e) that in most cases effects ofquantity of care are of a cumulative nature or effectsof time spent in nonmaternal care in the earliestmonths and years of life are contingent on amount ofcare experienced subsequently; and (f) that moretime in care not only predicts problem behaviormeasured on a continuous scale but at-risk (thoughnot clinical) levels of problem behavior, as well asassertiveness, disobedience, and aggression. Itshould also be noted that these correlational fndingsalso imply that lower levels of problems wereassociated with less time in child care.

Even though the effects of time spent in non-maternal care were statistically significant, it must beacknowledged that, in the main, effect sizes asso-ciated with them were limited (Cohen, 1977). Whenit comes to interpreting effect sizes, it is important toremember that Cohen offered conventions to guidepower analysis, not as a metric with which todismiss statistically significant findings. Evaluationsof the practical importance of research findings thatare modest in magnitude are not straightforward,because effect size estimates are affected by mea-surement, design, and method (McCartney & Ro-senthal, 2000). In the health domain, small effects aretaken seriously. Consider the fact that the effect ofaspirin on reducing heart attack is statistically verysmall (r25 .001, with corresponding r5 .034; Ro-senthal, 1994), yet the findings have influencedmedical practice.

In advancing the conclusions presented pre-viously, we acknowledge that despite the inclusionof controls for selection effects, it remains possiblethat the detected relations between time in care andproblem behavior could reflect effects of children’sbehavior on use of nonmaternal care. Conceivably,children who are more aggressive and disobedientthan others could be placed in child care at youngerages and for longer periods, and children who areshy and nonaggressive may be less likely to beplaced in child care, particularly with large groupsof children. Seemingly inconsistent with this analy-sis, however, is the fact that effects of time in careremained even when mother-reported difficult tem-perament at age 6 months was taken into considera-tion and were evident when just time in care in thefirst 6 months of life was used to predict socio-emotional adjustment. Nevertheless, the correla-tional nature of our longitudinal data does notpermit an unambiguous determination of causaldirection.

This observation highlights the need for futurework focused on mechanisms or processes ofinfluence, especially as most quantity effects re-mained significant, even if attenuated, when mater-nal sensitivity and other features of child care werecontrolled. In addition to future work focused onfamily interaction processes, including parenting,children’s physiological stress reactivity may also beworthy of investigation (Booth, Carver, & Granger,2000; Stansbury & Gunnar, 1994). This is becauserelations have been detected between separationfrom mother and children’s threshold for cortisolproduction as well as the size of the cortisol increasein response to a stressor (Gunnar, Mangelsdorf,Larsen, & Herstgaard, 1998). In fact, recent


research indicates that long days in child care areassociated with elevated levels of cortisol among 3-and 4-year-olds (Dettling, Gunnar, & Donzella, 1999;Tout, de Hann, Kipp Campbell, & Gunnar, 1998).Additional research on social processes taking placewithin child care environments also merits consid-eration.

When set in a larger context, the results summar-ized here regarding amount of time spent in childcare may have implications for school readiness andthe transition to school (Pianta & Cox, 1999).National surveys of kindergarten teachers revealthe emphasis they place on the importance of socialand emotional competenciesFfollowing directions,getting along with peers, cooperation with adults,and other markers of self-regulationFin determin-ing the degree to which a child is succeeding inmaking an adjustment to kindergarten (Rimm-Kauffman, Pianta, Cox, & Early, 2000). In light ofthese views, there is reason to wonder whetheramount of time in child care will prove related tochildren’s adjustment as they progress in elementaryschool (and beyond). The ongoing investigation ofthe NICHD sample will afford the opportunity toaddress this topic. Most important with regard to thefindings of the present study, we plan to examine theextent to which the significant increments inproblem behavior and adult–child conflict asso-ciated with amount of nonmaternal care are stableor interact with schooling experiences as childrenmove through their elementary school years. Just ashigh levels of maternal sensitivity and features ofchild care other than quantity attenuated some of theapparent consequences of quantity of care, high-quality classroom experiences could mitigate thenegative correlates of quantity of care that emergedin this study. On the other hand, it is also possiblethat classrooms characterized by poor managementand instructional practices could amplify theseassociations with more time in nonmaternal carethrough the infant, toddler, and preschool years.Finally, the detected relations of time in nonmaternalcare to more negative socioemotional adjustmentmay simply disappear as children get older, regard-less of their classroom experiences. Evidence con-sistent with the latter possibility comes from someearlier research (Egeland & Heister, 1995; Harvey,1999), though even if effects detected at 4.5 years andin kindergarten disappear, it may be problematicalto assume that they are gone forever. After all,significant relations between time spent in child careand socioemotional adjustment evident when chil-dren were 2 years of age in the NICHD sample wereabsent a year later (NICHD Early Child Care

Research Network, 1998), only to emerge once againa year and a half later and remain evident whenchildren were in kindergarten.

Despite the fact that there remains healthy debateabout the size and meaningfulness of virtually allchild care effects (Scarr, 1998), it must be remem-bered that more and more children are spendingmore and more time at younger and younger ages innonmaternal care arrangements in the United States.Even small effects, when experienced by manychildren, may have broad-scale implications forlarger policy discussions (Fabes, Martin, Hanish, &Updergraff, 2000; Jeffrey, 1989). Indeed, the detectedeffects may have no implications for how anyindividual child should be cared for or how anyindividual family functions, but could have implica-tions at broader levels concerning how classrooms,communities, and even societies operate. Clearly, itwill be important to continue to evaluate the extentto which this is the case for amount of child care andtime in center care, and for low-quality child careand, thus, whether the findings that emerged fromthis inquiry remain throughout the early years ofschool or whether they prove limited to the periodsurrounding the transition to school.

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