Pregnancy Increases Bmi In Adolescents Of A Population-based Birth Cohort Study

Community and International Nutrition

Pregnancy Increases BMI in Adolescents of a Population-Based Birth Cohort1 Denise P. Gigante,*†2 Kathleen M. Rasmussen,† and Cesar G. Victora* *Post-Graduate Program in Epidemiology, Universidade Federal de Pelotas, CP 464, 96001–970, Pelotas, RS, Brazil and †Division of Nutritional Sciences, Cornell University, Ithaca, NY 14853

KEY WORDS:

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adolescence

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pregnancy

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nutritional status

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BMI

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cohort studies

The consequences of teenage pregnancy for adult nutritional status have been little studied in developing countries as a result of the lack of longitudinal studies. Evidence from developed countries suggests that adolescents grow while pregnant and that their growth is associated with increased weight gain and fat storage. Scholl and colleagues (6), in the Camden study, showed that the amount of postpartum retained weight was significantly greater in still-growing gravidas than in other pregnant women. Growth in stature continues after menarche (7,8) and during teenage pregnancy. In the Camden Study, growth in stature was detected on the basis of a knee height measuring device to eliminate the effect of shrinkage in stature during pregnancy. Adolescents had significantly positive increments compared with mature gravidas (9). In Pelotas, a city in southern Brazil, girls have been studied since their birth in 1982. In 2001, the risk factors for childbearing during adolescence were studied in this cohort (10). A total of 16% of the girls who belonged to the cohort study gave birth up to 2001. Anthropometric variables were measured since birth. The purpose of the research reported here was to study the effects of teenage pregnancy on nutritional status at age 19 in a subsample of female adolescents that was followed from 1982 to 2001. The Pelotas Birth Cohort Study allows examining these associations, while accounting for some of the possible confounding factors that have not been considered in many studies.

The prevalence of obesity has increased worldwide as a result of the rapid nutritional transition observed in many countries (1). Among adolescents, the age group between 10 and 19 y old, increased weight and height have been documented even in less-developed regions (2). Data from developed countries suggest that the diet of adolescents puts them at risk for chronic diseases, such as cardiovascular disease, diabetes, cancer, and osteoporosis (3). From the perspective of prevention, the identification of groups of people at risk of becoming obese has been considered an important concern in research. In a systematic review that included studies of childhood predictors of adult obesity, Parsons and colleagues (4) concluded that the lack of longitudinal data from childhood to adulthood is the major research gap. Likewise, many developing countries are experiencing an increase in the rate of adolescent pregnancy. In Brazil, there were 8.0 births per 1000 adolescents in 1980, which increased 14% to 9.1 in 2000. Among all births, the increase in the proportion of adolescent mothers has been even sharper. In 1980, 9.1% of all births were among adolescent mothers, and this proportion more than doubled, to 19.4% in 2000 (5).

1 This study was supported by the Fundac¸a˜o de Amparo a` Pesquisa no Rio Grande do Sul (FAPERGS), by the Coordenac¸a˜o de Aperfeic¸oamento de Pessoal de Nı´vel Superior (CAPES) of Brazil, and by WHO. 2 To whom correspondence should be addressed. E-mail: [email protected].

0022-3166/05 $8.00 © 2005 American Society for Nutritional Sciences. Manuscript received 16 August 2004. Initial review completed 1 September 2004. Revision accepted 11 October 2004. 74

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ABSTRACT Evidence from developed countries suggests that adolescents grow while pregnant and that their growth is associated with increased weight gain and fat storage, but this has never been examined in girls from developing countries. Adolescents born in 1982 in Pelotas, Brazil, are being followed in a birth cohort study. Information on social and biological determinants of nutritional status was collected in early life. Both in 1997 and in 2001, 464 girls were located through household visits, 16% of whom had had a pregnancy in this period. Changes in height, weight, and BMI between 1997 and 2001 were analyzed in relation to the occurrence of pregnancy, after adjustment for previous anthropometric status, as well as social and biological characteristics. The average gains were 2.0 ⫾ 2.0 cm in height, 3.1 ⫾ 5.9 kg in weight, and 0.7 ⫾ 2.2 kg/m2 in BMI. Each pregnancy was associated with a reduction of 0.46 cm on height gain from 1997 to 2001 (P ⫽ 0.02). Girls who became pregnant gained 2.25 kg more than all others (P ⫽ 0.004). There was a clear association between pregnancy and BMI change. A single pregnancy was associated with an increase of 0.81 kg/m2 (P ⫽ 0.01) and 2 or more pregnancies were associated with an increase of 1.58 kg/m2 (P ⫽ 0.02). Teenage pregnancy was associated with an important increase in BMI. Given the growing epidemic of obesity in low- and middle-income countries, particularly among women, efforts to reduce teenage pregnancy may also contribute to preventing overweight. J. Nutr. 135: 74 – 80, 2005.

TEENAGE PREGNANCY AND NUTRITIONAL STATUS

METHODS

naire. In this later questionnaire, information about abortion could be more precise. The higher number of pregnancies was used in the analyses. The outcomes variables were changes in height, weight, and BMI between 1997 and 2001. Seven girls were excluded from the analyses because their change in height or weight was 3 SD or more above or below the mean change for the whole sample; these were likely to represent major measurement errors (16). Statistical methods. To describe the sample, frequency distributions were used to compare main variables collected in earlier phases of the cohort study for all girls and for those who were included in this analysis. The ␹2 test was used to compare these proportions. To examine associations, ANOVA and F test were used to compare mean values of anthropometric variables across different categories of explanatory variables. Multiple linear regression analyses were carried out for changes in height, weight, and BMI between 1997 and 2001. Potential confounding factors, including anthropometric measurements in childhood and age at menarche, were investigated and those that showed an association (P ⬍ 0.2) were taken to the multivariable analyses. Variables were included as continuous in the linear regression models. Age at menarche and teenage pregnancy were also assessed both as categorical variable (⬍12, 12, 13, or 14 y and none pregnancy, 1 or ⱖ2 pregnancies, respectively) and as dichotomous variables (ⱖ12 vs. ⬍12 y and yes vs. no, respectively). Interactions between socioeconomic and nutritional variables included in these models were tested jointly through the product of all variables. The same criterion of significance for confounding factors (P ⬍ 0.2) was used for interactions. Graphical analyses of the residuals were developed for the linear regression assumptions, and colinearity was tested through variance inflation factors analysis (17) and the STATA package (Intercooled Stata 8.0 for Windows, Stata) was used in the analyses. The regression model included all potential confounding variables, and results are shown for age at menarche and the occurrence of teenage pregnancy.

RESULTS The comparison between characteristics of girls included in the analysis and those identified in the original cohort shows that adolescents from the lowest income group, those born with a low birth weight, whose mothers were ⬍30 y old were less likely to have been identified in the recent follow-up and included in this analysis (Table 1). The incidence of pregnancy between 1997 and 2001 was inversely related to family income in 1982 (Table 2). There was a 10-fold difference between girls belonging to families in the lowest and the highest income groups. An inverse association was also observed between birth weight and pregnancy. In addition, pregnancy between 1997 and 2001 was less frequent among taller girls, with a 4-fold difference between the extreme height groups (Table 2). Means and standard deviations of heights and weights were 80.7 ⫾ 5.1 cm and 11.0 ⫾ 1.6 kg at age 2; 97.4 ⫾ 5.2 cm and 15.5 ⫾ 2.4 kg at age 4; 159.5 ⫾ 6.3 cm and 54.7 ⫾ 10.2 kg at age 15; and 161.5 ⫾ 6.3 cm and 58.3 ⫾ 12.9 kg at age 19. Height and weight at ages 2 and 4 were negatively associated with age at menarche—taller and fatter girls had earlier menarche (data not shown). Inverse associations were also observed between age at menarche and attained weights at ages 15 and 19. BMI values were 16.8 ⫾ 1.4 kg/m2, 16.2 ⫾ 1.5 kg/m2, 21.5 ⫾ 3.5 kg/m2, and 22.3 ⫾ 4.5 kg/m2 at ages 2, 4, 15, and 19, respectively. Higher BMI values at ages 4, 15, and 19 were observed among girls who had early menarche [P ⬍ 0.001 (data not shown)]. Family income, birth weight, and overweight in childhood were positively associated with attained height in 2001 (Table 3). An inverse association was observed between teenage

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Study population and sample. In 1982, births of all women living in the urban area of Pelotas were included in a populationbased, birth-cohort study. Information was obtained on 5914 liveborn infants, representing 99% of births in the year. This cohort has been followed several times, and the methods used in earlier phases have been described elsewhere (11–14). The research described here involved female adolescents who belonged to the Cohort Study and who were examined in both 1997 and 2001. In 1997, a systematic sample of 27% of all households in the city was visited in search of adolescents who were born in 1982. It resulted in the identification of 515 female adolescents, who were also visited in 2001. In this latest follow-up, 502 girls were located, representing 97.5% of those located in 1997. Of the subcohort of female adolescents who were seen in both 1997 and 2001, 38 were excluded because they were pregnant at the time of the 2001 interview (n ⫽ 24), had a child before 1997 (n ⫽ 4), had a child in the last 6 mo (n ⫽ 5), or had missing information in anthropometric variables (n ⫽ 5). All analyses were based on the remaining 464 young women. Data collection. In all phases of this cohort study, a group of field workers was trained in interview techniques and anthropometry. The enrollment of the Birth Cohort Study took place in the maternity hospital, and follow-ups were conducted in the girls’ households. Informed consent was obtained in all phases of the study, and the adolescent and her mother gave their consent in this phase of the cohort study. The study was approved by the maternity hospitals’ ethical committees and by the university ethical committee. Information concerning the mother and her pregnancy was obtained at enrollment in 1982 and was collected at the maternity hospitals up to 3 d after birth. Information concerning the children was obtained through standardized questionnaires given to their mothers in 1984 and in 1986 at mean ages of 2 and 4 y, respectively. Mothers of the study subjects were weighed and measured in the maternity hospitals. Their weight at the beginning of pregnancy was obtained by recall or, if available, collected from the antenatal card. Children were weighed in the maternity hospital and in all follow-up contacts. Length or height (as appropriate) was measured at each follow-up. In the 1997 and 2001 follow-up contacts, questionnaires were applied to both the adolescents and their mothers. Mean ages were 15 and 19 y in 1997 and 2001, respectively. In all phases, a fieldwork supervisor repeated 5% of the interviews. These findings were compared with the original, ensuring the quality control. Variables. This analysis included variables from different phases of this cohort study. The variables collected in 1982 were family income (the sum of monthly incomes of all working people living in the household in minimal wage, categorized as ⱕ1 minimal wage, 1.1 to 3, 3.1 to 6, and ⬎6 minimal wages), maternal skin color (white or nonwhite), maternal age (categorized as ⬍20; 20 to 29; and ⱖ30 y), pregestational weight (from the antenatal care register or by asking the mother about her weight before pregnancy), maternal height (measured soon after admission to the maternity hospital), and birth weight (measured in grams immediately after birth by trained interviewers, using regularly calibrated pediatric scales). Children were weighed using portable scales (CMS Weighing Equipment) and had their supine length (1984) or height (1986) measured using boards manufactured locally according to international specifications (AHRTAG). BMI at ages 2 and 4 y were calculated by dividing weight by the square of height measured in 1984 and 1986, respectively. To define overweight in childhood and adolescence, dichotomous variables were used according to the recommended cutoff values by age (15). In 1984, the age of children varied from 12 to 27 mo and in 1986, from 36 to 51 mo. In 1997 and 2001, these ranges were 19 and 20 mo, respectively. The variables collected during adolescence included BMI based on weight and height measured at ages 15 and 19 y, skin color reported by adolescent in 1997, and age at menarche (continuous variable, collected in 1997 and confirmed in 2001, divided into 4 categories: ⬍12, 12, 13, and ⱖ14 y). Whether there was a pregnancy between 1997 and 2001 was obtained by asking the adolescent in the interviewer-applied questionnaire and in a confidential question-

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TABLE 1 Characteristics of the original 1982 Pelotas birth cohort and adolescents included in the analyses. Pelotas, 1982–2001

Variable

Original cohort (n ⫽ 2876)

Included in analysis (n ⫽ 464)

P-value

% 0.04 21.7 46.4 19.1 12.8

15.8 49.0 21.5 13.7

82.5 17.5

84.7 15.3

15.4 57.3 27.4

12.5 54.7 32.8

10.1 26.7 37.9 21.3 4.0

6.7 25.4 37.3 25.2 5.4

83.2 16.8

84.5 15.5

81.4 18.6

82.4 17.6

0.2 0.03

0.05

TABLE 2

0.6 0.8

1 Income expressed as multiple of the minimum wage.

pregnancy and attained height (P ⫽ 0.002). Skin color, age at menarche, and maternal age were not associated with height. Maternal age, birth weight, and overweight in childhood were positively associated with attained weight in 2001, but there was an inverse association with age at menarche (Table 3). BMI in 2001 was positively associated with age of the adolescent’s mother, with overweight during childhood, and with early menarche. There was a trend (P ⫽ 0.07) toward greater BMI among nonwhite adolescents. Height change between 1997 and 2001 was 2.0 ⫾ 2.0 cm. Of the girls, 11% had negative changes in height of up to 3 cm, which were likely due to measurement error; these values were retained in the analyses, but the results were unchanged when they were excluded. The change in weight was 3.1 ⫾ 5.9 kg, and for BMI it was 0.7 ⫾ 2.2 kg/m2. Changes in height were smaller among girls who were overweight in 1986 and for white adolescents (Table 3). A positive association was observed between height change and age at menarche (P ⬍ 0.001). Height change was similar among adolescents who had zero or one pregnancy, but was smaller among those with 2 or more pregnancies. Change in weight was positively associated with age of the adolescent’s mother. Weight change was also greater among adolescents who were not overweight in 1986 (P ⫽ 0.01) and among those with later menarche (P ⬍ 0.001). The effect of pregnancy on weight change was not linear; weight gain was most pronounced among adolescents who had a single pregnancy (Table 3). BMI change between 1997 and 2001 was inversely related to family income and directly related to age at menarche. Teenage pregnancy was positively associated with BMI change (P ⫽ 0.001), with a significant linear trend (Table 3).

Incidence of pregnancy between 1997 and 2001 according to selected covariates included in the analysis. Pelotas, 1982–2001 Variable Family income2 in 1982 (minimum wage) ⱕ1 1.1–3 3.1–6 ⬎6 Maternal age in 1982, y ⬍20 20–29 ⱖ30 Birth weight, g ⬍2500 2500–2999 3000–3499 3500–3999 ⱖ4000 Skin color White Nonwhite Age at menarche, y ⬍12 12 13 ⱖ14 Z score weight for age in 1997 ⬍⫺1 ⫺1 to 1 ⬎1 Z score height for age in 1997 ⬍⫺1 ⫺1 to 1 ⬎1 BMI in 1997, kg/m2 ⬍18.5 18.5–24.9 25–29.9 ⱖ30

n

% Pregnant

73 226 99 63

31.5 19.9 9.1 3.2

58 254 152

12.1 20.9 13.2

31 118 173 117 25

16.1 24.6 17.3 12.8 4.0

373 91

15.8 23.1

119 130 120 95

20.2 16.2 16.7 15.8

64 326 72

17.2 18.4 12.5

109 316 37

22.9 16.8 5.4

77 320 51 15

13.0 19.4 11.8 13.3

P-value1 ⬍0.001 ⬍0.0013

0.07 0.53 0.06 0.023

0.1 0.8 0.43

0.5 0.43 0.05 0.023 0.4 0.93

1 Pearson ␹2 test. 2 Income expressed as number of times the minimum wage. 3 Linear trend.

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Family income1 (minimum wage) ⱕ1 1.1–3 3.1–6 ⬎6 Maternal skin color White Nonwhite Maternal age, y ⬍20 20–29 ⱖ30 Birth weight, g ⬍2500 2500–2999 3000–3499 3500–3999 ⱖ4000 Overweight in 1984 No Yes Overweight in 1986 No Yes

In the adjusted analysis, height changes were significantly larger among girls with later menarche (Table 4). Each year for which menarche was delayed resulted in an increase of 0.47 cm in height (P ⬍ 0.001). When age at menarche was treated as a categorical variable, the effect was restricted to ages 13 y or greater. The dichotomous variable was also significant (P ⫽ 0.001). The continuous variable expressing the number of pregnancies showed that adolescents who had been pregnant were shorter than those who had never been pregnant (P ⫽ 0.02). Girls who were never pregnant were 0.40 cm (P ⫽ 0.11) taller than those with one or more pregnancies. Finally, categorical analysis showed that, compared with girls who were never pregnant, those who had 1 pregnancy were 0.18 cm shorter (P ⫽ 0.5) and those with 2 or more pregnancies were 1.60 cm shorter (P ⫽ 0.006). Therefore, the effect of pregnancy was restricted to those with more than 1 pregnancy. Weight changes were also larger among girls with later menarche (Table 5). Each year for which menarche was delayed, adolescents weighed 0.79 kg more than the others (P ⬍ 0.001). When age at menarche was treated as a categorical

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77

TABLE 3 Height, weight, and BMI at age 19 and height, weight, and BMI change between ages from 15 to 19 according to selected covariates included in the analysis. Pelotas, 1982–2001 Variable

Weight

BMI

Height change

Weight change

BMI change

cm

kg

kg/m2

cm

kg

kg/m2

73 226 99 63

159.3 ⫾ 6.8 161.2 ⫾ 5.9 162.5 ⫾ 6.6 163.8 ⫾ 6.1 ⬍0.001 ⬍0.001

56.1 ⫾ 9.8 59.3 ⫾ 15.1 58.7 ⫾ 11.4 57.0 ⫾ 8.4 0.2 0.9

22.1 ⫾ 3.8 22.8 ⫾ 5.3 22.2 ⫾ 4.0 21.2 ⫾ 2.5 0.10 0.10

2.2 ⫾ 2.1 2.0 ⫾ 2.1 2.2 ⫾ 1.9 1.9 ⫾ 1.9 0.8 0.8

3.7 ⫾ 6.3 3.5 ⫾ 6.1 2.4 ⫾ 6.0 2.5 ⫾ 4.7 0.3 0.09

1.0 ⫾ 2.5 0.8 ⫾ 2.3 0.4 ⫾ 2.3 0.4 ⫾ 1.7 0.2 0.05

58 254 152

160.2 ⫾ 5.8 161.5 ⫾ 6.5 162.0 ⫾ 6.2 0.19 0.09

56.0 ⫾ 10.0 57.5 ⫾ 12.6 60.4 ⫾ 14.0 0.03 0.01

21.8 ⫾ 3.8 22.0 ⫾ 4.5 23.0 ⫾ 4.8 0.09 0.04

1.9 ⫾ 2.0 2.0 ⫾ 2.1 2.2 ⫾ 1.9 0.5 0.2

1.9 ⫾ 5.3 3.0 ⫾ 6.1 3.8 ⫾ 5.9 0.11 0.04

0.3 ⫾ 1.8 0.7 ⫾ 2.4 0.9 ⫾ 2.3 0.3 0.11

31 118 173 117 25

157.5 ⫾ 6.2 159.8 ⫾ 5.8 161.6 ⫾ 6.1 163.1 ⫾ 6.2 166.1 ⫾ 6.1 ⬍0.001 ⬍0.001

53.6 ⫾ 9.6 57.9 ⫾ 13.1 57.5 ⫾ 11.9 60.5 ⫾ 14.1 61.4 ⫾ 11.4 0.04 0.006

21.6 ⫾ 4.4 22.6 ⫾ 4.7 22.0 ⫾ 4.5 22.7 ⫾ 4.6 22.2 ⫾ 4.0 0.6 0.6

2.2 ⫾ 1.8 2.1 ⫾ 2.0 1.9 ⫾ 2.0 2.1 ⫾ 2.1 2.2 ⫾ 2.1 0.9 0.9

2.9 ⫾ 5.4 4.0 ⫾ 7.2 3.0 ⫾ 5.1 3.1 ⫾ 5.8 0.8 ⫾ 11.4 0.18 0.11

0.6 ⫾ 2.2 0.9 ⫾ 2.8 0.6 ⫾ 1.9 0.6 ⫾ 2.1 ⫺0.2 ⫾ 1.9 0.2 0.13

366 67

161.2 ⫾ 6.3 163.4 ⫾ 6.7 0.007

57.3 ⫾ 12.8 63.3 ⫾ 11.9 ⬍0.001

22.0 ⫾ 4.6 23.7 ⫾ 4.2 0.005

2.0 ⫾ 2.1 2.2 ⫾ 1.8 0.6

2.9 ⫾ 5.9 3.4 ⫾ 5.6 0.5

0.6 ⫾ 2.2 0.7 ⫾ 2.2 0.8

356 76

161.4 ⫾ 6.3 163.2 ⫾ 6.4 0.02

57.3 ⫾ 13.2 63.9 ⫾ 10.9 ⬍0.001

22.0 ⫾ 4.7 24.0 ⫾ 4.1 ⬍0.001

2.2 ⫾ 2.1 1.6 ⫾ 1.6 0.02

3.5 ⫾ 5.9 1.6 ⫾ 6.2 0.01

0.8 ⫾ 2.2 0.2 ⫾ 2.4 0.07

373 91

161.6 ⫾ 6.3 161.2 ⫾ 6.4 0.6

57.8 ⫾ 12.2 60.1 ⫾ 15.3 0.13

22.1 ⫾ 4.3 23.1 ⫾ 5.4 0.07

2.0 ⫾ 1.9 2.5 ⫾ 2.4 0.03

2.9 ⫾ 6.0 4.1 ⫾ 5.7 0.10

0.6 ⫾ 2.3 0.9 ⫾ 2.2 0.2

119 130 120 95

160.9 ⫾ 6.8 161.3 ⫾ 6.0 161.9 ⫾ 6.3 162.0 ⫾ 6.3 0.5 0.14

62.5 ⫾ 16.1 57.0 ⫾ 12.3 57.6 ⫾ 10.8 55.7 ⫾ 10.4 ⬍0.001 ⬍0.001

24.0 ⫾ 5.4 21.9 ⫾ 4.7 21.9 ⫾ 3.6 21.2 ⫾ 3.7 ⬍0.001 ⬍0.001

1.5 ⫾ 1.7 1.7 ⫾ 1.9 2.1 ⫾ 1.9 3.2 ⫾ 2.2 ⬍0.001 ⬍0.001

2.5 ⫾ 6.8 1.7 ⫾ 5.0 3.9 ⫾ 5.9 4.9 ⫾ 5.6 ⬍0.001 ⬍0.001

0.6 ⫾ 2.6 0.2 ⫾ 2.0 1.0 ⫾ 2.2 0.9 ⫾ 2.0 0.03 0.05

384 68 12

161.8 ⫾ 6.4 160.7 ⫾ 5.5 155.6 ⫾ 5.9 0.002 0.002

58.2 ⫾ 13.1 58.8 ⫾ 12.0 56.9 ⫾ 12.8 0.9 0.9

22.2 ⫾ 4.5 22.8 ⫾ 4.3 23.6 ⫾ 5.9 0.4 0.17

2.1 ⫾ 2.0 2.2 ⫾ 2.0 0.0 ⫾ 1.4 0.002 0.04

2.7 ⫾ 5.7 5.3 ⫾ 6.8 3.9 ⫾ 7.7 0.005 0.005

0.5 ⫾ 2.1 1.4 ⫾ 2.5 1.7 ⫾ 3.1 0.004 0.001

1 Values are means ⫾ SD. 2 Income expressed as multiple of the minimum wage.

variable, the association was restricted to ages 14 y or more. When expressed as a continuous variable, weight was higher by 1.56 kg per pregnancy (P ⫽ 0.01). When expressed as a dichotomous variable, girls who were never pregnant weighed 2.25 kg less than those who had become pregnant (P ⫽ 0.004). Compared with girls who were never pregnant, those who had 1 pregnancy weighed 2.24 kg more (P ⫽ 0.007). Those with 2 or more pregnancies were 2.31 kg heavier (P ⫽ 0.2). Therefore, the effect of pregnancy was restricted to those with 1 pregnancy (Table 5). BMI change and age at menarche were not significantly associated in either the crude or the adjusted analyses (Table 6). There was a clear association between pregnancy and BMI

change. Girls who had ever been pregnant were 0.92 kg/m2 fatter than all others (P ⫽ 0.002). For each pregnancy, BMI increased by 0.71 kg/m2 (P ⫽ 0.002). Finally, a single pregnancy was associated with an increase of 0.81 kg/m2 (P ⫽ 0.01), and 2 or more pregnancies were associated with an increase of 1.58 kg/m2 (P ⫽ 0.02). No interactions were observed in the models, and graphic analyses of the residuals indicated that the assumptions of linear regression were not violated. Colinearity was found when 2 measurements of weight (1984 and 1986) were included in the multiple linear regression analyses. The results were similar when just 1 or 2 measurements were included in the analyses.

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Family income2 (minimum wage) ⱕ1 1.1–3 3.1–6 ⬎6 t test Linear trend Maternal age, y ⬍20 20–29 ⱖ30 t test Linear trend Birth weight, g ⬍2500 2500–2999 3000–3499 3500–3999 ⱖ4000 t test Linear trend Overweight at age 2 y No Yes t test Overweight at age 4 y No Yes t test Adolescent skin color White Nonwhite t test Age at menarche, y ⬍12 12 13 ⱖ14 t test Linear trend Teenage pregnancy None 1 2 or 3 t test Linear trend

Height1

n

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TABLE 4 Multiple linear regression analysis of the effects of menarche and pregnancy on height change. Pelotas, 1982–20011 Adjusted2

Crude

␤ Age at menarche, y Continuous ⱖ12 vs. ⬍12 ⬍12 12 13 ⱖ14 Pregnancy Continuous Yes vs. no None 1 ⱖ2

SE

P-value

␤

0.48 0.79 0.00 0.22 0.66 1.76

0.06 0.21 — 0.24 0.25 0.26

⬍0.001 ⬍0.001 — 0.4 0.008 ⬍0.001

⫺0.44 ⫺0.25 0.00 0.08 ⫺2.07

0.19 0.25 — 0.26 0.58

0.02 0.3 — 0.8 ⬍0.001

SE

P-value

0.47 0.78 0.00 0.24 0.70 1.70

0.07 0.22 — 0.25 0.26 0.28

⬍0.001 0.001 — 0.4 0.007 ⬍0.001

⫺0.46 ⫺0.40 0.00 ⫺0.18 ⫺1.60

0.19 0.25 — 0.27 0.57

0.02 0.11 — 0.5 0.006

1 n ⫽ 457. 2 Adjusted for skin color, maternal age, weight, and length at 2 y, and for weight and height at 4 y. Pregnancy was also adjusted for age at

menarche as a continuous variable.

In this study, we investigated the occurrence of pregnancy from 1997 to 2001 and the changes on nutritional status during this period in a sample of girls in the Pelotas Birth Cohort Study who have been followed since their birth in 1982. Longitudinal data allowed us to assess the association of teenage pregnancy with nutritional status of female adolescents. Nutritional status was assessed several times since birth, and this analysis included girls who were measured in 1997 and 2001. Very few studies from developing countries permit this type of analysis (18). In addition, information bias was reduced by collecting information close to the time at which events occurred and by collecting weight and height during childhood and adolescence. This study included all girls belonging to the cohort study who were identified in a cluster sample of 27% of all households in the city in 1997. We estimate that these

represent 72% of the cohort adolescents who should have located in this sample of the city, in the absence of outmigration. Losses to follow-up were more common among the poorest families, when the adolescent herself had been born to an adolescent mother, and for low-birth-weight girls, which may be partly explained by survival bias. However, differences between those located and those lost to follow-up were not marked. A more detailed analysis of losses to follow-up is available elsewhere (14). Among female adolescents interviewed in 1997, 96.5% were seen again in 2001. Of all girls included in this study, only 4 adolescents had a term delivery before 1997. These teenage mothers were excluded in this analysis because pregestational nutritional status was unknown. According to the official information systems for live and stillbirth in Pelotas, childbearing during adolescence was reported for 16.2% of all girls belonging to the original cohort.

TABLE 5 Multiple linear regression analysis of the effects of menarche and pregnancy on weight change. Pelotas, 1982–20011 Adjusted2

Crude

Age at menarche, y Continuous ⱖ12 vs. ⬍12 ⬍12 12 13 ⱖ14 Pregnancy Continuous Yes vs. no None 1 ⱖ2

␤

SE

P-value

␤

SE

P-value

0.66 0.85 0.00 ⫺0.79 1.39 2.35

0.20 0.64 — 0.75 0.76 0.81

0.001 0.18 — 0.3 0.07 0.004

0.79 0.99 0.00 ⫺0.66 1.47 2.77

0.21 0.68 — 0.79 0.80 0.87

⬍0.001 0.14 — 0.4 0.07 0.002

1.46 2.34 0.00 2.54 1.17

0.57 0.72 — 0.78 1.73

0.01 0.001 — 0.001 0.5

1.56 2.25 0.00 2.24 2.31

0.62 0.78 — 0.83 1.88

0.01 0.004 — 0.007 0.2

1 n ⫽ 460. 2 Adjusted for family income, skin color, birth weight, weight and length at 2 y, and for weight and height at 4 y. Pregnancy was also adjusted for

age at menarche as a continuous variable.

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DISCUSSION

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TABLE 6 Multiple linear regression analysis of the effects of menarche and pregnancy on BMI change. Pelotas, 1982–20011 Adjusted2

Crude

Age at menarche, y Continuous ⱖ12 vs. ⬍12 ⬍12 12 13 ⱖ14 Pregnancy Continuous Yes vs. no None 1 ⱖ2

␤

SE

P-value

␤

SE

P-value

0.11 0.10 0.00 ⫺0.36 0.39 0.36

0.07 0.24 — 0.28 0.29 0.31

0.13 0.7 — 0.2 0.17 0.2

0.14 0.13 0.00 ⫺0.33 0.37 0.47

0.08 0.25 — 0.30 0.30 0.32

0.07 0.6 — 0.3 0.2 0.15

0.55 0.91 0.00 0.86 1.18

0.32 0.27 — 0.29 0.64

0.09 0.001 — 0.004 0.07

0.71 0.92 0.00 0.81 1.58

0.23 0.30 — 0.31 0.70

0.002 0.002 — 0.01 0.02

1 n ⫽ 454. 2 Adjusted for family income, maternal age, birth weight, weight at age 2 y, and for weight at 4 y. Pregnancy was also adjusted for age at menarche

as a continuous variable.

positive increments in stature were observed in adolescents during pregnancy (7,8,23). Hediger and colleagues (8) have documented that adolescent pregnancy is associated with larger gestational weight gains. They found that, after week 28, growing adolescents on average failed to lose fat and tended to continue to accrue fat in their upper arm fat area. In contrast, mature women and the nongrowing adolescents both lost fat from their upper arms and back. Thus, continued subcutaneous fat accrual after week 28, at a time when reductions in the subcutaneous fat mass are expected, appears to be characteristic of maternal growth during pregnancy. In the present study, linear association between BMI gain and pregnancy was related to both reduced height and increased weight. In another Brazilian study, Sichieri and co-workers (27) used 1996 Demographic Health Survey data on 2297 women aged 20 to 45 y and showed that short stature was associated with an increased risk of weight gain with pregnancy in the developed areas of Brazil. Some limitations must be considered when interpreting these data. Only 12 adolescents had had 2 or more pregnancies, and the possibility of lack of statistical power must be considered, particularly for the analyses of weight change. Also, the 1997 follow-up was restricted to 27% of the urban area, and 28% of the original cohort was lost to follow-up. If we had been able to include all of the female adolescents who were born in 1982 (2876 girls) in this follow-up, the number of adolescents who were pregnant and the power of the study would have been greater, which may have permitted us to identify additional associations with weight change. In summary, the negative effect of pregnancy on height change was restricted to adolescents who had 2 or more pregnancies, while greater gains in weight and BMI were observed among all of those who became pregnant. Both by reducing adult height and increasing overweight, teenage pregnancy may contribute to the prevalence of obesity and the incidence of chronic diseases from adolescence to adulthood (28,29). Using existing data on the distribution of BMI among adult Pelotas women (30), it is possible to estimate the impact of teenage pregnancy. Currently, 54.2% of Pelotas women are overweight. If one assumes a normal distribution of BMI, a shift of 1.58 kg/m2 in a group of this population and, consequently, a shift of 0.29 Z score, would increase the prevalence

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Among girls included in the present analyses, 16.9% delivered a baby between 1997 and 2001. Teenage pregnancy is a time of nutritional risk (19). Adolescents most likely to become pregnant are often those with inadequate nutritional status and unfavorable socioeconomic background (20,21). Scholl and Hediger (22) showed that there is a competition for nutrients between the growing pregnant adolescent and her fetus. The present results confirmed a strong inverse association between pregnancy and change of height, especially among those who had 2 or more pregnancies. Growth in stature continues after menarche and during teenage pregnancy (7,8,23). In this study, mean attained heights at the end of adolescence were similar for girls with zero or one pregnancy, but markedly shorter for those with 2 or more gestations. This suggests that there was a cumulative effect on growth in height that only became apparent for girls with 2 pregnancies. Researchers have found that there is considerable weight gain and increase in subcutaneous fatness at central sites when pregnancy occurs during the final phase of adolescent growth (6,23,24). Some authors have documented that adolescent pregnancy is associated with larger gestational weight gains and increased risk for weight retention (25). The present results showed that having one or more pregnancies during adolescence was associated with weighing 2 kg more at age 19 y than not having any pregnancies. Changes in BMI were a result of the combination of relative gains in weight and height. Girls with a single pregnancy gained more weight but did not gain more height than those who did not get pregnant. On the other hand, those with 2 or more pregnancies gained about the same amount of weight as those with a single pregnancy but gained considerably less height. Thus, BMI change was linearly related to the number of pregnancies. In adult women, the findings are inconclusive with respect to excess weight gain and risk of becoming overweight associated with pregnancy. In adolescents, maternal growth accounts for some of the weight increase during pregnancy (26). Although most of the pubertal gains in stature in girls are around the time of peak height velocity, and, prior to menarche, growth continues during late adolescence, small and

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of overweight in this group to 65.5%. So, efforts to reduce teenage pregnancy may also contribute to prevent overweight, taking into account the growing epidemic of obesity in lowand middle-income countries, particularly among women. LITERATURE CITED

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1. Popkin, B. M. & Doak, C. M. (1998) The obesity epidemic is a worldwide phenomenon. Nutr. Rev. 56: 106 –114. 2. Schneider, D. (2000) International trends in adolescent nutrition. Soc. Sci. Med. 51: 955–967. 3. Lytle, L. A. (2002) Nutritional issues for adolescents. J. Am. Diet. Assoc. 102: S8 –12. 4. Parsons, T. J., Power, C., Logan, S. & Summerbell, C. D. (1999) Childhood predictors of adult obesity: a systematic review. Int. J. Obes. Relat. Metab. Disord. 23: S1–S107. 5. IBGE. Censo Demografico 2000: fecundidade e mortalidade infantil. Resultados preliminares. IBGE, Rio de Janeiro, Brazil. 6. Scholl, T. O., Hediger, M. L., Schall, J. I., Khoo, C. S. & Fisher, R. L. (1994) Maternal growth during pregnancy and the competition for nutrients. Am. J. Clin. Nutr. 60: 183–188. 7. Frisancho, A. R. (1997) Reduction of birth weight among infants born to adolescents: maternal-fetal growth competition. Ann. N.Y. Acad. Sci. 817: 272–280. 8. Hediger, M. L., Scholl, T. O. & Schall, J. I. (1997) Implications of the Camden Study of adolescent pregnancy: interactions among maternal growth, nutritional status, and body composition. Ann. N.Y. Acad. Sci. 817: 281–291. 9. Scholl, T. O., Hediger, M. L. & Schall, J. I. (1997) Maternal growth and fetal growth: pregnancy course and outcome in the Camden Study. Ann. N.Y. Acad. Sci. 817: 292–301. 10. Gigante, D. P., Victora, C. G., Gonc¸alves, H., Lima, R. C., Barros, F. C. & Rasmussen, K. M. (2004) Risk factors of childbearing during adolescence in a population-based birth cohort in southern Brazil. Rev. Panam. Salud Publica 16: 1–10. 11. Victora, C. G., Barros, F. C. & Vaughan, J. P. (1989) . Epidemiologia da Desigualdade. Hucitec, Sao Paulo, Brazil. 12. Barros, F. C., Victora, C. G. & Vaughan, J. P. (1990) The Pelotas birth cohort study, 1982–1987. Strategies for following up 6,000 children in a developing country. Paediatr. Perinat. Epidemiol. 4: 267–282. 13. Barros, F. C., Victora, C. G., Vaughan, J. P., Tomasi, E., Horta, B. L., Cesar, J. A., Menezes, A.M.B., Halpern, R., Post, C. L. & Garcia, M. M. (2001) The epidemiologic transition in maternal and child health in a Brazilian city, 1982–1993: a comparison of two population-based cohorts. Paediatr. Perinat. Epidemiol. 15: 4 –11. 14. Victora, C. G, Barros, F. C., Lima, R. C., Behague, D., Gonc¸alves, H.,

Horta, B. L., Gigante, D. P. & Vaughan, J. P. (2003) The Pelotas (Brazil) Birth Cohort Study, 1982–2001. Cad. Saude Publica 19: 1241–1256. 15. Cole, T. J., Bellizzi, M. C., Flegal, K. M. & Dietz, W. H. (2000) Establishing a standard definition for child overweight and obesity worldwide: international survey. Br. Med. J. 320: 1240 –1243. 16. [WHO] World Health Organization (1995) Physical status: the use and interpretation of anthropometry. Technical Report Series 854. WHO, Geneva, Switzerland. 17. Kleinbaum, D. G., Kupper, L. L., Muleer, K. E. & Nizam, A. (1998) Applied Regression Analysis and Other Multivariate Methods. 3rd ed. Duxbury Press, Pacific Grove, CA. 18. Harpham, T., Huttly, S., Wilson, I. & De Wet, T. (2003) Linking public issues with private troubles: panel studies in developing countries. J. Int. Dev. 15: 1–11. 19. Lederman, S. A. (1997) Nutritional support for the pregnant adolescent. Ann. N.Y. Acad. Sci. 817: 304 –312. 20. Lawlor, D. A. & Shaw, M. (2002) Too much too young? Teenage pregnancy is not a public health problem. Int. J. Epidemiol. 31: 552–554. 21. Smith, S. (2002) Too much too young? In Nepal more a case of too little, too young. Int. J. Epidemiol. 31: 557–558. 22. Scholl, T. O. & Hediger, M. L. (1993) A review of the epidemiology of nutrition and adolescent pregnancy: maternal growth during pregnancy and its effect on the fetus. J. Am. Coll. Nutr. 12: 101–107. 23. Garn, S. M., Lavelle, M., Pesick, S. D. & Ridella, S. A. (1984) Are pregnant teenagers still in rapid growth? Am. J. Dis. Child. 138: 32–34. 24. Scholl, T. O., Hediger, M. L., Schall, J. I., Mead, J. P. & Fischer, R. L. (1995) Maternal growth during adolescent pregnancy. J. Am. Med. Assoc. 274: 26 –27. 25. Olson, C. M., Strawderman, M. S., Hinton, P. S. & Pearson, T. A. (2003) Gestational weight gain and postpartum behaviors associated with weight change from early pregnancy to 1 y postpartum. Int. J. Obes. Relat. Metab. Disord. 27: 117–127. 26. Gunderson, E. P. & Abrams, B. (2000) Epidemiology of gestational weight gain and body weight changes after pregnancy. Epidemiol. Rev. 22: 261–274. 27. Sichieri, R., Silva, C.V.C. & Moura, A. S. (2003) Combined effect of short stature and socioeconomic status on body mass index and weight gain during reproductive age in Brazilian women. Braz. J. Med. Biol. Res. 36: 1319 – 1325. 28. Allebeck, P. & Bergh, C. (1992) Height, body mass index and mortality: do social factors explain the association? Public Health 106: 375–382. 29. Must, A., Jacques, P. F., Dallal, G. E., Bajema, C. J. & Dietz, W. H. (1992) Long-term morbidity and mortality of overweight adolescents. A follow-up of the Harvard Growth Study of 1922 to 1935. N. Engl. J. Med. 327: 1350 –1355. 30. Gigante, D. P., Barros, F. C., Post, C. L. & Olinto, M. T. (1997) Prevalence and risk factors of obesity in adults. Rev. Saude Publica 31: 236 –246.