Younger maternal age (at initiation of childbearing) and recurrent perinatal mortality

Article Outline

• Abstract
• 1. Introduction
• 2. Materials and methods
  • 2.1. Data source
  • 2.2. Cohort composition
  • 2.3. Variables
  • 2.4. Statistical analysis
• 3. Results
• 4. Comment
• Acknowledgements
• References
• Copyright

Abstract 

Objective

To assess whether young maternal age at initiation of childbearing is associated with recurrence of perinatal mortality (PM), as well as its components: stillbirth and neonatal death.

Study design

We conducted a population-based, retrospective cohort study on the Missouri maternally linked longitudinal data files comprising adolescent (10–19 years; n=73,533) or mature (20–24 years; n=78,618) mothers in their first pregnancy with follow-up in their second pregnancy to document the occurrence of PM or its components. The study covered the period 1989–2005. We used unconditional logistic regression modeling to generate odds ratios and to control for confounding.

Results

A history of perinatal mortality, stillbirth, or neonatal mortality increased the risk of a recurrence by 4–5 times. Among women with a history of PM or stillbirth in the first pregnancy, maternal age at initiation of pregnancy was not a risk factor for subsequent PM or its components. However, adolescent mothers with a history of neonatal mortality in the first pregnancy were about 5 times as likely to experience stillbirth in the second pregnancy, as compared to their mature counterparts.

Conclusions

Young maternal age at the initiation of childbearing is not associated with an overall increased risk of recurrent perinatal loss. However, prior history of neonatal mortality among teen mothers is strongly predictive of subsequent stillbirth.

Keywords: Recurrent perinatal mortality, Neonatal death, Stillbirth, Adolescent pregnancy

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1. Introduction 

Adolescent pregnancies have long been associated with increased rates of medical complications, including low birth weight, prematurity, and increased perinatal mortality [1], [2], [3]. In 2005, it was reported that the rates of perinatal mortality (PM) in the United States for mothers in the 15–17, 18–19 and 20–24 age groups were 51%, 32% and 7% higher, respectively, than for women aged 25–29 years [4]. Numerous studies have tried to find whether the elevated risk is due to young age per se or to other predisposing factors [5], [6], [7]. The findings from the literature range from age having an adverse effect [5], [6], [8] to having no effect [9], [10], [11], [12]. Few studies, however, have specifically examined the recurrence of adverse outcomes in age-specific populations, particularly among teens.

Previous research has shown that women experiencing a previous adverse pregnancy outcome are more likely to have recurrence of the negative outcomes in subsequent pregnancies [13]. One study found a 2-fold increase in risk for PM and a 4-fold increase for neonatal mortality among adolescents with a previous adverse event as compared to nulliparous adolescents [14]. A notable limitation of that study was the lack of comparison of the adolescent groups with multiparous mature mothers with a history of perinatal loss. The objective of our study is to assess the impact of young maternal age at initiation of childbirth on the recurrence of PM. Given that teenage mothers are at risk of having repeat pregnancies [15], it is important to understand the role of young age in repeated perinatal losses in order to inform intervention strategies aimed at reducing perinatal mortality in the population.

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2. Materials and methods 

2.1. Data source 

The Missouri maternally linked cohort data files covering the period from 1989 through 2005 were utilized for this study. Live birth, fetal and infant death data have been longitudinally linked to the biological mothers using unique identifiers. The methods used in the linkage process and validation have been reported in detail in previous studies [16], [17].

2.2. Cohort composition 

The data file contained 1,035,547 records between 1989 and 2005. The study sample was limited to 766,710 mothers with two consecutive pregnancies. We sequentially excluded the following records: 37,529 (3.6%) multiple gestations, 41,117 (4.0%) with gestational ages outside the range of 20–44 weeks, and 121,458 (11.7%) infants without siblings. An additional 240,828 (42.5%) records of women whose age at first pregnancy was not between 10 and 24 years were removed from analysis. Furthermore, we removed 21,912 records (6.7%) from the adolescent cohort and 151,198 (46.4%) records from the mature cohort so that the ages at second pregnancy would be no greater than 24 years and 28 years respectively. Other exclusions included unknown demographic variables such as unknown maternal age, missing birth weight, and inter-pregnancy interval of less than 28 days, totaling 517 (0.3%) records. Our final sample consisted of 152,151 sibling-matched birth records.

2.3. Variables 

The main exposure of interest in this study was young maternal age at initiation of childbearing (first pregnancy), defined as ages 10–19 years. We created a comparison group comprised of mature mothers, whom we defined as women aged 20–24 years in their first pregnancy. In previous studies examining the potential negative effects of teen pregnancies, we similarly defined our exposed and comparison groups [8], [18]. It is pertinent to explain that the age of exposure was defined in the first pregnancy knowing that age varies over time. An adolescent woman (<20 years) at her first pregnancy would be some years older in her second. However, since age at initiation of childbearing was the exposure of interest, the same woman would be designated as belonging to the exposed group (<20 years) during her second pregnancy. This approach permits follow-up of the same cohort in the subsequent pregnancy. Furthermore, we have limited the age at second pregnancy to ≤24 years for the exposed group and ≤28 years for the non-exposed to make the cohorts more comparable.

The main outcome of interest was recurrence of PM in the second pregnancy. Perinatal mortality was defined as fetal deaths of 20+ weeks and infant deaths of <28 days of age. We also examined the components of PM, stillbirth and neonatal mortality, individually. Gestational age was computed in weeks as the interval between the last menstrual period and the date of delivery of the fetus. When the estimated gestational age was inconsistent with the observed birth weight, we used the clinical estimate of gestational age [19], [20]. The clinical estimates were less than 5% of the records in the Missouri Vital Statistics.

In analyzing the risk of recurrent perinatal mortality, we examined the following second pregnancy variables that might affect this risk: race/ethnicity, education, marital status, smoking during pregnancy, adequacy of prenatal care, body mass index (BMI) and inter-pregnancy interval. Adequacy of prenatal care was assessed using the revised graduated index (R-GINDEX) to describe the level of prenatal care utilization [21], [22]. This index assesses the adequacy of care on the basis of the trimester when prenatal care began, number of visits and the infant's gestational age at birth. Inadequate prenatal care utilization in the study refers to women who had missing prenatal care information, received sub-optimal prenatal care, or had no prenatal care at all. BMI (weight in kg divided by height in m²) was derived from the mother's height recorded at the first prenatal visit and self-reported pre-pregnancy weight at the first visit of each pregnancy. BMI was categorized according to the Institute of Medicine (IOM) guidelines as the following: <18.5kg/m² (underweight), 18.5–24.9kg/m² (normal weight), 25–29.9kg/m² (overweight) and ≥30.0kg/m² (obese) [23]. Inter-pregnancy interval was calculated as the interval between the birth of the first and second child minus the gestational age of the second child. It was grouped as 1st (≤1.2 years), 2nd (1.3–2.1 years), 3rd (2.2–3.4 years) and 4th (≥3.5 years) quartiles [24]. Other covariates considered were common maternal obstetric complications: pre-eclampsia, eclampsia, chronic hypertension, placental abruption, placenta previa, diabetes (insulin-dependent, and other forms of diabetes) and anemia. A composite variable of obstetric complications was further constructed to denote the presence of at least one of these complications.

2.4. Statistical analysis 

Crude rates for PM and stillbirth were computed as number of events divided by live birth and fetal death counts and multiplied by 1000. Neonatal death rates were expressed as number of deaths within the first 28 days divided by total live births multiplied by 1000. Chi-square test was used to assess crude frequency differences in second pregnancy socio-demographic characteristics and obstetric complications between the two groups. Mature women with and without prior perinatal loss were set as the referent category. We used multivariate logistic regression to generate adjusted odds ratios to approximate relative risks. Variables included in the final model were those that showed a significant association with the exposure of interest in a bivariate analysis. The model, therefore, controlled for these maternal characteristics in the second pregnancy: race, education, marital status, smoking, inadequate prenatal care, BMI and inter-pregnancy interval. We did not adjust for medical/obstetric complications because we regard these as pathways or intermediate variables linking exposure to outcome. Further sub-analysis was conducted for components of recurrent PM: recurrent stillbirth and neonatal mortality. All statistical analyses were performed using SAS version 9.1 (SAS Institute, Cary, NC). The study was approved by the Office of the Institutional Review Board at the University of South Florida.

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3. Results 

Of the 152,151 women analyzed, 73,533 (48.3%) were teen mothers aged 10–19 years at initiation of childbearing. The distribution of second pregnancy characteristics by maternal age at initiation of childbearing is shown in Table 1. Approximately one-third of the teen cohort was still under 20 years of age when they delivered a second time. Socio-demographic features typically associated with increased risk for adverse pregnancy outcomes (e.g., Black race, lower level of education, singleness, inadequate prenatal care, smoking during pregnancy) were more prevalent in teen mothers (P<0.0001) (Table 1). Furthermore, these initially adolescent mothers were more likely to have shorter inter-pregnancy intervals than more mature mothers (P<0.0001).

Table 1. Distribution of maternal characteristics in second pregnancy by maternal age in the first pregnancy, Missouri 1989–2005.

Maternal age at first pregnancy
Characteristics in second pregnancy	10–19 years N=73,533 (48.0%)a	20–24 years N=78,618 (51.7%)a	P-value
Age (years)
10–14	22 (0.03)	b–	<0.0001
15–19	23,678 (32.2)	–
20–24	49,833 (67.8)	36,740 (46.7)
25–29	0 (0.0)	41,878 (53.3)
Race
White	53,902 (73.3)	67,339 (85.7)	<0.0001
Black	18,737 (25.5)	9867 (12.6)
Others	894 (1.2)	1412 (1.8)
Education (years)
<12	29,943 (40.7)	9916 (12.6)	<0.0001
≥12	42,967 (58.4)	68,231 (86.8)
Unknown	623 (0.85)	471 (0.6)
Marital status
Single	41,563 (56.5)	20,478 (26.1)	<0.0001
Married	31,953 (43.5)	58,095 (73.9)
Unknown	17 (0.02)	45 (0.06)
Smoked in pregnancy
No	50,867 (69.2)	62,469 (79.5)	<0.0001
Yes	22,505 (30.6)	15,980 (20.3)
Alcohol use
No	72,866 (99.1)	77,849 (99.0)	0.1658
Yes	480 (0.7)	579 (0.7)
Unknown	187 (0.3)	190 (0.2)
Adequate prenatal care
No	42,472 (57.8)	37,627 (47.9)	<0.0001
Yes	31,061 (42.2)	40,991 (52.1)
BMI
Underweight	7141 (9.7)	5532 (6.8)	<0.0001
Normal weight	36,606 (50.0)	37,770 (48.4)
Overweight	14,939 (20.5)	16,650 (21.4)
Obese	12,178 (16.7)	16,235 (20.8)
Unknown	2123 (2.9)	2004 (2.6)
Inter-pregnancy interval (years)
<1.2	22,227 (30.2)	23,297 (29.6)	<0.0001
1.3–2.1	15,626 (21.3)	20,299 (25.8)
2.2–3.4	16,354 (22.2)	21,000 (26.7)
≥3.5	19,326 (26.3)	14,022 (17.8)

Table 2 shows the distribution of medical and obstetric complications in the second pregnancy across maternal age at initiation of childbearing. There were a number of differences between the two cohorts. A higher proportion of complications due to diabetes (insulin-dependent and other), chronic hypertension, cardiac disease, pre-eclampsia, and placenta abruption were found among the mature cohort, while anemia was more prevalent in the adolescent cohort. However, the frequency of occurrence of a composite variable denoting any of the aforementioned pregnancy complications was higher in the mature compared to teen mothers (8.74% vs. 8.25%, respectively, P=0.0006) (data not shown).

Table 2. Pregnancy complications in the second pregnancy by maternal age in the first pregnancy, Missouri 1989–2005.

Maternal age at first pregnancy
Complications in second pregnancy	10–19 years N=73,533 (%)*	20–24 years N=78,618 (%)*	P-value
Anemia
Yes	1673 (2.3)	1211 (1.5)	<0.0001
No	71,828 (97.7)	77,371 (98.4)
Unknown	36 (0.05)	32 (0.04)
Insulin-dependent diabetes
Yes	318 (0.4)	547 (0.7)	<0.0001
No	73,183 (99.5)	78,035 (99.3)
Unknown	32 (0.04)	36 (0.05)
Other diabetes
Yes	852 (1.2)	1388 (1.8)	<0.0001
No	72,649 (98.8)	77,194 (98.2)
Unknown	32 (0.05)	36 (0.04)
Chronic hypertension
Yes	341 (0.5)	598 (0.8)	<0.0001
No	73,160 (99.5)	77,984 (99.2)
Unknown	32 (0.04)	36 (0.05)
Cardiac disease
Yes	306 (0.4)	437 (0.6)	0.0005
No	73,195 (99.5)	87,142 (99.4)
Unknown	33 (0.04)	37 (0.04)
Pre-eclampsia
Yes	2133 (2.9)	2492 (3.2)	0.0092
No	71,368 (97.1)	76,090 (96.8)
Unknown	32 (0.04)	36 (0.05)
Eclampsia
Yes	63 (0.09)	45 (0.06)	0.1121
No	73,438 (99.9)	78,537 (99.9)
Unknown	32 (0.05)	36 (0.04)
Placenta abruption
Yes	616 (0.8)	570 (0.7)	0.0072
No	72,900 (99.1)	78,016 (99.2)
Unknown	17 (0.02)	32 (0.04)
Renal disease
Yes	236 (0.3)	188 (0.2)	0.0101
No	73,265 (99.6)	78,394 (99.7)
Unknown	32 (0.04)	36 (0.05)

Table 3 summarizes the risk for perinatal mortality, stillbirth, and neonatal mortality in the second pregnancy among adolescent mothers. For stillbirth, adjusted risk estimate was lower for adolescent mothers (P=0.03). The two groups had, however, comparable risk levels for perinatal mortality and neonatal mortality (P=0.11 and P=0.92, respectively). Table 4 presents the association between maternal age at initiation of childbirth and the recurrence of PM. A history of PM, stillbirth, or neonatal mortality increased the risk of a recurrent mortality by 4–5 times compared to not having a history of PM or its components. Among women with a history of PM or stillbirth in the first pregnancy, maternal age at initiation of pregnancy was not a risk factor for subsequent PM or its components. However, adolescent mothers with a history of neonatal mortality in the first pregnancy were about 5 times as likely to experience stillbirth in the second pregnancy as compared to their mature counterparts. We re-analyzed the data to control for maternal obstetric complications; the results did not change.

Table 3. Risk of perinatal mortality in the second pregnancy in association with low maternal age at initiation of childbirth, Missouri 1989–2005.

Table 4. Risk of second pregnancy perinatal mortality (PM) in association with low maternal age at initiation of childbirth stratified by history of PM, Missouri 1989–2005.

	Age at 1st pregnancy	Perinatal mortality		Stillbirth		Neonatal mortality
		Nb	Adjusted OR (95% CI)a	Nb	Adjusted OR (95% CI)	Nb	Adjusted OR (95% CI)
Hx of perinatal mortality
Overallc		58	5.13 (3.91–6.73)*	34	5.23 (3.66–7.46)*	25	4.78 (3.18–7.19)*
Yes	10–19	28	0.92 (0.51–1.66)	16	0.99 (0.46–2.12)	12	0.83 (0.33–2.02)
	20–24	30	1.00 (reference)	17	1.00 (reference)	13	1.00 (reference)
No	10–19	571	0.91 (0.80–1.03)	298	0.83 (0.70–0.98)	273	1.03 (0.85–1.23)
	20–24	540	1.00 (reference)	318	1.00 (reference)	222	1.00 (reference)
Hx of stillbirth
Overall		27	4.90 (3.31–7.25)*	18	5.87 (3.64–9.47)*	9	3.53 (1.81–6.86)**
Yes	10–19	8	0.44 (0.18–1.10)	4	0.28 (0.08–0.93)	4	1.01 (0.24–4.21)
	20–24	19	1.00 (reference)	14	1.00 (reference)	5	1.00 (reference)
No	10–19	591	0.92 (0.81–1.05)	310	0.85 (0.72–1.01)	281	1.01 (0.84–1.23)
	20–24	551	1.00 (reference)	321	1.00 (reference)	230	1.00 (reference)
Hx of neonatal mortality
Overall		31	5.10 (3.54–7.36)*	16	4.40 (2.62–7.39)*	16	5.73 (3.46–9.50)***
Yes	10–19	20	1.89 (0.81–4.46)	12	4.94 (1.24–19.59)**	8	0.75 (0.23–2.42)
	20–24	11	1.00 (reference)	3	1.00 (reference)	8	1.00 (reference)
No	10–19	579	0.89 (0.78–1.01)	302	0.80 (0.68–0.95)	277	1.02 (0.84–1.24)
	20–24	559	1.00 (reference)	332	1.0 (reference)	227	1.00 (reference)

Hx, History.

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4. Comment 

Our findings indicate that the risk for perinatal and neonatal mortality is comparable for the adolescent and mature cohorts. As for stillbirth, we observed a minimal reduction in risk in the second pregnancy for mothers who initiated childbearing at ages 10–19 years compared to mothers who were 20–24 at their first pregnancies. Although previous studies have demonstrated that adolescents have an increased risk of PM compared to mature gravidae [8], the present study does not indicate that the age of initiation of childbearing is a factor in the association. Previous research has shown that the risk is not intrinsic in young maternal age alone [9], [10], [11] rather adverse risk outcomes are mainly related to poor socio-economic and behavioral factors [12], [25], [26]. When predisposing factors such as inadequate prenatal care are improved, evidence has shown offspring of teenagers do not have elevated risks of low birth weight or high mortality [27], [28].

Another pertinent finding in this study is the impact of previous history on subsequent perinatal outcomes. Our findings indicate an overall 4- to 5-fold increased risk for PM in women initiating childbearing between the ages of 10 and 24 years and having a history of PM. However, when the data were analyzed based on adolescence versus non-adolescence, no significantly increased risks were revealed, except for mothers with a previous history of neonatal mortality. Overall, however, prior history of perinatal event, rather than maternal age, was more predictive of subsequent PM. These findings are in agreement with previous reports on successive pregnancy outcomes [29], [30].

The linkage between events in one pregnancy to those of a subsequent gestation has been explained theoretically in what is known as “event memory hypothesis,” which postulates that fetal death repetition occurs as a result of memory recall of programmed apoptosis cascade [31]. The new theory posits that a woman's birth outcome in her initial pregnancy tends to be repeated completely, partially or as a delayed event in a subsequent pregnancy. However, the dataset used for this analysis does not contain relevant information that would allow us to delineate the pathways through which recurrence of PM could have occurred, thus creating an important limitation of this study.

In recent years, the life course perspective has emerged as a compelling framework in the inquiry of plausible explanations for disparities in pregnancy outcomes [31], [32]. This longitudinal approach posits that exposures not only during pregnancy, but also throughout the life cycle, may result in stress responses and physiologic changes that affect reproductive health experiences, including pregnancy outcomes [31], [32], [33]. Analyses utilizing this perspective incorporate contextual data on social networks, social supports, and societal factors, such as poverty, racism, access to health care and education, insecurity in communities and neighborhoods, parental, familial, and peer influences, and media exposures. Furthermore, the paternal role on maternal and infant health is a growing area of investigation [34], [35], [36], [37]. The dataset utilized in this study lacks the information to conduct analyses of this type.

The clinical implications of these findings are obvious. While young maternal age has been consistently highlighted as a risk factor for adverse pregnancy outcome, our analysis shows that this does not hold for recurrence of the outcome subsequently, except for a previous history of neonatal mortality. Our results could probably be explained by maternal age optimization in the subsequent pregnancy. In the clinical setting, therefore, counseling of mothers who are adolescents and have experienced an adverse event should emphasize the role of the adverse event rather than the young maternal age per se.

As previously reported, recurrence risk is limited by selection bias such as a couple's decision to achieve a desired family size, resulting in selective fertility [38]. Specifically, couples that experience perinatal losses in the first pregnancy tend to go on to have additional pregnancies in order to achieve a desired family size. This phenomenon operates across all parity levels, but is particularly a concern in higher-order births and older women. It is suggested that selective fertility leads to an overrepresentation of high-risk women at higher birth orders [38], leading to skewed positive results. Our data could not determine the extent to which unmeasured pregnancy intendedness impacted our results. It is, however, reasonable to suspect that unplanned pregnancy would be greater among adolescents than mature women. Hence, the fact that maternal age (given the association with pregnancy intention) did not predict subsequent adverse birth outcome (with one exception) represents some assurance that selection bias resulting from desired family size could not have explained our results. Another limitation of our analysis is that certain sub-group analyses might have suffered from lack of power due to paucity of numbers for the outcomes. This is, especially, true for our analyses on the impact of previous stillbirth and differential outcomes based on maternal age in the second pregnancy. Therefore, such associations may more likely be due to chance.

Despite these limitations, the study bears several points of merit. To our knowledge, no studies have previously evaluated the impact of young maternal age at childbirth initiation on recurrence of perinatal mortality. Our findings represent an important addition to current knowledge in this area. Another point of strength in the study lies in the population-based nature of vital records, which reduced the likelihood of selection bias as compared to data collected from referral facilities or centers. The findings can, therefore, be reasonably generalized. Furthermore, the large volume of records studied permitted the finding of sufficient cases of recurrent PM. While the number of cases was few, the detection of significant differences is evidence of a strong association. Little is known about recurrent PM and even less is known about its relationship to adolescent age at initiation of childbirth. Our findings could therefore, serve as an impetus for further studies in this area.

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Acknowledgements 

This study was funded by a grant from the Flight Attendant Medical Research Institute (FAMRI: 024008) to the first author (Hamisu Salihu, MD, PhD). The funding agency did not play any role in any aspect of the study. We thank the Missouri Department of Health and Senior Services for providing the data files used in this study.

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References 

1. Hediger ML, Scholl TO, Schall JI, Krueger PM. Young maternal age and preterm labor. Ann Epidemiol. 1997;7(6):400–406
   • View In Article
   • Abstract
   • Full-Text PDF (1012 KB)
   • CrossRef
2. Olausson PO, Cnattingius S, Haglund B. Teenage pregnancies and risk of late fetal death and infant mortality. Br J Obstet Gynaecol. 1999;106(2):1116–1121
   • View In Article
   • MEDLINE
3. Haldre K, Rahu K, Harro H, Rahu M. Is a poor pregnancy outcome related to young maternal age? A study of teenagers in Estonia during the period of major socio-economic changes (from 1992 to 2002). Eur J Obstet Gynecol Reprod Biol. 2007;131(1):145–151
   • View In Article
4. MacDorman MF, Kirmeyer S. Fetal and perinatal mortality, United States, 2005. Natl Vital Stat Rep. 2009;57(8):1–19
   • View In Article
5. Chen XK, Wen SW, Fleming N, Demissie K, Rhoads GG, Walker M. Teenage pregnancy and adverse birth outcomes: a large population based retrospective cohort study. Int J Epidemiol. 2007;2:368–373
   • View In Article
6. Fraser AM, Brockert JE, Ward RH. Association of young maternal age with adverse reproductive outcomes. N Engl J Med. 1995;332(17):1113–1117
   • View In Article
   • MEDLINE
   • CrossRef
7. Orvos H, Nyirati I, Hajdú J, Pál A, Nyári T, Kovács L. Is adolescent pregnancy associated with adverse perinatal outcome?. J Perinat Med. 1999;27(3):199–203
   • View In Article
   • MEDLINE
   • CrossRef
8. Wilson RE, Alio AP, Kirby RS, Salihu HM. Young maternal age and risk of intrapartum stillbirth. Arch Gynecol Obstet. 2008;278(3):231–236
   • View In Article
   • CrossRef
9. Scholl TO, Hediger ML, Huang J, Johnson FE, Smith W, Ances IG. Young maternal age and parity. Influences on pregnancy outcome. Ann Epidemiol. 1992;2:565–575
   • View In Article
   • MEDLINE
   • CrossRef
10. Lee MC, Suhng LA, Lu TH, Chou MC. Association of parental characteristics with adverse outcomes of adolescent pregnancy. Fam Pract. 1998;15:336–342
    • View In Article
    • MEDLINE
    • CrossRef
11. Reichman NE, Pagnini DL. Maternal age and birth outcomes: data from New Jersey. Fam Plann Perspect. 1997;29(268–72):295
    • View In Article
    • CrossRef
12. Makinson C. The health consequences of teenage fertility. Fam Plann Perspect. 1985;17:132–139
    • View In Article
    • MEDLINE
    • CrossRef
13. Lykke JA, Paidas MJ, Langhoff-Roos J. Recurring complications in second pregnancy. Obstet Gynecol. 2009;113(6):1217–1224
    • View In Article
14. Reime B, Schucking BA, Wenzlaff P. Reproductive outcomes in adolescents who had a previous birth or an induced abortion compared to adolescents’ first pregnancies. BMC Pregnancy Childbirth. 2008;8:4
    • View In Article
    • CrossRef
15. Boardman LA, Allsworth J, Phipps MG, Lapane KL. Risk factors for unintended versus intended rapid repeat pregnancies among adolescents. J Adolesc Health. 2006;39(4):597.e1–597.e8
    • View In Article
    • CrossRef
16. Herman AA, McCarthy BJ, Bakewell JM. Data linkage methods used in maternally-linked birth and infant death surveillance data sets from the United States (Georgia, Missouri, Utah and Washington State), Israel, Norway, Scotland and Western Australia. Paediatr Perinat Epidemiol. 1997;11(Suppl. 1):5–22
    • View In Article
17. Martin J, Curtin S, Saulnier M, Mousavi J. Development of the matched multiple birth file. In: 1995–1998 matched multiple birth dataset. NCHS CD-ROM series 21, no. 13a. Hyattsville, MD: National Center for Health Statistics; 2003;
    • View In Article
18. Aliyu MH, Salihu HM, Alio AP, Wilson RE, Chakrabarty S, Clayton HB. Prenatal smoking among adolescents and risk of fetal demise before and during labor. J Pediatr Adolesc Gynecol. 2010;(February):[Epub ahead of print]
    • View In Article
19. lexander GR, Himes JH, Kaufman RB, Mor J, Kogan M. A United States national reference for fetal growth. Obstet Gynecol. 1996;87(2):163–168
    • View In Article
    • MEDLINE
    • CrossRef
20. affel S, Johnson D, Heuser R. A method of imputing length of gestation on birth certificates. Vital Health Stat 2. 1982;93:1–11
    • View In Article
21. Alexander GR, Cornely DA. Prenatal care utilization: its measurement and relationship to pregnancy outcome. Am J Prev Med. 1987;3(5):243–253
    • View In Article
    • MEDLINE
22. Alexander GR, Kotelchuck M. Quantifying the adequacy of prenatal care: a comparison of indices. Public Health Rep. 1996;111(5):408–418
    • View In Article
    • MEDLINE
23. Parker JD, Abrams B. Prenatal weight gain advice: an examination of the recent prenatal weight gain recommendations of the Institute of Medicine. Obstet Gynecol. 1992;79(5 (Pt 1)):664–669
    • View In Article
    • MEDLINE
24. Al-Jasmi F, Al-Mansoor F, Alsheiba A, Carter AO, Carter TP, Hossain MM. Effect of interpregnancy interval on risk of spontaneous preterm birth in Emirati women, United Arab Emirates. Bull World Health Organ. 2002;80(11):871–875
    • View In Article
    • MEDLINE
25. Cunnington A. What's so bad about teenage pregnancy?. J Fam Plann Reprod Health Care. 2001;27:36–41
    • View In Article
    • MEDLINE
    • CrossRef
26. Hemminki E, Gissler M. Births by younger and older mothers in a population with late and regulated childbearing: Finland 1991. Acta Obstet Gynecol Scand. 1996;75(1):19–27
    • View In Article
    • MEDLINE
    • CrossRef
27. Olds DL, Henderson CR, Tatelbaum R, Chamberlin R. Improving the delivery of prenatal care and outcomes of pregnancy: a randomized trial of nurse home visitation. Pediatrics. 1986;77(1):16–28
    • View In Article
    • MEDLINE
28. Neeson JD, Patterson KA, Mercer RT, May KA. Pregnancy outcome for adolescents receiving prenatal care by nurse practitioners in extended roles. J Adolesc Health Care. 1983;4(2):94–99
    • View In Article
    • Abstract
    • Full-Text PDF (478 KB)
    • CrossRef
29. Sharma PP, Salihu HM, Kirby RS. Stillbirth recurrence in a population of relatively low risk mothers. Paediatr Perinat Epidemiol. 2007;21(Suppl. 1):24–30
    • View In Article
30. Smith GC, Pell JP. Teenage pregnancy and risk of adverse perinatal outcomes associated with first and second birth: population based retrospective cohort study. BMJ. 2001;323(7311):476
    • View In Article
    • MEDLINE
    • CrossRef
31. Salihu HM. Fetal death repetition: the event memory hypothesis. Med Hypotheses. 2008;70(3):567–571
    • View In Article
    • Abstract
    • Full Text
    • Full-Text PDF (99 KB)
    • CrossRef
32. Lu MC, Halfon N. Racial and ethnic disparities in birth outcomes: a life-course perspective. Matern Child Health J. 2003;7(1):1092–7875
    • View In Article
33. Misra DP, Guyer B, Allston A. Integrated perinatal health framework: a multiple determinants model with a life span approach. Am J Prev Med. 2003;25(1):65–75
    • View In Article
    • Abstract
    • Full Text
    • Full-Text PDF (187 KB)
    • CrossRef
34. Rich-Edwards JW, Grizzard TA. Psychosocial stress and neuroendocrine mechanisms in preterm delivery. AJOG. 2005;192(5S):S30–S35
    • View In Article
35. Alio AP, Mbah AK, Kornosky JL, Wathington D, Marty PJ, Salihu HM. Assessing the impact of paternal involvement on racial/ethnic disparities in infant mortality rates. J Commun Health. 2010;[Epub ahead of print]
    • View In Article
36. Alio AP, Salihu HM, Kornosky JL, Richman AM, Marty PJ. Feto-infant health and survival: does paternal involvement matter?. Matern Child Health J. 2009;[Epub ahead of print]
    • View In Article
37. Alio AP, Kornosky JL, Mbah AK, Marty PJ, Salihu HM. The impact of paternal involvement on feto-infant morbidity among Whites, Blacks and Hispanics. Matern Child Health J. 2009;[Epub ahead of print]
    • View In Article
38. Ananth CV. Epidemiologic approaches for studying recurrent pregnancy outcomes: challenges and implications for research. Semin Perinatol. 2007;31(3):196–201
    • View In Article
    • Abstract
    • Full Text
    • Full-Text PDF (98 KB)
    • CrossRef