Epidemiology
DOI: 10.1111/1471-0528.12883 www.bjog.org
Temporal trends in stillbirth in the United States, 1992–2004: a population-based cohort study X Zhang,a MS Kramera,b a Department of Pediatrics, b Department of Epidemiology, Biostatistics and Occupational Health, Faculty of Medicine, McGill University, Montreal, QC, Canada Correspondence: MS Kramer, 2300 Tupper Street (Les Tourelles), Montreal, QC, Canada H3H 1P3. Email [email protected]
Accepted 2 March 2014. Published Online 27 May 2014.
Objective To examine temporal trends in stillbirth and its risk
factors in the United States (US), and to assess the contribution of labour induction and caesarean delivery to the stillbirth rate. Design Population-based cohort study based on linked
birth-infant death and fetal death data files from the US National Vital Statistics System. Setting Complete data were available for 44 states and the District
of Columbia. Population or Sample Singleton births from 1992 to 2004. Methods We assessed changes in stillbirth rates from 1992–1994 to 2002–2004 before and after adjustment for changes in maternal characteristics including maternal age, education, smoking, and medical risk factors, using Cox regression models. We also carried out an ecological study, using states as the units of analysis, to assess the impact on the stillbirth rate of increasing induction and caesarean delivery. Race-specific subgroup analyses were performed and included non-Hispanic Whites and non-Hispanic Blacks.
Main outcome measure Stillbirth rate. Results The stillbirth rate among non-Hispanic White singleton births decreased 11.5% from 1992–1994 (5.2 per 1000) to 2002– 2004 (4.6 per 1000). After adjustment for maternal risk factors, the hazard ratio (HR) for 2002–2004 was 1.01 (0.99, 1.03) for gestational age (GA) ≤39 weeks, but 0.92 (0.86, 0.99) at 40 or more weeks. The ecologic analysis revealed a nonsignificant negative correlation of 0.17 ( 0.44, 0.13) between state-level changes in stillbirth at GA ≥40 weeks and labour induction. A nonsignificant positive correlation of 0.23 ( 0.07, 0.49) was observed between changes in stillbirth at all GAs and caesarean delivery and did not differ at GA ≤39 versus ≥40 weeks. Results were similar among non-Hispanic Blacks. Conclusions Changes in maternal risk factors explained the
reduction in stillbirth at GA ≤39 weeks but not at ≥40 weeks. The rise in labour induction and caesarean delivery rates did not explain the reduction in stillbirth ≥40 weeks of gestation. Keywords Caesarean delivery, labour induction, stillbirth.
Please cite this paper as: Zhang X, Kramer MS. Temporal trends in stillbirth in the United States, 1992–2004: a population-based cohort study. BJOG 2014; 121:1229–1236.
In the last two decades, the rate of preterm birth (41 weeks reduces the risks of stillbirth and perinatal mortality.18 We are aware of only two randomised trials of induction versus expectant management in late preterm or early term gestations: one in women with gestational hypertension, the other in women with gestational diabetes.19,20 Both trials were small, however, and neither observed any stillbirths. To our knowledge, no randomised trials at any gestational age have compared prelabour cesarean delivery with expectant management.
1229
Zhang, Kramer
In this study, we examine the temporal trends in stillbirth in the United States from 1992 to 2004, compare stillbirth rates between 1992–94 and 2002–2004, and assess the impacts on those trends of contemporaneous changes in maternal risk factors and in rates of labour induction and caesarean delivery.
Methods Data source and study population Our study is based on linked birth-infant death and fetal death data files from the National Vital Statistics System of the United States, which are available online from Centers for Disease Control and Prevention (CDC) (www.cdc.gov/ nchs/data_access/Vitalstatsonline.htm). The linked birth– infant death data provide demographic and health data for births occurring during a calendar year, based on information abstracted from birth certificates filed in vital statistics offices of each state and the District of Columbia.21 The fetal death files include information from all reports of fetal deaths.11 Most states report fetal deaths at 20 weeks or more of gestation and/or ≥350 g in birthweight,11 but a few states report fetal deaths for all periods of gestation.11 Demographic data include variables such as date of birth, age and educational attainment of the parents, marital status, live-birth order, race, sex, and geographic area. Health data include items such as birthweight, gestational age and prenatal care. Since 1989, a clinical estimate of gestational age has also been recorded,2 as has the use of labour induction.2,21 Data on caesarean delivery have also become available since the 1989 revision, but no information is recorded as to whether the caesarean was performed before or after the onset of labour.21 Until 2004, linked live birth and fetal death data files included geographic data, including state of birth occurrence; since 2005, however, those geographic data are no longer publicly available. Because a substantial number of states did not report the new (1989) data items from 1989 to 1991,10,22 our study is based on the singleton live birth and fetal death data from 1992 through 2004, the last year when the state identification data were publicly available. Six states were excluded from our analyses for all time periods, owing to missing data on Hispanic origin in fetal death data files in the 1992–94 period: Louisiana, Maryland, Massachusetts, New Hampshire, Oklahoma and Rhode Island. Moreover, Maryland and Oklahoma had missing LMP estimates of gestational age in >40% of fetal deaths and did not report a clinical estimate of gestational age; Louisiana did not report plurality on fetal death data files. Our overall study sample therefore comprised 6 650 475 births in the 1992–94 period and 6 114 413 births in the 2002–2004 period from 44 states and the District of Columbia, 91.8% of total singleton deliveries
1230
among non-Hispanic Whites in the two periods. The following ten states were also excluded for multivariate analysis, owing to missing data on maternal medical risk factors or smoking for >40% of fetal deaths: Alabama, California, Connecticut, District of Columbia, Hawaii, Indiana, New York, South Dakota, Texas and Virginia. This reduced sample for multivariate analysis comprised 4 536 115 births in the 1992–94 period and 4 275 460 births in the 2002– 2004 period, 64.5% of total singleton deliveries among non-Hispanic Whites in the two periods.
Gestational age estimation From the US birth certificate, gestational age (GA) is usually calculated from the first day of the mother’s last menstrual period (LMP). Gestational age derived from the LMP estimate is prone to error.23,24 A clinical estimate of gestation (CE) has also been available since the 1989 revision, except for the state of California. Although the CE is based on the managing clinician’s best estimate, including menstrual history, physical findings, laboratory values, and (if available) sonography,2,3 no instructions were provided prior to the 2003 revision of the US birth certificate for specifying the basis of the CE.2 Several methods for validating the LMP-based gestational age have been proposed to reduce misclassification, including that of Platt et al.,25 Zhang and Bowes,26 Alexander et al.27 and Qin et al.28 In Qin et al.’s28 ‘LMP/CE’ editing method, LMP is replaced by the CE when the two estimates differ by more than 2 weeks. The LMP/ CE method has been shown to reduce misclassification of GA28 and was the basis for gestational age estimation in this study. California did not report a clinical estimate before 2005, and thus LMP cannot be edited in that state.
Statistical analysis Our principal analysis was based on singleton births among non-Hispanic Whites. The effect of increased induction might be less effective in reducing stillbirth than it would be in Blacks or Hispanics because Whites already have high induction rates. The restriction by race was also intended to control for potential confounding due to differences in induction and caesarean rates by race6 and the changing racial composition over time. We compared stillbirth rates, as well as maternal demographic and medical risk factors, between two time periods 10 years apart: 2002–2004 versus 1992–94. As all surviving fetuses are at risk of subsequent stillbirth, our analytical approach was based on fetuses at risk rather than live births at each gestational week. In analyses based on individual women, Cox regression models were used to adjust for fetal sex, parity, and maternal demographic and medical risk factors including maternal age, education, smoking, diabetes, chronic and gestational hypertension. The association between stillbirth and labour induction or caesarean delivery cannot be ascertained in analyses
ª 2014 Royal College of Obstetricians and Gynaecologists
Temporal trends in stillbirth
based on individual women because it would be highly confounded by the medical indication for the procedure. Although maternal risk factors are reported on the US birth certificate, the specific clinical indications for induction or caesarean are not, and thus they cannot be controlled for adequately at the individual level. To reduce confounding by clinical indication, we carried out an ecological analysis10,29–31 to assess the impacts of labour induction and caesarean delivery on stillbirth, based on 45 ecologic units in our overall study sample: 44 states plus the District of Columbia. The ecologic study is based on changes in rates of stillbirth, labour induction, and caesarean delivery between 1992–94 and 2002–2004; ecologic correlations between the change in rates of stillbirth and of labour induction and caesarean delivery were calculated, with weighting for the number of births in each state. Although a few states have implemented the 2003 revision of birth certificates since 2003, coding for the 1989 revision was also available in those states, and the comparison between the two periods was therefore based on the 1989 coding. As a secondary analysis to further reduce confounding by the clinical indication for labour induction or caesarean delivery, we restricted the study sample to women with low-risk pregnancies, defined as maternal age 20–34 years and the absence of diabetes, chronic hypertension and pregnancy-induced hypertension.32–34 This classification of low risk has been used consistently in our previous studies7,35 and by other investigators,32–34 both in the USA and Canada. All other pregnancies were considered not to be at low risk. Ecologic analysis was repeated and stratified by low versus
nonlow risk. Finally, to assess the generalisability of our findings and examine effect modification by race, we repeated our analysis among non-Hispanic Blacks. All data were analysed using SAS version 9.2 (SAS Institute, Cary, NC, USA).
Results Maternal demographic and clinical characteristics As shown in Supporting Information Table S1 for the overall study sample, mothers were older among 2002–2004 births than among 1992–94 births; the proportion of mothers ≥35 years increased from 11.4 to 14.3%. Mothers had higher education in the later period, with the proportion of college degree or higher increasing from 24.7% in 1992–94 to 30.8% in 2002–2004, Maternal smoking was reduced from 15.2% in 1992–94 to 12.1% in 2002–2004, while the proportion with medical risk factors increased; the prevalence of diabetes, chronic hypertension, and gestational hypertension increased from 2.6, 0.6, and 3.3%, respectively, to 3.2, 0.9, and 4.1%, respectively. Labour induction increased by 61% between the two time periods, from 16.0% in 1992–94 to 25.7% in 2002–2004, while caesarean delivery increased by 23%, from 20.9% to 25.7%. Maternal characteristics and risk factors in the reduced sample for regression analysis were similar to those in the overall study sample in the two periods (Table S1).
Main findings Figure 1 shows the GA-specific stillbirth rates, based on fetuses at risk, for the two time periods. As shown in
Risk per 10,000 fetuses alive at each GA
15
NHW 1992-4 NHW 2002-4
10
5
0
20 21 22 23 24 25 26 27 28 29 30 31 32 33 34 35 36 37 38 39 40 41 42+
Gestational age in completed weeks Figure 1. GA-specific stillbirth rates based on fetuses at risk among US non-Hispanic White (NHW) singleton births. Six states (Louisiana, Maryland, Massachusetts, New Hampshire, Oklahoma and Rhode Island) were excluded, owing to missing data on gestational age and Hispanic origin in fetal death data files.
ª 2014 Royal College of Obstetricians and Gynaecologists
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Zhang, Kramer
Table S1, the stillbirth rate among non-Hispanic White singleton births decreased from 5.2 per 1000 total births in 1992–94 to 4.6 per 1000 in 2002–2004, a decrease of 11.5% between the two study periods. The stillbirth rate decreased modestly for GA ≤39 completed weeks over the two periods, from 4.7 per 1000 fetuses at risk in 1992–94 to 4.3 per 1000 fetuses at risk in 2002–2004 respectively, a relative 8.5% decrease; while for GA ≥40 completed weeks, the stillbirth rate decreased from 1.0 per 1000 in 1992–94 to 0.8 per 1000 in 2002–2004, a 20% relative reduction. Unadjusted hazard ratios (HR) for stillbirth (2002–2004 versus 1992–94) and their 95% confidence intervals (CI) were 0.92 (0.91, 0.93) overall, 0.93 (0.92, 0.95) for GA ≤39 completed weeks, and 0.83 (0.78, 0.88) for GA ≥40 completed weeks. Stillbirth rates among the reduced sample were also similar to those among the overall sample (Table S1), with the unadjusted HRs for stillbirth (2002–2004 versus 1992– 94) 0.93 (0.91, 0.94) for all GAs, 0.93 (0.91, 0.95) for GA ≤39 completed weeks, and 0.87 (0.81, 0.93) for GA ≥40 completed weeks. Parameter estimates and their 95% confidence intervals from the Cox regression models are presented in Table 1. Although the unadjusted hazard ratio (95% confidence interval) for total stillbirths for 2002–2004 versus 1992–94 was 0.93 (0.91, 0.94), after adjustment for maternal demographic and medical risk factors and fetal sex and parity, the overall risk of stillbirth did not decrease (HR = 1.00 [0.98, 1.02]). The unadjusted reduction in risk for GA ≤39 completed weeks (HR = 0.93 [0.91, 0.95]) disappeared after adjustment (HR = 1.01 [0.99, 1.03]) (Table 1). The reduced risk of stillbirth for GA ≥40 completed weeks was mitigated after adjustment but remained significant (unadjusted HR = 0.83 [0.78, 0.88]; adjusted HR = 0.92 [0.86, 0.99]).
The trend toward higher rates of labour induction and caesarean delivery from 1992–94 to 2002–2004 was observed in nearly all states. Among the 45 ecological units, only three states (Alaska, Hawaii and Wisconsin) had a decrease in induction rates ( 5.5, 0.4 and 4.6%, respectively), and all had an increase in caesarean delivery; most states (except Connecticut, Maine, Missouri, Nevada, New Jersey, New Mexico and Virginia) had a reduction in stillbirth. As shown in Table 2, no association was observed between changes in stillbirth and induction rates. The ecologic correlation between changes in stillbirth at GA ≤39 completed weeks and in induction was nearly zero (r = 0.02 [ 0.31, 0.28]), while a nonsignificant negative correlation (r = 0.17 [ 0.44, 0.13]) was observed between changes in stillbirth at GA ≥40 weeks and induction. A nonsignificant positive ecologic correlation (r = 0.23 [ 0.07, 0.49]) was observed between changes in stillbirth and caesarean delivery (Table 2). The ecologic correlations between changes in stillbirth and changes in caesarean were similar for stillbirth at GA ≤39 and ≥40 weeks.
Additional results Among low-risk non-Hispanic White births, stillbirth rates were lower than the overall rates in both periods: 4.0 per 1000 in 1992–94 and 2.5 per 1000 in 2002–2004 (adjusted HR = 0.64 [0.62, 0.66] for the later versus earlier period). Rates of caesarean delivery were also lower in both periods (19.3 and 23.4%, respectively) but rates of labour induction were similar to the overall rates. Ecologic analysis showed no significant association between changes in stillbirth and labour induction or caesarean delivery among low-risk women; all weighted ecologic correlation coefficients were
Table 1. Adjusted* hazard ratios and their 95% confidence intervals among US non–Hispanic White singleton births**
2002–2004 versus 1992–94 Maternal age
DOI: 10.1111/1471-0528.12883 www.bjog.org
Temporal trends in stillbirth in the United States, 1992–2004: a population-based cohort study X Zhang,a MS Kramera,b a Department of Pediatrics, b Department of Epidemiology, Biostatistics and Occupational Health, Faculty of Medicine, McGill University, Montreal, QC, Canada Correspondence: MS Kramer, 2300 Tupper Street (Les Tourelles), Montreal, QC, Canada H3H 1P3. Email [email protected]
Accepted 2 March 2014. Published Online 27 May 2014.
Objective To examine temporal trends in stillbirth and its risk
factors in the United States (US), and to assess the contribution of labour induction and caesarean delivery to the stillbirth rate. Design Population-based cohort study based on linked
birth-infant death and fetal death data files from the US National Vital Statistics System. Setting Complete data were available for 44 states and the District
of Columbia. Population or Sample Singleton births from 1992 to 2004. Methods We assessed changes in stillbirth rates from 1992–1994 to 2002–2004 before and after adjustment for changes in maternal characteristics including maternal age, education, smoking, and medical risk factors, using Cox regression models. We also carried out an ecological study, using states as the units of analysis, to assess the impact on the stillbirth rate of increasing induction and caesarean delivery. Race-specific subgroup analyses were performed and included non-Hispanic Whites and non-Hispanic Blacks.
Main outcome measure Stillbirth rate. Results The stillbirth rate among non-Hispanic White singleton births decreased 11.5% from 1992–1994 (5.2 per 1000) to 2002– 2004 (4.6 per 1000). After adjustment for maternal risk factors, the hazard ratio (HR) for 2002–2004 was 1.01 (0.99, 1.03) for gestational age (GA) ≤39 weeks, but 0.92 (0.86, 0.99) at 40 or more weeks. The ecologic analysis revealed a nonsignificant negative correlation of 0.17 ( 0.44, 0.13) between state-level changes in stillbirth at GA ≥40 weeks and labour induction. A nonsignificant positive correlation of 0.23 ( 0.07, 0.49) was observed between changes in stillbirth at all GAs and caesarean delivery and did not differ at GA ≤39 versus ≥40 weeks. Results were similar among non-Hispanic Blacks. Conclusions Changes in maternal risk factors explained the
reduction in stillbirth at GA ≤39 weeks but not at ≥40 weeks. The rise in labour induction and caesarean delivery rates did not explain the reduction in stillbirth ≥40 weeks of gestation. Keywords Caesarean delivery, labour induction, stillbirth.
Please cite this paper as: Zhang X, Kramer MS. Temporal trends in stillbirth in the United States, 1992–2004: a population-based cohort study. BJOG 2014; 121:1229–1236.
In the last two decades, the rate of preterm birth (41 weeks reduces the risks of stillbirth and perinatal mortality.18 We are aware of only two randomised trials of induction versus expectant management in late preterm or early term gestations: one in women with gestational hypertension, the other in women with gestational diabetes.19,20 Both trials were small, however, and neither observed any stillbirths. To our knowledge, no randomised trials at any gestational age have compared prelabour cesarean delivery with expectant management.
1229
Zhang, Kramer
In this study, we examine the temporal trends in stillbirth in the United States from 1992 to 2004, compare stillbirth rates between 1992–94 and 2002–2004, and assess the impacts on those trends of contemporaneous changes in maternal risk factors and in rates of labour induction and caesarean delivery.
Methods Data source and study population Our study is based on linked birth-infant death and fetal death data files from the National Vital Statistics System of the United States, which are available online from Centers for Disease Control and Prevention (CDC) (www.cdc.gov/ nchs/data_access/Vitalstatsonline.htm). The linked birth– infant death data provide demographic and health data for births occurring during a calendar year, based on information abstracted from birth certificates filed in vital statistics offices of each state and the District of Columbia.21 The fetal death files include information from all reports of fetal deaths.11 Most states report fetal deaths at 20 weeks or more of gestation and/or ≥350 g in birthweight,11 but a few states report fetal deaths for all periods of gestation.11 Demographic data include variables such as date of birth, age and educational attainment of the parents, marital status, live-birth order, race, sex, and geographic area. Health data include items such as birthweight, gestational age and prenatal care. Since 1989, a clinical estimate of gestational age has also been recorded,2 as has the use of labour induction.2,21 Data on caesarean delivery have also become available since the 1989 revision, but no information is recorded as to whether the caesarean was performed before or after the onset of labour.21 Until 2004, linked live birth and fetal death data files included geographic data, including state of birth occurrence; since 2005, however, those geographic data are no longer publicly available. Because a substantial number of states did not report the new (1989) data items from 1989 to 1991,10,22 our study is based on the singleton live birth and fetal death data from 1992 through 2004, the last year when the state identification data were publicly available. Six states were excluded from our analyses for all time periods, owing to missing data on Hispanic origin in fetal death data files in the 1992–94 period: Louisiana, Maryland, Massachusetts, New Hampshire, Oklahoma and Rhode Island. Moreover, Maryland and Oklahoma had missing LMP estimates of gestational age in >40% of fetal deaths and did not report a clinical estimate of gestational age; Louisiana did not report plurality on fetal death data files. Our overall study sample therefore comprised 6 650 475 births in the 1992–94 period and 6 114 413 births in the 2002–2004 period from 44 states and the District of Columbia, 91.8% of total singleton deliveries
1230
among non-Hispanic Whites in the two periods. The following ten states were also excluded for multivariate analysis, owing to missing data on maternal medical risk factors or smoking for >40% of fetal deaths: Alabama, California, Connecticut, District of Columbia, Hawaii, Indiana, New York, South Dakota, Texas and Virginia. This reduced sample for multivariate analysis comprised 4 536 115 births in the 1992–94 period and 4 275 460 births in the 2002– 2004 period, 64.5% of total singleton deliveries among non-Hispanic Whites in the two periods.
Gestational age estimation From the US birth certificate, gestational age (GA) is usually calculated from the first day of the mother’s last menstrual period (LMP). Gestational age derived from the LMP estimate is prone to error.23,24 A clinical estimate of gestation (CE) has also been available since the 1989 revision, except for the state of California. Although the CE is based on the managing clinician’s best estimate, including menstrual history, physical findings, laboratory values, and (if available) sonography,2,3 no instructions were provided prior to the 2003 revision of the US birth certificate for specifying the basis of the CE.2 Several methods for validating the LMP-based gestational age have been proposed to reduce misclassification, including that of Platt et al.,25 Zhang and Bowes,26 Alexander et al.27 and Qin et al.28 In Qin et al.’s28 ‘LMP/CE’ editing method, LMP is replaced by the CE when the two estimates differ by more than 2 weeks. The LMP/ CE method has been shown to reduce misclassification of GA28 and was the basis for gestational age estimation in this study. California did not report a clinical estimate before 2005, and thus LMP cannot be edited in that state.
Statistical analysis Our principal analysis was based on singleton births among non-Hispanic Whites. The effect of increased induction might be less effective in reducing stillbirth than it would be in Blacks or Hispanics because Whites already have high induction rates. The restriction by race was also intended to control for potential confounding due to differences in induction and caesarean rates by race6 and the changing racial composition over time. We compared stillbirth rates, as well as maternal demographic and medical risk factors, between two time periods 10 years apart: 2002–2004 versus 1992–94. As all surviving fetuses are at risk of subsequent stillbirth, our analytical approach was based on fetuses at risk rather than live births at each gestational week. In analyses based on individual women, Cox regression models were used to adjust for fetal sex, parity, and maternal demographic and medical risk factors including maternal age, education, smoking, diabetes, chronic and gestational hypertension. The association between stillbirth and labour induction or caesarean delivery cannot be ascertained in analyses
ª 2014 Royal College of Obstetricians and Gynaecologists
Temporal trends in stillbirth
based on individual women because it would be highly confounded by the medical indication for the procedure. Although maternal risk factors are reported on the US birth certificate, the specific clinical indications for induction or caesarean are not, and thus they cannot be controlled for adequately at the individual level. To reduce confounding by clinical indication, we carried out an ecological analysis10,29–31 to assess the impacts of labour induction and caesarean delivery on stillbirth, based on 45 ecologic units in our overall study sample: 44 states plus the District of Columbia. The ecologic study is based on changes in rates of stillbirth, labour induction, and caesarean delivery between 1992–94 and 2002–2004; ecologic correlations between the change in rates of stillbirth and of labour induction and caesarean delivery were calculated, with weighting for the number of births in each state. Although a few states have implemented the 2003 revision of birth certificates since 2003, coding for the 1989 revision was also available in those states, and the comparison between the two periods was therefore based on the 1989 coding. As a secondary analysis to further reduce confounding by the clinical indication for labour induction or caesarean delivery, we restricted the study sample to women with low-risk pregnancies, defined as maternal age 20–34 years and the absence of diabetes, chronic hypertension and pregnancy-induced hypertension.32–34 This classification of low risk has been used consistently in our previous studies7,35 and by other investigators,32–34 both in the USA and Canada. All other pregnancies were considered not to be at low risk. Ecologic analysis was repeated and stratified by low versus
nonlow risk. Finally, to assess the generalisability of our findings and examine effect modification by race, we repeated our analysis among non-Hispanic Blacks. All data were analysed using SAS version 9.2 (SAS Institute, Cary, NC, USA).
Results Maternal demographic and clinical characteristics As shown in Supporting Information Table S1 for the overall study sample, mothers were older among 2002–2004 births than among 1992–94 births; the proportion of mothers ≥35 years increased from 11.4 to 14.3%. Mothers had higher education in the later period, with the proportion of college degree or higher increasing from 24.7% in 1992–94 to 30.8% in 2002–2004, Maternal smoking was reduced from 15.2% in 1992–94 to 12.1% in 2002–2004, while the proportion with medical risk factors increased; the prevalence of diabetes, chronic hypertension, and gestational hypertension increased from 2.6, 0.6, and 3.3%, respectively, to 3.2, 0.9, and 4.1%, respectively. Labour induction increased by 61% between the two time periods, from 16.0% in 1992–94 to 25.7% in 2002–2004, while caesarean delivery increased by 23%, from 20.9% to 25.7%. Maternal characteristics and risk factors in the reduced sample for regression analysis were similar to those in the overall study sample in the two periods (Table S1).
Main findings Figure 1 shows the GA-specific stillbirth rates, based on fetuses at risk, for the two time periods. As shown in
Risk per 10,000 fetuses alive at each GA
15
NHW 1992-4 NHW 2002-4
10
5
0
20 21 22 23 24 25 26 27 28 29 30 31 32 33 34 35 36 37 38 39 40 41 42+
Gestational age in completed weeks Figure 1. GA-specific stillbirth rates based on fetuses at risk among US non-Hispanic White (NHW) singleton births. Six states (Louisiana, Maryland, Massachusetts, New Hampshire, Oklahoma and Rhode Island) were excluded, owing to missing data on gestational age and Hispanic origin in fetal death data files.
ª 2014 Royal College of Obstetricians and Gynaecologists
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Zhang, Kramer
Table S1, the stillbirth rate among non-Hispanic White singleton births decreased from 5.2 per 1000 total births in 1992–94 to 4.6 per 1000 in 2002–2004, a decrease of 11.5% between the two study periods. The stillbirth rate decreased modestly for GA ≤39 completed weeks over the two periods, from 4.7 per 1000 fetuses at risk in 1992–94 to 4.3 per 1000 fetuses at risk in 2002–2004 respectively, a relative 8.5% decrease; while for GA ≥40 completed weeks, the stillbirth rate decreased from 1.0 per 1000 in 1992–94 to 0.8 per 1000 in 2002–2004, a 20% relative reduction. Unadjusted hazard ratios (HR) for stillbirth (2002–2004 versus 1992–94) and their 95% confidence intervals (CI) were 0.92 (0.91, 0.93) overall, 0.93 (0.92, 0.95) for GA ≤39 completed weeks, and 0.83 (0.78, 0.88) for GA ≥40 completed weeks. Stillbirth rates among the reduced sample were also similar to those among the overall sample (Table S1), with the unadjusted HRs for stillbirth (2002–2004 versus 1992– 94) 0.93 (0.91, 0.94) for all GAs, 0.93 (0.91, 0.95) for GA ≤39 completed weeks, and 0.87 (0.81, 0.93) for GA ≥40 completed weeks. Parameter estimates and their 95% confidence intervals from the Cox regression models are presented in Table 1. Although the unadjusted hazard ratio (95% confidence interval) for total stillbirths for 2002–2004 versus 1992–94 was 0.93 (0.91, 0.94), after adjustment for maternal demographic and medical risk factors and fetal sex and parity, the overall risk of stillbirth did not decrease (HR = 1.00 [0.98, 1.02]). The unadjusted reduction in risk for GA ≤39 completed weeks (HR = 0.93 [0.91, 0.95]) disappeared after adjustment (HR = 1.01 [0.99, 1.03]) (Table 1). The reduced risk of stillbirth for GA ≥40 completed weeks was mitigated after adjustment but remained significant (unadjusted HR = 0.83 [0.78, 0.88]; adjusted HR = 0.92 [0.86, 0.99]).
The trend toward higher rates of labour induction and caesarean delivery from 1992–94 to 2002–2004 was observed in nearly all states. Among the 45 ecological units, only three states (Alaska, Hawaii and Wisconsin) had a decrease in induction rates ( 5.5, 0.4 and 4.6%, respectively), and all had an increase in caesarean delivery; most states (except Connecticut, Maine, Missouri, Nevada, New Jersey, New Mexico and Virginia) had a reduction in stillbirth. As shown in Table 2, no association was observed between changes in stillbirth and induction rates. The ecologic correlation between changes in stillbirth at GA ≤39 completed weeks and in induction was nearly zero (r = 0.02 [ 0.31, 0.28]), while a nonsignificant negative correlation (r = 0.17 [ 0.44, 0.13]) was observed between changes in stillbirth at GA ≥40 weeks and induction. A nonsignificant positive ecologic correlation (r = 0.23 [ 0.07, 0.49]) was observed between changes in stillbirth and caesarean delivery (Table 2). The ecologic correlations between changes in stillbirth and changes in caesarean were similar for stillbirth at GA ≤39 and ≥40 weeks.
Additional results Among low-risk non-Hispanic White births, stillbirth rates were lower than the overall rates in both periods: 4.0 per 1000 in 1992–94 and 2.5 per 1000 in 2002–2004 (adjusted HR = 0.64 [0.62, 0.66] for the later versus earlier period). Rates of caesarean delivery were also lower in both periods (19.3 and 23.4%, respectively) but rates of labour induction were similar to the overall rates. Ecologic analysis showed no significant association between changes in stillbirth and labour induction or caesarean delivery among low-risk women; all weighted ecologic correlation coefficients were
Table 1. Adjusted* hazard ratios and their 95% confidence intervals among US non–Hispanic White singleton births**
2002–2004 versus 1992–94 Maternal age
