Abstract
Despite considerable lay attention on the regulation and legislation of abortion in the United States, important gaps remain in our understanding of its incidence and health and social ...consequences since its legalization in 1973. Many of these gaps in knowledge can be attributed to a lack of access to high-quality, individual-level abortion data over the past 46 years. Herein, we review the strengths and limitations of different, currently available methods for enumerating abortions in the United States and discuss how lack of access to high-quality data limits our surveillance and research activities of not only abortion but other important reproductive and perinatal health outcomes. We conclude by discussing some potential opportunities for improved access to high-quality abortion data in the United States.
Prevalence statistics for pregnancy complications identified through screening such as gestational diabetes usually assume universal screening. However, rates of screening completion in pregnancy are ...not available in many birth registries or hospital databases. We validated screening-test completion by comparing public insurance laboratory and radiology billing records with medical records at three hospitals in British Columbia, Canada.
We abstracted a random sample of 140 delivery medical records (2014-2019), and successfully linked 127 to valid provincial insurance billings and maternal-newborn registry data. We compared billing records for gestational diabetes screening, any ultrasound before 14 weeks gestational age, and Group B streptococcus screening during each pregnancy to the gold standard of medical records by calculating sensitivity and specificity, positive predictive value, negative predictive value, and prevalence with 95% confidence intervals (CIs).
Gestational diabetes screening (screened vs. unscreened) in billing records had a high sensitivity (98% 95% CI = 93, 100) and specificity (>99% 95% CI = 86, 100). The use of specific glucose screening approaches (two-step vs. one-step) were also well characterized by billing data. Other tests showed high sensitivity (ultrasound 97% 95% CI = 92, 99; Group B streptococcus 96% 95% CI = 89, 99) but lower negative predictive values (ultrasound 64% 95% CI = 33, 99; Group B streptococcus 70% 95% CI = 40, 89). Lower negative predictive values were due to the high prevalence of these screening tests in our sample.
Laboratory and radiology insurance billing codes accurately identified those who completed routine antenatal screening tests with relatively low false-positive rates.
OBJECTIVE:To quantify the contribution of placenta accreta to the rate of postpartum hemorrhage and severe postpartum hemorrhage.
METHODS:All hospital deliveries in Canada (excluding Quebec) for the ...years 2009 and 2010 (N=570,637) were included in a retrospective cohort study using data from the Canadian Institute for Health Information. Placenta accreta included placental adhesion to the uterine wall, musculature, and surrounding organs (accreta, increta, or percreta). Severe postpartum hemorrhage included postpartum hemorrhage with blood transfusion, hysterectomy, or other procedures to control bleeding (including uterine suturing and ligation or embolization of pelvic arteries). Rates, rate ratios, population-attributable fractions (ie, incidence of postpartum hemorrhage attributable to placenta accreta), and 95% confidence intervals (CIs) were estimated. Logistic regression was used to quantify associations between placenta accreta and risk factors.
RESULTS:The incidence of placenta accreta was 14.4 (95% CI 13.4–15.4) per 10,000 deliveries (819 cases among 570,637 deliveries), whereas the incidence of placenta accreta with postpartum hemorrhage was 7.2 (95% CI 6.5–8.0) per 10,000 deliveries. Postpartum hemorrhage among women with placenta accreta was predominantly third-stage hemorrhage (41% of all cases). Although placenta accreta was strongly associated with postpartum hemorrhage (rate ratio 8.3, 95% CI 7.7–8.9), its low frequency resulted in a small population-attributable fraction (1.0%, 95% CI 0.93–1.16). However, the strong association between placenta accreta and postpartum hemorrhage with hysterectomy (rate ratio 286, 95% CI 226–361) resulted in a population-attributable fraction of 29.0% (95% CI 24.3–34.3).
CONCLUSION:Placenta accreta is too infrequent to account for the recent temporal increase in postpartum hemorrhage but contributes substantially to the proportion of postpartum hemorrhage with hysterectomy.
LEVEL OF EVIDENCE:II
Weight gain in early pregnancy may influence a woman’s risk of developing preeclampsia. However, the consequences of weight gain throughout pregnancy up to the diagnosis of preeclampsia are unknown. ...The aim of this study was to determine whether pregnancy weight gain before the diagnosis of preeclampsia is associated with increased risks of preeclampsia (overall and by preeclampsia subtype). The study population included nulliparous pregnant women in the Swedish counties of Gotland and Stockholm, 2008 to 2013, stratified by early pregnancy body mass index category. Electronic medical records were linked with population inpatient and outpatient records to establish date of preeclampsia diagnosis (classified as any, early preterm <34 weeks, late preterm 34–36 weeks, or term ≥37 weeks). Antenatal weight gain measurements were standardized into gestational age-specific z scores. Among 62 705 nulliparous women, 2770 (4.4%) developed preeclampsia. Odds of preeclampsia increased by ≈60% with every 1 z score increase in pregnancy weight gain among normal weight and overweight women and by 20% among obese women. High pregnancy weight gain was more strongly associated with term preeclampsia than early preterm preeclampsia (eg, 64% versus 43% increased odds per 1 z score difference in weight gain in normal weight women, and 30% versus 0% in obese women, respectively). By 25 weeks, the weight gain of women who subsequently developed preeclampsia was significantly higher than women who did not (eg, 0.43 kg in normal weight women). In conclusion, high pregnancy weight gain before diagnosis increases the risk of preeclampsia in nulliparous women and is more strongly associated with later-onset preeclampsia than early-onset preeclampsia.
Objective
To estimate the effect of antenatal corticosteroids on newborn respiratory morbidity in twins.
Design
Regression discontinuity applied to population‐based birth registry data.
Setting
...British Columbia, Canada, 2008–2018.
Population
Twin pregnancies admitted for birth between 31+0 and 36+6 weeks of gestation.
Methods
During our study period, Canadian clinical practice guidelines recommended antenatal corticosteroid administration for imminent preterm birth up to 33+6 weeks. We used a logistic model to compare the predicted risks of our outcomes among pregnancies admitted for birth immediately before this clinical cut‐point (higher probability of exposure to antenatal corticosteroids) versus immediately after it (lower probability).
Main outcome measures
Our primary outcome was a composite of newborn respiratory distress or in‐hospital death. Our secondary outcome was a composite of newborn respiratory intervention or in‐hospital death.
Results
Among 2524 pregnancies (5035 liveborn twins), 47% of admissions before 34+0 weeks of gestation were exposed to antenatal corticosteroids but only 4.2% of admissions after this cut‐point were exposed. The risk of newborn respiratory distress or in‐hospital mortality increased abruptly at 34+0 weeks, corresponding to a protective effect of treatment (risk ratio RR 0.69, 95% CI 0.53–0.90; risk difference RD −12 cases per 100 births, 95% CI −20 to −4.1). There was no clear evidence for or against an effect on newborn respiratory intervention or in‐hospital death (RR 0.89, 95% CI 0.70–1.13; RD −4.2 per 100, 95% CI −13 to +4.2).
Conclusions
Our findings provide evidence for the effectiveness of antenatal corticosteroids in preventing adverse newborn respiratory outcomes in twins.
Folic acid supplementation is recommended during pregnancy to support healthy fetal development; (6S)-5-methyltetrahydrofolic acid ((6S)-5-MTHF) is available in some commercial prenatal vitamins as ...an alternative to folic acid, but its effect on blood folate status during pregnancy is unknown. To address this, we randomised sixty pregnant individuals at 8–21 weeks’ gestation to 0·6 mg/d folic acid or (6S)-5-MTHF × 16 weeks. Fasting blood specimens were collected at baseline and after 16 weeks (endline). Erythrocyte and serum folate were quantified via microbiological assay (as globally recommended) and plasma unmetabolised folic acid (UMFA) via LC-MS/MS. Differences in biochemical folate markers between groups were explored using multivariable linear/quantile regression, adjusting for baseline concentrations, dietary folate intake and gestational weeks. At endline (n 54), the mean values and standard deviations (or median, inter-quartile range) of erythrocyte folate, serum folate and plasma UMFA (nmol/l) in those supplemented with (6S)-5-MTHF v. folic acid, respectively, were 1826 (sd 471) and 1998 (sd 421); 70 (sd 13) and 78 (sd 17); 0·5 (0·4, 0·8) and 1·3 (0·9, 2·1). In regression analyses, erythrocyte and serum folate did not differ by treatment group; however, concentrations of plasma UMFA in pregnancy were 0·6 nmol/l higher (95 % CI 0·2, 1·1) in those supplementing with folic acid as compared with (6S)-5-MTHF. In conclusion, supplementation with (6S)-5-MTHF may reduce plasma UMFA by ∼50 % as compared with supplementation with folic acid, the biological relevance of which is unclear. As folate is currently available for purchase in both forms, the impact of circulating maternal UMFA on perinatal outcomes needs to be determined.
Objective
Reference charts for classifying and monitoring pregnancy weight gain in severely obese women do not exist. The goal was to construct pregnancy weight‐gain‐for‐gestational‐age z‐score ...charts for overweight and obese mothers, stratified by severity of obesity.
Methods
Serial weight gain measurements were ed from 1047, 1202, 1267, and 730 overweight, class I, II, and III obese women, respectively, delivering uncomplicated term pregnancies at Magee‐Womens Hospital in Pittsburgh, PA. Multi‐level linear regression models were used to express serial weight gain measurements as a function of gestational age.
Results
There were a median interquartile range of 11 9‐12 and 11 9‐13 serial weight measurements for overweight and obese (class I, II, and III) women, respectively. The rate of weight gain was minimal until 15‐20 weeks and then increased in a slow, linear manner until term. The slope of weight gain flattened as pre‐pregnancy BMI increased. Charts were created describing the mean, standard deviation, and select percentiles of weight gain in class I, II, and III obese and overweight pregnancies.
Conclusions
These charts are an innovative tool for studying the association between gestational weight gain and adverse pregnancy outcomes.
The Institute of Medicine pregnancy weight gain guidelines were developed without evidence linking high weight gain to maternal cardiometabolic disease and child obesity. The upper limit of current ...recommendations may be too high for the health of the pregnant individual and child.
The aim of this study was to identify the range of pregnancy weight gain for pregnancies within a normal body mass index (BMI) range that balances the risks of high and low weight gain by simultaneously considering 10 different health conditions.
We used data from an United States prospective cohort study of nulliparae followed until 2 to 7 y postpartum (N = 2344 participants with a normal BMI). Pregnancy weight gain z-score was the main exposure. The outcome was a composite consisting of the occurrence of ≥1 of 10 adverse health conditions that were weighted for their seriousness. We used multivariable Poisson regression to relate weight gain z-scores with the weighted composite outcome.
The lowest risk of the composite outcome was at a pregnancy weight gain z-score of -0.6 SD (standard deviation) (equivalent to 13.1 kg at 40 wk). The weight gain ranges associated with no more than 5%, 10%, and 20% increase in risks were -1.0 to -0.2 SD (11.2-15.3 kg), -1.4 to 0 SD (9.4-16.4 kg), and -2.0 to 0.4 SD (7.0-18.9 kg). When we used a lower threshold to define postpartum weight increase in the composite outcome (>5 kg compared with >10 kg), the ranges were 1.6 to -0.7 SD (8.9-12.6 kg), -2.2 to -0.3 SD (6.3-14.7 kg), and ≤0.2 SD (≤17.6 kg). Compared with the ranges of the current weight gain guidelines (-0.9 to -0.1 SD, 11.5-16 kg), the lower limits from our data tended to be lower while upper limits were similar or lower.
If replicated, our results suggest that policy makers should revisit the recommended pregnancy weight gain range for individuals within a normal BMI range.
Rates of gestational diabetes are reported to be increasing in many jurisdictions, but the reasons for this are poorly understood. We sought to evaluate the relative contribution of screening ...practices for gestational diabetes (including completion and methods of screening) and population characteristics to risk of gestational diabetes in British Columbia, Canada, from 2005 to 2019.
We used a population-based cohort from a provincial registry of perinatal data, linked to laboratory billing records. We used data on screening completion, screening method (1-step 75-g glucose test or 2-step approach of 50-g glucose screening test, followed by a diagnostic test for patients who screen positive) and demographic risk factors. We modelled predicted annual risk for gestational diabetes, sequentially adjusted for screening completion, screening method and risk factors.
We included 551 457 pregnancies in the study cohort. The incidence of gestational diabetes more than doubled over the study period, from 7.2% in 2005 to 14.7% in 2019. Screening completion increased from 87.2% in 2005 to 95.5% in 2019. Use of 1-step screening methods increased from 0.0% in 2005 to 39.5% in 2019 among those who were screened. Unadjusted models estimated a 2.04 (95% confidence interval CI 1.94-2.13) increased risk of gestational diabetes in 2019 (v. 2005). This increase was 1.89 (95% CI 1.81-1.98) after accounting for the rise in screening completion and 1.34 (95% CI 1.28-1.40) after accounting for changes in screening methods. Further accounting for demographic risk factors (e.g., age, body mass index, prenatal care) had a small impact (increase of 1.25, 95% CI 1.19-1.31).
Most of the observed increase in the incidence of gestational diabetes was attributable to changes in screening practices (primarily changes in screening methods) rather than changing population factors. Our findings highlight the importance of understanding variation in screening practices when monitoring incidence rates for gestational diabetes.
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