Abstract
Background:
β-Blockers are a class of antihypertensive medications that are commonly used in pregnancy.
Objective:
To estimate the risks for major congenital malformations associated with first-trimester exposure to β-blockers.
Design:
Cohort study.
Setting:
Health registries in the 5 Nordic countries and the U.S. Medicaid database.
Patients:
Pregnant women with a diagnosis of hypertension and their offspring.
Measurements:
First-trimester exposure to β-blockers was assessed. Outcomes were any major congenital malformation, cardiac malformations, cleft lip or palate, and central nervous system (CNS) malformations. Propensity score stratification was used to control for potential confounders.
Results:
Of 3577 women with hypertensive pregnancies in the Nordic cohort and 14 900 in the U.S. cohort, 682 (19.1%) and 1668 (11.2%), respectively, were exposed to β-blockers in the first trimester. The pooled adjusted relative risk (RR) and risk difference per 1000 persons exposed (RD1000) associated with β-blockers were 1.07 (95% CI, 0.89 to 1.30) and 3.0 (CI, −6.6 to 12.6), respectively, for any major malformation; 1.12 (CI, 0.83 to 1.51) and 2.1 (CI, −4.3 to 8.4) for any cardiac malformation; and 1.97 (CI, 0.74 to 5.25) and 1.0 (CI, −0.9 to 3.0) for cleft lip or palate. For CNS malformations, the adjusted RR was 1.37 (CI, 0.58 to 3.25) and the RD1000 was 1.0 (CI, −2.0 to 4.0) (based on U.S. cohort data only).
Limitation:
Analysis was restricted to live births, exposure was based on dispensed medication, and cleft lip or palate and CNS malformations had few outcomes.
Conclusion:
The results suggest that maternal use of β-blockers in the first trimester is not associated with a large increase in the risk for overall malformations or cardiac malformations, independent of measured confounders.
Primary Funding Source:
The Eunice Kennedy Shriver National Institute of Child Health and Human Development and the Söderström König Foundation.
Chronic hypertension is increasingly prevalent in pregnancy (1), likely because of the higher prevalence of obesity in women of reproductive age and increasing maternal age at the time of pregnancy. Thus, antihypertensive medications have become a common exposure in early pregnancy (2–4). β-Blockers are one of the most frequently used classes of antihypertensive medication in pregnancy. An estimated 0.5% to 1% of pregnancies are exposed to β-blockers during the first trimester in the United States (2, 3), and the United Kingdom (4), and 0.1% in Sweden (5). β-Blockers (together with calcium-channel blockers and methyldopa) are recommended as a first-line therapy for the treatment of hypertension in pregnancy by professional organizations and societies (6, 7).
The association between β-blocker exposure in early pregnancy and risk for congenital malformations is controversial. β-Blockers cross the placenta (8), and at least some studies in animal models suggest a potential teratogenic effect (9). A recent systematic review and meta-analysis of studies involving pregnant women (10) did not find increased risk for congenital malformations overall but did report significantly increased risks for cardiac defects, cleft lip or palate, and neural tube defects. However, most included studies had significant limitations and were vulnerable to several sources of bias, including confounding by the mother’s underlying hypertension. Confounding by indication is particularly problematic in evaluating this association because studies suggest that maternal chronic hypertension increases risk for malformations (11–13).
Because antihypertensive medications are used frequently in early pregnancy, understanding whether β-blockers increase the risk for congenital malformations is a research priority. Thus, we examined the association between first-trimester exposure to β-blockers in utero and congenital malformations overall and organ-specific malformations that have previously been associated with β-blockers, including cardiac defects, cleft lip or palate, and central nervous system (CNS) malformations. To reduce potential confounding by indication, we restricted the analysis to pregnancies with a diagnosis of hypertension. To maximize study power, we used data from national health registries in the 5 Nordic countries (Denmark, Finland, Iceland, Norway, and Sweden) and the Medicaid database in the United States through the InPreSS (International Pregnancy Safety Study) consortium.
Methods
Study Overview
The InPreSS consortium is a collaborative among investigators from the 5 Nordic countries and the United States with access to high-quality data on medication exposure during pregnancy and infant outcomes (14). These data resources are analyzed with harmonized protocols to provide timely and accurate information on medication safety during pregnancy. The current study was based on 2 large population-based cohorts of births in the Nordic countries and the United States. The research was approved by applicable ethics review boards in Sweden, Iceland, Finland, and the United States (Appendix, available at Annals.org). No ethics approval was required in Denmark or Norway.
Data Sources and Study Population
The Nordic cohort included women from Denmark, Finland, Iceland, Norway, and Sweden who gave birth to a live singleton infant between 1996 and 2010 (N = 2 310 825 pregnancies) according to nationwide health registries (15, 16). These registries contain prospectively collected health information on all inhabitants, including data on health conditions, medications, and outcomes (such as congenital malformations). A civil personal registration number is assigned to each resident at birth or immigration, permitting data linkage at the individual level among national registries. Reporting to the registries is mandatory and regulated by national laws (17). According to data availability, each Nordic country contributed data from different periods (Denmark, 1997 to 2010; Finland, 1996 to 2006; Iceland, 2003 to 2007; Norway, 2005 to 2010; and Sweden, 2006 to 2010). The Appendix gives details about the Nordic data sources used in the analysis.
The US MAX cohort (N = 1 358 708 pregnancies) was identified from the 2000-to-2010 Medicaid Analytic eXtract (MAX), a nationwide Medicaid database. The method used to create this cohort has previously been described in detail (18). The cohort has been used extensively to study medication safety during pregnancy (19–21).
Eligibility Criteria
The Nordic cohort included all hypertensive women with singleton pregnancies resulting in a live-born infant. The US MAX cohort included hypertensive pregnant women aged 12 to 55 years who were continuously enrolled in Medicaid from 3 months before the date of their last menstrual period (LMP) to 1 month after delivery; women were linked to their live-born infants using a deterministic algorithm (18). Infants were also required to have Medicaid insurance for 3 months after their date of birth, unless they died, in which case a shorter eligibility period was allowed. Both cohorts excluded pregnancies in which a woman filled a prescription for a known teratogenic medication (Appendix Table 1, available at Annals.org) between 30 days before LMP and day 90 of pregnancy (first trimester) or in which a fetal chromosomal abnormality was detected in the infant in the first year of life in the Nordic cohort and before day 90 of life in the US MAX cohort (Figure 1).
Figure 1. Nordic and US MAX cohort selection.
ATC = Anatomical Therapeutic Chemical; LMP = last menstrual period; US MAX = U.S. Medicaid Analytic eXtract. * From 30 d before LMP through first trimester. These included wafarin, antineoplastic agents (actinomycin, busulfan, chlorambucil, colchicine, cyclophosphamide, doxorubicin, mercaptopurine, methotrexate, vinblastine, and vincristine), lithium, isotretinoin, misoprostol, and thalidomide for the US MAX cohort and retinoids, antineoplastic agents, and immunosuppressants (ATC codes D05BB, L01, and L04) for the Nordic cohort. † From 90 d before LMP through first trimester (ATC codes C02, C03, C04, C05, C08, and C09), or (for US MAX) a Beta-blocker in the 90 d before LMP but not in the first trimester. ‡ Excluded because of missing data on parity, i.e., the checkbox for parity on the maternity record had not been filled in.
Ascertainment of β-Blocker Exposure
We determined medication exposure through the nationwide prescription drug registries in the Nordic cohort and by review of pharmacy dispensing records in the US MAX cohort. Medications are recorded according to the World Health Organization’s Anatomical Therapeutic Chemical classification in the Nordic data and by generic drug name in the U.S. data.
In this study, a pregnancy was considered exposed if a woman filled a prescription for a β-blocker (Anatomical Therapeutic Chemical code C07) during the first trimester of pregnancy—that is, covering the period of embryogenesis. We excluded a pregnancy if an antihypertensive medication other than a β-blocker was dispensed between 90 days before LMP and the end of the first trimester (Figure 1) because some of these medications may also be associated with an increased risk for congenital malformations (22, 23). A pregnancy was considered unexposed if a woman did not fill a prescription for a β-blocker from the LMP to the end of the first trimester.
Congenital Malformations
We collected information on infants’ congenital malformations overall and organ-specific malformations that previous studies suggested were associated with β-blocker exposure, including cardiac defects, cleft lip or palate, and CNS malformations. In the Nordic cohort, we retrieved this information from the medical birth, patient, and cause-of-death registries. During the study period, congenital malformations were recorded according to the International Classification of Diseases, 10th Revision (ICD-10), in Denmark, Norway, Sweden, and Iceland and with ICD, Ninth Revision, Clinical Modification (ICD-9-CM) codes (Atlanta Clinical Modification for birth defects) in Finland (Appendix Table 2, available at Annals.org). A malformation was defined if an ICD-10 or ICD-9-CM diagnosis was recorded for the infant in any of the available registers within 1 year after birth. In the US MAX cohort, we used algorithms based on inpatient and outpatient ICD-9-CM diagnoses and procedure codes in the maternal or infant record to identify congenital malformations diagnosed within 90 days after birth (Appendix Table 2). Codes from the maternal record were considered because infant Medicaid claims are sometimes recorded under the mother’s name before the infant’s eligibility for Medicaid is activated. This approach has been previously validated to identify certain major congenital malformations in the MAX data with high positive predictive value (24).
Covariates
We considered a broad range of covariates in our analyses. To take advantage of the data available in each data set, we allowed the assessed covariates to differ between the Nordic and US MAX cohorts. In both, we defined maternal age at delivery, calendar year of birth, parity, diabetes, and diabetes treatments in early pregnancy, including insulin and oral hypoglycemic agents. In the Nordic data, we assessed country of origin. In the US MAX data, we defined maternal race/ethnicity, renal disease, overweight or obesity, smoking, alcohol abuse, illicit drug use, multiple gestations, and exposure to potential teratogens. We also defined health care use variables (which may be markers of overall comorbidity burden) in the US MAX data, including the number of physician visits for any reason and number of distinct non-antihypertensive prescription drugs during the 3 months before LMP.
Statistical Analysis
All analyses were run separately in the Nordic and US MAX cohorts, and the resulting effect estimates in each cohort were then combined using a meta-analytic approach. To compare pregnancies in which women were treated with β-blockers versus not treated with any antihypertensive medication during the first trimester, we calculated the crude prevalence of overall and organ-specific congenital malformations in the infant as well as unadjusted relative risks (RRs) (that is, prevalence ratio) with 95% CIs. We then further controlled for all available confounding variables using an approach based on propensity scores (PSs).
The PS was estimated using logistic regression and included all covariates shown in Table 1 and Appendix Table 3 (available at Annals.org) without further selection. No interaction terms were considered. After generating the PS, we excluded observations from the nonoverlapping regions of the PS distributions and created 50 PS strata of equal size based on the distribution among treated women. In the Nordic cohort, 1 patient from the exposed group and 25 from the unexposed group were excluded because of trimming. In the US MAX cohort, 3 from the exposed group and 18 from the unexposed group were excluded. The untreated pregnancies were then weighted using the distribution of the treated women among PS strata (25). We assessed the balance of baseline characteristics in the weighted population using the standardized mean difference (the balance of a covariate was deemed acceptable if the absolute value of the standardized mean difference was <10%). Adjusted RRs and risk differences were estimated using generalized linear models (PROC GENMOD in SAS [SAS Institute] with a weight statement). The model estimated the average treatment effect in exposed patients (that is, the increase in risk for malformations associated with β-blocker exposure compared with nonexposure in patients with the characteristics of those exposed to β-blockers).
Table 1.
Characteristics of Births With Hypertension, by β-Blocker Exposure in the First Trimester of Pregnancy*
| Characteristic | Nordic Cohort | US MAX Cohort | ||||||||
|---|---|---|---|---|---|---|---|---|---|---|
| Crude, n (%) | PS-Adjusted, n (%) | Standardized Difference, % | Crude, n (%) | PS-Adjusted, n (%) | Standardized Difference, % | |||||
| Exposed to β-Blockers (n = 682) | Unexposed to Antihypertensive Medications (n = 2895) | Exposed to β-Blockers (n = 681) | Unexposed to Antihypertensive Medications (n = 2870) | Exposed to β-Blockers (n = 1668) | Unexposed to Antihypertensive Medications (n = 13 232) | Exposed to β-Blockers (n = 1665) | Unexposed to Antihypertensive Medications (n = 13 214) | |||
| Maternal age | ||||||||||
| ≤19 y | 3 (0.4) | 16 (0.6) | 3 (0.4) | 11 (0.4) | 0.8 | 71 (4.3) | 1538 (11.6) | 71 (4.3) | 524 (4.0) | 1.5 |
| 20–24 y | 30 (4.4) | 205 (7.1) | 29 (4.3) | 123 (4.3) | −0.2 | 303 (18.2) | 4240 (32.0) | 303 (18.2) | 2254 (17.1) | 3.0 |
| 25–29 y | 129 (18.9) | 787 (27.2) | 129 (18.9) | 563 (19.6) | −1.7 | 529 (31.7) | 3764 (28.4) | 528 (31.7) | 4193 (31.7) | −0.1 |
| 30–34 y | 256 (37.5) | 1109 (38.3) | 256 (37.6) | 1053 (36.7) | 1.9 | 418 (25.1) | 2189 (16.5) | 417 (25.0) | 3408 (25.8) | −1.7 |
| 35–39 y | 203 (29.8) | 624 (21.6) | 203 (29.8) | 858 (29.9) | −0.2 | 257 (15.4) | 1151 (8.7) | 257 (15.4) | 2120 (16.0) | −1.7 |
| ≥40 y | 61 (8.9) | 154 (5.3) | 61 (9.0) | 261 (9.1) | −0.5 | 90 (5.4) | 350 (2.6) | 89 (5.3) | 714 (5.4) | −0.3 |
| Year of birth | ||||||||||
| 1997 | 20 (2.9) | 49 (1.7) | 20 (2.9) | 87 (3.0) | −0.6 | NA | NA | NA | NA | NA |
| 1998 | <20 (<2.9) | 62 (2.1) | 17 (2.5) | 63 (2.2) | 2.0 | NA | NA | NA | NA | NA |
| 1999 | 20 (2.9) | 95 (3.3) | 20 (2.9) | 76 (2.6) | 1.7 | NA | NA | NA | NA | NA |
| 2000 | 37 (5.4) | 104 (3.6) | 37 (5.4) | 148 (5.2) | 1.2 | <11 (<07) | 15 (0.1) | <11 (<0.7) | 18 (0.1) | −0.5 |
| 2001 | 30 (4.4) | 134 (4.6) | 30 (4.4) | 124 (4.3) | 0.4 | 55 (3.3) | 566 (4.3) | 55 (3.3) | 434 (3.3) | 0.1 |
| 2002 | 30 (4.4) | 133 (4.6) | 30 (4.4) | 138 (4.8) | −1.9 | 78 (4.7) | 696 (5.3) | 78 (4.7) | 597 (4.5) | 0.8 |
| 2003 | 22 (3.2) | 126 (4.4) | 22 (3.2) | 104 (3.6) | −2.2 | 107 (6.4) | 1034 (7.8) | 107 (6.4) | 843 (6.4) | 0.2 |
| 2004 | 36 (5.3) | 175 (6.0) | 36 (5.3) | 150 (5.2) | 0.3 | 134 (8.0) | 1345 (10.2) | 134 (8.0) | 1030 (7.8) | 0.9 |
| 2005 | 45 (6.6) | 182 (6.3) | 45 (6.6) | 191 (6.7) | −0.1 | 147 (8.8) | 1509 (11.4) | 147 (8.8) | 1182 (8.9) | −0.4 |
| 2006 | 76 (11.1) | 290 (10.0) | 76 (11.2) | 331 (11.5) | −1.1 | 193 (11.6) | 1563 (11.8) | 193 (11.6) | 1572 (11.9) | −1.0 |
| 2007 | 74 (10.9) | 361 (12.5) | 74 (10.9) | 304 (10.6) | 0.8 | 218 (13.1) | 1602 (12.1) | 217 (13.0) | 1755 (13.3) | −0.7 |
| 2008 | 98 (14.4) | 399 (13.8) | 98 (14.4) | 412 (14.4) | 0.1 | 222 (13.3) | 1605 (12.1) | 222 (13.3) | 1755 (13.3) | 0.2 |
| 2009 | 100 (14.7) | 462 (16.0) | 100 (14.7) | 429 (14.9) | −0.8 | 266 (15.9) | 1810 (13.7) | 265 (15.9) | 2074 (15.7) | 0.6 |
| 2010 | 76 (11.1) | 323 (11.2) | 76 (11.2) | 313 (10.9) | 0.9 | 246 (14.7) | 1487 (11.2) | 245 (14.7) | 1954 (14.8) | −0.2 |
| Multiparous | 461 (67.6) | 1891 (65.3) | 439 (64.5) | 1821 (63.4) | 1.0 | 1467 (87.9) | 10 782 (81.5) | 1464 (87.9) | 11 641 (88.1) | −0.5 |
| Mother diagnosed with diabetes | 37 (5.4) | 142 (4.9) | 36 (5.3) | 159 (5.5) | −0.2 | 166 (10.0) | 1372 (10.4) | 166 (10.0) | 1340 (10.1) | −0.6 |
| Maternal use of insulin | 29 (4.3) | 112 (3.9) | 28 (4.1) | 126 (4.4) | −0.3 | 84 (5.0) | 535 (4.0) | 84 (5.0) | 687 (5.2) | −0.7 |
| Maternal use of other antidiabetic agents | 12 (1.8) | 37 (1.3) | 12 (1.8) | 59 (2.1) | −0.3 | 76 (4.6) | 355 (2.7) | 76 (4.6) | 632 (4.8) | −1.0 |
NA = not available; PS = propensity score; US MAX = U.S. Medicaid Analytic eXtract.
Percentages may not sum to 100 due to rounding.
Finally, the effect estimates in the Nordic and US MAX data were combined for each outcome by fixed-effects meta-analysis.
Sensitivity Analyses
We did several sensitivity analyses to test the robustness of our findings. First, to reduce the possibility of exposure misclassification, we required women to have filled a prescription for a β-blocker at least twice during the first trimester. Second, to capture women with plausible exposure to β-blockers in the first trimester, we widened the window to include those who filled at least 1 prescription between 30 days before LMP and the end of the first trimester in the Nordic data and in the 90 days before LMP with days’ supply that overlapped the first trimester in the US MAX data. Third, to reduce potential dependency between multiple pregnancies, we restricted the analysis to pregnancies resulting in a woman’s first birth observed in the study data. The resulting effect estimates from the Nordic and US MAX data were combined by fixed-effects meta-analysis, as in the main analysis. Fourth, we changed the reference group of the main analysis to compare the prevalence of congenital malformations in infants exposed to β-blockers versus calcium-channel blockers in the first trimester of pregnancy and then calculated the unadjusted and PS-adjusted RRs with 95% CIs.
Finally, because the cohort was restricted to pregnancies that resulted in live birth, we sought to quantify the potential effect of missing pregnancies not resulting in live birth (that is, induced abortion, spontaneous abortion, or stillbirth) using methods previously described in detail (19, 20, 26) (Appendix).
Role of the Funding Source
The funders had no role in the design, conduct, or interpretation of the study.
Results
Our analysis involved 3577 women in the Nordic cohort and 14 900 in the US MAX cohort with hypertensive pregnancies resulting in live birth (Figure 1); 500 women in the Nordic data and 195 in the MAX data contributed more than 1 pregnancy. Among pregnancies in which women had a recorded hypertension diagnosis, 682 (19.1%) in the Nordic cohort and 1668 (11.2%) in the US MAX cohort were exposed to β-blockers. Appendix Table 4 (available at Annals.org) shows the specific β-blockers dispensed.
Women treated with β-blockers in the first trimester were older than untreated women, were more likely to have given birth previously, and were more likely to use medication for diabetes (Table 1 and Appendix Table 3). In the US MAX cohort, treated women were more likely to smoke, be white, have additional indications for β-blockers (including migraines, anxiety, cardiac arrhythmia, congestive heart failure, and ischemic heart disease), and have filled prescriptions for propylthiouracil. Appendix Figure 1 (available at Annals.org) shows the distribution of the PSs in the exposed and unexposed groups. After stratification on the PS, covariates were well balanced in exposed and unexposed women across the cohorts, with a standardized difference less than 10% for all variables (Table 1 and Appendix Table 3).
The fully adjusted results are as follows. The RRs in the Nordic and US MAX cohorts were 1.22 (95% CI, 0.88 to 1.71) and 1.01 (CI, 0.80 to 1.27), respectively, for malformations overall; 0.98 (CI, 0.52 to 1.84) and 1.16 (CI, 0.82 to 1.63) for cardiac defects; and 2.26 (CI, 0.47 to 10.8) and 1.81 (CI, 0.52 to 6.33) for cleft lip or palate (Table 2). The risk differences per 1000 persons exposed (RD1000) in the Nordic and US MAX cohorts were 13.0 (CI, −8.0 to 34.1) and 0.4 (CI, −10.4 to 11.2), respectively, for malformations overall; −0.5 (CI, −12.8 to 11.8) and 3.0 (CI, −4.4 to 10.5) for cardiac defects; and 2.4 (CI, −2.8 to 7.6) and 0.8 (CI, −1.3 to 2.9) for cleft lip or palate. In the US MAX cohort, the RR for CNS malformations was 1.37 (CI, 0.58 to 3.25) and the RD1000 was 1.0 (CI, −2.0 to 4.0); in the Nordic cohort, the low number of CNS malformation events precluded a corresponding analysis. Pooling these estimates for the Nordic and US MAX cohorts resulted in RRs of 1.07 (CI, 0.89 to 1.30) for any major malformation, 1.12 (CI, 0.83 to 1.51) for any cardiac malformation, and 1.97 (CI, 0.74 to 5.25) for cleft lip or palate (Figure 2). The pooled RD1000 associated with β-blockers was 3.0 (CI, −6.6 to 12.6) for any major malformation, 2.1 (CI, −4.3 to 8.4) for any cardiac malformation, and 1.0 (CI, −0.9 to 3.0) for cleft lip or palate.
Table 2.
Risk for Congenital Malformations After Exposure to β-Blockers in the First Trimester of Pregnancy
| Congenital Malformation | Prevalence, N (%) | RR (95% CI) | RD1000 (95% CI) | |||
|---|---|---|---|---|---|---|
| Exposed (n = 682) | Unexposed (n = 2895) | Unadjusted | PS-Adjusted | Unadjusted | PS-Adjusted | |
| Nordic cohort | ||||||
| Overall | 48 (7.0) | 152 (5.3) | 1.34 (0.98 to 1.83) | 1.22 (0.88 to 1.71) | 17.9 (−3.0 to 38.7) | 13.0 (−8.0 to 34.1) |
| Cardiac | 15 (2.2) | 55 (1.9) | 1.16 (0.66 to 2.04) | 0.98 (0.52 to 1.84) | 3.0 (−9.1 to 15.1) | −0.5 (−12.8 to 11.8) |
| CNS | * | * | * | * | * | * |
| Cleft lip/palate | 3 (0.4) | 4 (0.1) | 3.18 (0.71 to 14.19) | 2.26 (0.47 to 10.8) | 3.0 (−2.1 to 8.2) | 2.4 (−2.8 to 7.6) |
| Exposed (n = 1668) | Unexposed (n = 13 232) | |||||
| US MAX cohort† | ||||||
| Overall | 78 (4.7) | 534 (4.0) | 1.16 (0.92 to 1.46) | 1.01 (0.80 to 1.27) | 6.4 (−4.3 to 17.1) | 0.4 (−10.4 to 11.2) |
| Cardiac | 37 (2.2) | 224 (1.7) | 1.31 (0.93 to 1.85) | 1.16 (0.82 to 1.63) | 5.3 (−2.1 to 12.7) | 3.0 (−4.4 to 10.5) |
| CNS | <11 (<0.7) | 36 (0.3) | 1.32 (0.56 to 3.13) | 1.37 (0.58 to 3.25) | 0.9 (−2.1 to 3.9) | 1.0 (−2.0 to 4.0) |
| Cleft lip/palate | <11 (<0.7) | 13 (0.1) | 1.83 (0.52 to 6.42) | 1.81 (0.52 to 6.33) | 0.8 (−1.3 to 2.9) | 0.8 (−1.3 to 2.9) |
CNS = central nervous system; PS = propensity score; RD1000 = risk difference per 1000 persons exposed; RR = relative risk; US MAX = U.S. Medicaid Analytic eXtract.
Cannot be estimated because of restrictions on reporting observations with small cell sizes.
Cell sizes <11 cannot be presented because of restrictions on reporting small numbers in US MAX data.
Figure 2. Meta-analytic pooled estimates for the Nordic and US MAX cohorts.
Risk for congenital malformations after exposure to β-blockers in the first trimester of pregnancy, adjusted RR (95% CI), and adjusted RD1000 (95% CI). Cell sizes <11 cannot be presented for the US MAX data because of restrictions on reporting small numbers. RD1000 = risk difference per 1000 persons exposed; RR = relative risk; US MAX = U.S. Medicaid Analytic eXtract.
Relative risk estimates across the sensitivity analyses were consistent with those of the main analysis for malformations overall and cardiac malformations (Table 3). As in the main analysis, CIs were wide for cleft lip or palate in both cohorts. Changing the reference group from pregnancies unexposed to any antihypertensive agents to those exposed to calcium-channel blockers yielded imprecise estimates for specific malformations because of low numbers (Appendix Table 5, available at Annals.org). The adjusted point estimates for overall malformations comparing patients exposed to β-blockers versus calcium-channel blockers suggested similar risk for the Nordic cohort (RR, 1.01 [CI, 0.48 to 2.14]) and US MAX cohort (RR, 1.22 [CI, 0.79 to 1.90]), but the CIs were wide.
Table 3.
Sensitivity Analyses Examining the Risk for Congenital Malformations After Exposure to β-Blockers in the First Trimester of Pregnancy
| Congenital Malformation | Hypertension-Restricted Cohort | Meta-analysis of PS-Adjusted RR (95% CI) | |||||||||||
|---|---|---|---|---|---|---|---|---|---|---|---|---|---|
| Nordic Cohort | US MAX Cohort | ||||||||||||
| Total, n | Outcome Events, n | RR (95% CI) | Total, n | Outcome Events, n | RR (95% CI) | ||||||||
| Exposed | Unexposed | Exposed | Unexposed | Unadjusted | PS-Adjusted | Exposed | Unexposed | Exposed | Unexposed | Unadjusted | PS-Adjusted | ||
| Exposure based on ≥2 dispensings in the first trimester | |||||||||||||
| Overall | 283 | 2895 | 22 | 152 | 1.48 (0.96–2.28) | 1.35 (0.70–2.61) | 717 | 13 232 | 36 | 534 | 1.24 (0.90–1.73) | 1.10 (0.79–1.52) | 1.14 (0.86–1.53) |
| Cardiac | 283 | 2895 | 6 | 55 | 1.12 (0.49–2.57) | 0.97 (0.26–3.58) | 717 | 13 232 | 17 | 224 | 1.40 (0.86–2.28) | 1.23 (0.76–2.00) | 1.19 (0.76–1.88) |
| CNS | * | * | * | * | * | * | 717 | 13 232 | <11 | 36 | 1.54 (0.47–4.98) | 1.82 (0.56–5.93) | * |
| Cleft lip/palate | 283 | 2895 | <3 | * | 5.11 (0.94–27.81) | 2.40 (0.15–38.07) | 717 | 13 232 | <11 | 13 | 2.84 (0.64–12.56) | 2.46 (0.56–10.76) | 2.45 (0.67–8.99) |
| Exposure based on ≥1 dispensing 30 d before the LMP through the end of the first trimester (Nordic cohort) or the days’ supply overlapping the first trimester (US MAX cohort) | |||||||||||||
| Overall | 750 | 2827 | 51 | 149 | 1.29 (0.95–1.75) | 1.16 (0.82–1.66) | 1886 | 13 232 | 85 | 534 | 1.12 (0.89–1.40) | 0.97 (0.78–1.21) | 1.02 (0.84–1.23) |
| Cardiac | 750 | 2827 | 15 | 55 | 1.03 (0.58–1.81) | 0.98 (0.51–1.86) | 1886 | 13 232 | 38 | 224 | 1.19 (0.85–1.67) | 1.03 (0.74–1.45) | 1.02 (0.75–1.38) |
| CNS | * | * | * | * | * | * | 1886 | 13 232 | <11 | 36 | 1.36 (0.61–3.06) | 1.41 (0.63–3.18) | * |
| Cleft lip/palate | 750 | 2827 | 3 | 4 | 2.83 (0.63–12.61) | 1.29 (0.21–7.82) | 1886 | 13 232 | <11 | 13 | 1.62 (0.46–5.68) | 1.62 (0.46–5.67) | 1.50 (0.54–4.21) |
| Cohorts restricted to the mother’s first observed birth | |||||||||||||
| Overall | 608 | 2409 | 43 | 127 | 1.34 (0.96–1.87) | 1.49 (1.02–2.17) | 1618 | 13 072 | 78 | 523 | 1.20 (0.96–1.52) | 1.04 (0.83–1.31) | 1.15 (0.94–1.40) |
| Cardiac | 608 | 2409 | 13 | 46 | 1.12 (0.61–2.06) | 1.19 (0.61–2.33) | 1618 | 13 072 | 37 | 221 | 1.35 (0.96–1.91) | 1.19 (0.85–1.68) | 1.19 (0.88–1.62) |
| CNS | * | * | * | * | * | * | 1618 | 13 072 | <11 | 36 | 1.35 (0.57–3.19) | 1.40 (0.59–3.33) | * |
| Cleft lip/palate | 608 | 2409 | 3 | 3 | 3.96 (0.80–19.58) | 4.53 (0.59–34.85) | 1618 | 13 072 | <11 | 12 | 2.02 (0.57–7.15) | 2.15 (0.60–7.66) | 2.65 (0.90–7.79) |
CNS = central nervous system; LMP = last menstrual period; PS = propensity score; RR = relative risk; US MAX = U.S. Medicaid Analytic eXtract.
Cannot be estimated because of restrictions on reporting observations with small cell sizes.
Finally, we considered the potential effect of missing information on pregnancies that do not result in live births (Appendix Figure 2, available at Annals.org). Under the most extreme scenarios modeled, the pooled adjusted estimate of RR would shift from 1.07 to 1.26 for overall malformations, 1.12 to 1.32 for cardiac malformations, 1.97 to 2.32 for cleft lip or palate, and 1.37 to 2.40 for CNS malformations.
Discussion
In the pooled estimates from this cohort study of 3.6 million pregnancies drawn from the registries of the 5 Nordic countries and the MAX data set in the United States, the upper bound of the 95% CI suggests an excess risk per 1000 persons exposed to β-blockers in the first trimester of no more than 12.6 for congenital malformations overall, 8.4 for cardiac malformations, 3.0 for cleft lip or palate, and 4.0 for CNS malformations after adjustment for confounding factors, including maternal hypertension. The risk estimates were generally consistent in the Nordic and U.S. data and across several sensitivity analyses in which we varied the exposure definition.
Several prior studies yielded conflicting results about the association between β-blocker exposure in utero and major congenital malformations. A recent meta-analysis summarized the results of 13 case–control or cohort studies that evaluated this association (10). The pooled results did not show an increase in risk for major congenital malformations overall but showed a 2-fold increase in risk for cardiac malformations and an approximately 3-fold increase in risks for cleft lip or palate and neural tube defects. However, the authors noted that the included studies were subject to several potential biases, including recall, confounding, and publication biases. The potential for confounding by indication was of particular concern because several recent studies have suggested that maternal hypertension itself may be a risk factor for malformations (11, 13). Subsequent to the meta-analysis, a case–control study from the National Birth Defects Prevention Study reported a nearly 2-fold increase in risk for congenital heart defects associated with early β-blocker exposure in pregnancy (22). However, this association was present for other classes of antihypertensive drug and for untreated hypertension, suggesting that it may be explained (at least in part) by the underlying hypertension. These findings (22) may also be vulnerable to recall bias because exposure data were collected up to 2 years after delivery. Another retrospective, population-based cohort study using data from Kaiser Permanente Southern California showed a 2.5-fold increase in risk for cardiac defects associated with first-trimester exposure that shifted to the null with adjustment for measured confounders (27). A case–malformed control study from the EUROCAT (European Concerted Action on Congenital Anomalies and Twins) registries failed to find an association with congenital heart defects, oral clefts, or neural tube defects (28).
Our study overcomes many of the limitations associated with prior studies in this area. Conducted by the InPreSS consortium, our study pooled data from large cohorts drawn from 6 countries, representing some of the largest cohorts assembled for the study of drug safety in pregnancy. Medication use was defined on the basis of exposure during the first trimester, an etiologically relevant window when assessing a medication’s effect on organogenesis. Information on medication exposures was based on filled prescriptions and is thus free from recall bias. We restricted the analysis to women with hypertension, addressing the important issue of confounding by this common indication for β-blocker use. The data sources contained rich patient-level information, including maternal demographic characteristics, medical conditions, and medication exposures, and we used PS-based methods to adjust for such factors in the analysis.
Our observational study is also subject to certain limitations. β-Blocker exposure during the first trimester of pregnancy was based on filled prescriptions during this period, and we do not know whether the patient consumed the medication. To help overcome this limitation, we did a sensitivity analysis in which we redefined exposure as 2 or more dispensings of a β-blocker during the first trimester; the rationale was that if a woman refills a prescription, she likely consumed the medication. This sensitivity analysis yielded very similar point estimates to those of the main analysis for the assessment of overall and cardiac malformations. Another potential limitation of our study is that the covariates used for adjustment differed between the Nordic and US MAX data sets. However, we deliberately did the analysis in this manner to take advantage of all potentially relevant information available in each data set in an effort to generate the least biased effect estimates before pooling. Of note, effect estimates derived from each data set were similar. Our study considered exposure to β-blockers as a class; future studies should examine the risk associated with individual β-blockers and how it may vary with dose.
Our analysis was also restricted to pregnancies ending in live birth. We therefore explored the potential effect of missing information on pregnancies resulting in stillbirth or spontaneous or therapeutic abortion on our risk estimates in a series of sensitivity analyses. For the analyses of overall and cardiac malformations, changes to the risk estimates were relatively modest under the scenarios modeled. We also did a comparative safety analysis with calcium-channel blockers; however, risk estimates were generally imprecise because relatively few patients were exposed to calcium-channel blockers in the first trimester. Finally, despite large base cohorts, the risks for cleft lip or palate and neural tube defects are imprecisely estimated. Future, well-controlled studies done in similarly large cohorts will be needed to further refine these risk estimates.
β-Blockers are a first-line therapy for the treatment of hypertension in pregnancy (6) and are also commonly used in women of reproductive age who may inadvertently become pregnant (29). Our analysis was able to exclude a large increase in the relative and absolute risks for overall major malformations and for the most common class of birth defects, cardiac malformations. In the setting of small numbers of outcomes, our study cannot exclude an increase in the RR for the less common malformation types, cleft lip or palate and CNS malformations. However, the point estimates from our analysis suggest a more modest increase in the RR for these malformations than earlier publications have reported (10). The potential risks to the fetus must be balanced against the risks to the mother associated with untreated hypertension during pregnancy.
β-Blockers are among the most commonly used antihypertensive medications during pregnancy, but concerns have been raised about their potential teratogenic effects. These authors analyzed data from data from nationwide health registries in 5 Nordic countries and the U.S. Medicaid population to examine the risk for congenital malformations in offspring of women with hypertension who were exposed to β-blockers during the first trimester of pregnancy.
Acknowledgment:
The authors thank Loreen Straub for her assistance.
Grant Support: By grant R21HD092879 from the Eunice Kennedy Shriver National Institute of Child Health and Human Development and the Söderström König Foundation. Dr. Huybrechts was supported by career development grant K01MH099141 from the National Institute of Mental Health. Dr. Bateman was supported by career development grant K08HD075831 from the Eunice Kennedy Shriver National Institute of Child Health and Human Development. The Drugs and Pregnancy project was funded by the Finnish Medicines Agency, the National Institute for Health and Welfare, and the Social Insurance Institution of Finland.
Disclosures: Dr. Bateman reports grants from the National Institutes of Health during the conduct of the study and grants from Lilly, GlaxoSmithKline, Baxalta, Pacira, and Pfizer outside the submitted work. Dr. Hernandez-Diaz reports grants from the National Institutes of Health during the conduct of the study; grants from Lilly, Pfizer, and GlaxoSmithKline outside the submitted work; and being an advisor for the Antipsychotics Pregnancy Registry and epidemiologist for the North American Antiepileptics Pregnancy Registry, both at Massachusetts General Hospital. Dr. Kieler reports that her institution receives or received money from several drug companies for performing post-approval safety studies and comparative effectiveness studies; the companies, which her institution assists, are not marketing authorization holders of any of the studied drugs. Dr. Huybrechts reports grants from the National Institute of Mental Health and Eunice Kennedy Shriver National Institute of Child Health and Human Development during the conduct of the study and grants from Lilly, GlaxoSmithKline, Pfizer, Boehringer Ingelheim outside the submitted work. Authors not named here have disclosed no conflicts of interest. Disclosures can also be viewed at www.acponline.org/authors/icmje/ConflictOfInterestForms.do?msNum=M18-0338.
Appendix: Additional Information
Ethics Review
Research using the Nordic data was approved by the Regional Ethical Research Board in Stockholm, Sweden; the National Bioethics Committee and Data Protection Authority in Iceland; and the steering committee of the Drugs and Pregnancy project, the National Institute for Health and Welfare, and the Social Insurance Institute in Finland. No ethics approval was required in Denmark and Norway. Research using the U.S. data was approved by the Institutional Review Board of Brigham and Women’s Hospital, which granted a waiver of informed consent.
Details on the Nordic Databases Used in the Analyses
The Nordic cohort participants were identified through birth registries in Denmark, Iceland, Norway, and Sweden and through the Drugs and Pregnancy database in Finland. The Finnish database includes information from the Medical Birth Registry, the Malformation Registry, and the Registry on Induced Abortions combined with the Drug Reimbursement Registry. In the other Nordic countries, information on medications used during pregnancy was obtained directly from each country’s prescription drug registry and data on congenital malformations were obtained by combining information from the birth registries, patient registries, and cause-of-death registries.
Effect of Restriction to Live Births
The cohorts included in the study contained information on live births only and not on pregnancies that ended in stillbirth or spontaneous or therapeutic abortion. If live birth frequency is the same in pregnancies exposed and unexposed to β-blockers within the levels of the covariates included as potential confounders, then the RR estimates obtained from our analyses will be unbiased. However, if nonlive births occur more frequently in pregnancies exposed to β-blockers, estimates from our main analysis may be biased toward the null (that is, the true risk underestimated by the reported RR and RD). We therefore sought to quantify the potential effect of missing nonlive births using methods previously described in detail and used in our prior evaluations of the teratogenic potential of medications (19, 26, 30). In this sensitivity analysis, we assumed that the frequency of nonlive birth (which included stillbirth and spontaneous or therapeutic abortion) in nonmalformed fetuses is 20%. Then, on the basis of literature-based estimates of the frequency of termination with recognized malformations, we modeled live birth probabilities in unexposed, malformed fetuses. Probabilities ranged from 75% to 55% for overall malformations, cardiac malformations, and cleft lip or palate and from 55% to 35% for CNS malformations given the higher frequency of termination expected for fetuses diagnosed with CNS malformations (31). We then assessed the potential effect of a 10% to 20% higher frequency of nonlive birth among those exposed to β-blockers (again, within levels of the covariates adjusted for through the use of PSs) on the RRs estimated in the main pooled estimate for the fully adjusted analysis.
The results of these analyses are shown in Appendix Figure 2. For overall malformations, cardiac malformations, and cleft lip or palate, the most extreme scenario modeled was a probability of live birth of 55% (assuming a 20% frequency of termination, spontaneous abortion, or stillbirth for reasons other than malformations and a 25% frequency of nonlive birth because of a malformation) and a 20% absolute decrease in the probability of live birth in patients exposed to β-blockers compared with those unexposed (for both malformed and nonmalformed fetuses). Based on these assumptions, the pooled adjusted estimate of RR would shift from 1.07 to 1.26 for overall malformations, 1.12 to 1.32 for cardiac malformations, and 1.97 to 2.32 for cleft lip or palate. When considering CNS malformations, we modeled a higher frequency of terminations. Here, the most extreme scenario considered was a 35% probability of live birth in the unexposed group with malformations (20% frequency of termination, spontaneous abortion, or still birth for reasons other than malformations and 45% frequency of nonlive birth because of a malformation) and a 20% absolute decrease in the probability of live birth in patients exposed to β-blockers. In this situation, the RR from the pooled adjusted estimate shifts from 1.37 to 2.40.
Appendix Figure 1.
Histograms Showing the Distribution of the Propensity Score (PS) in the β-Blocker Exposed and Unexposed in the First Trimester of Pregnancy
Appendix Figure 2.
Impact of restriction to livebirths.
Appendix Table 1.
Teratogenic Medications
| Database | Drugs |
|---|---|
| Medicaid Analytic eXtract (MAX) | Warfarin, antineoplastic agents (actinomycin, busulfan, chlorambucil, colchicine, cyclophosphamide, doxorubicin, mercaptopurine, methotrexate, vinblastine, vincristine), lithium, isotretinoin, misoprostol, thalidomide |
| Nordic cohort | Retinoids, antineoplastic agents and immunosuppressants (ATC codes D05BB, L01, L04) |
ATC = Anatomical Therapeutic Chemical classification system.
Appendix Table 2.
Ascertainment of Congenital Malformations
| Malformation Group | US MAX Data (ICD-9)* | Nordic Data (ICD-9 and ICD-10) |
|---|---|---|
| Overall | ≥1 major malformation | ICD-9-CM codes: 740.xx-759.xx ICD-10 codes: Q00.0-Q89.9 (except Q17.0, Q38.1, Q53, Q65.0-Q65.6, Q82.5, Q86) |
| 1. Central nervous system defects | 740.xx-742.xx | ICD-9 codes: 740.xx-742.xx ICD-10 codes: Q00-Q07 |
| 2. Eye anomalies | 743.xx (excluded if only 743.6x and 743.8x) | |
| 3. Ear anomalies | 744.xx (excluded if only 744.1x, 744.21, 744.29, and 744.4x-744.9x) | |
| 4. Cardiac malformations | 745.xx-746.xx, 747.0x-747.4x, 747.83 (excluded if only 745.5 AND preterm, 746.02 AND preterm, 746.4x, 746.6x, 746.99, 747.0x AND preterm, 747.3 AND preterm) | ICD-9 codes: 745–747 ICD-10 codes: Q20.0-Q26.4, Q26.8, Q26.9 |
| 5. Vascular (non-cardiac) malformations | 747.6x-747.9x (excluded if only 747.83) | |
| 6. Respiratory malformations | 748.xx (did not count if only 748.1x) | |
| 7. Oral cleft | 749.xx | ICD-9 codes: 749.xx ICD-10 codes: Q35–37 |
| 8. Gastrointestinal malformations | 750.xx-751.xx (did not count if only 750.0x, 750.1x, 750.50, 751.0x) | |
| 9. Genital (male and female) malformations | 752.xx (did not count if only 752.42, 752.52) (in addition, did not count 752.5x if preterm) | |
| 10. Urinary malformations | 753.xx (did not include if only 753.7x) | |
| 11. Musculoskeletal malformations | 754.xx and 756.xx (did not count if only 754.3x, 754.81, 754.82, 756.2x) | |
| 12. Limb defects | 755.xx (excluded if only 755.65) | |
| 13. Other malformations | 757.xx; 759.xx (excluded if only 757.2–757.6, 759.81–759.83) |
ICD-9 = International Classification of Diseases, Ninth Revision; ICD-9-CM = International Classification of Diseases, Ninth Revision, Clinical Modification; ICD-10 = International Classification of Diseases, 10th Revision; US MAX = U.S. Medicaid Analytic eXtract; x = any possible digit.
In the US MAX cohort, a malformation was defined if any 1 of the following 3 criteria were met: an ICD-9-CM diagnosis was recorded for a specific malformation on >1 date, a diagnosis for a specific malformation was recorded on ≥1 dates and a relevant corrective surgery or procedure code was recorded, or a diagnosis for a specific malformation was recorded on ≥1 dates and the infant died within 30 d of birth. If the malformation was defined on the basis of codes from the maternal record alone (and not the infant record) and these codes were also present during the first 105 d of pregnancy (the time point after which anomalies may begin to be identified in-utero), the outcome was excluded because the diagnosis likely reflects a maternal congenital malformation.
Appendix Table 3.
Additional Characteristics of Births With Hypertension, by β-Blocker Exposure in the First Trimester of Pregnancy
| Characteristic | Nordic Cohort | US MAX Cohort | ||||||||
|---|---|---|---|---|---|---|---|---|---|---|
| Crude, n (%) | PS-Adjusted, n (%)* | Standardized Difference, % | Crude, n (%) | PS-Adjusted, n (%)* | Standardized Difference, % | |||||
| Exposed to β-Blockers | Unexposed to Antihypertensive Medications | Exposed to β-Blockers | Unexposed to Antihypertensive Medications | Exposed to β-Blocker s | Unexposed to Antihypertensive Medications | Exposed to β-Blockers | Unexposed to Antihypertensive Medications | |||
| Nordic country of birth | ||||||||||
| Denmark | 455 (66.7) | 2030 (70.1) | 455 (66.8) | 1910 (66.5) | 0.6 | * | * | * | * | – |
| Finland | <25 | 50 (1.7) | < 25 (3.4) | 111 (3.9) | −2.7 | * | * | * | * | – |
| Iceland | <3 | 13 (0.4) | <3 (0.1) | 1 (0.0) | 3.3 | * | * | * | * | – |
| Norway | 45 (6.6) | 300 (10.4) | 45 (6.6) | 183 (6.4) | 1.0 | * | * | * | * | – |
| Sweden | 157 (23.0) | 502 (17.3) | 157 (23.1) | 665 (23.2) | −0.3 | * | * | * | * | – |
| Maternal race/ethnicity | ||||||||||
| Caucasian | * | * | * | * | – | 744 (44.6) | 4657 (35.2) | 743 (44.6) | 5974 (45.2) | −1.2 |
| African American | * | * | * | * | – | 626 (37.5) | 5737 (43.4) | 624 (37.5) | 4896 (37.1) | 0.9 |
| Hispanic | * | * | * | * | – | 144 (8.6) | 1518 (11.5) | 144 (8.7) | 1113 (8.4) | 0.8 |
| Asian | * | * | * | * | – | 48 (2.9) | 318 (2.4) | 48 (2.9) | 384 (2.9) | −0.1 |
| Native American | * | * | * | * | – | 21 (1.3) | 215 (1.6) | 21 (1.3) | 159 (1.2) | 0.5 |
| Other | * | * | * | * | – | 54 (3.2) | 508 (3.8) | 54 (3.2) | 428 (3.2) | 0.0 |
| Unknown | * | * | * | * | – | 31 (1.9) | 279 (2.1) | 31 (1.9) | 259 (2.0) | −0.7 |
| Parity | ||||||||||
| 1 | 242 (35.5) | 1074 (37.1) | 242 (35.5) | 1049 (36.6) | −1.0 | * | * | * | * | – |
| 2 | 277 (40.6) | 1188 (41.0) | 276 (40.5) | 1087 (37.9) | 2.7 | * | * | * | * | – |
| 3 | 99 (14.5) | 429 (14.8) | 99 (14.5) | 447 (16.6) | −2.1 | * | * | * | * | – |
| ≥4 | 64 (9.4) | 204 (7.0) | 64 (9.4) | 257 (9.0) | 0.4 | * | * | * | * | – |
| Maternal medical/obstetric conditions | ||||||||||
| Renal disease | * | * | * | * | – | 35 (2.1) | 264 (2) | 35 (2.1) | 285 (2.2) | −0.4 |
| Overweight/obese | * | * | * | * | – | 143 (8.6) | 1074 (8.1) | 143 (8.6) | 1140 (8.6) | −0.1 |
| Alcohol use | * | * | * | * | – | 15 (0.9) | 116 (0.9) | 15 (0.9) | 117 (0.9) | 0.2 |
| Illicit drug use | * | * | * | * | – | 30 (1.8) | 294 (2.2) | 28 (1.7) | 228 (1.7) | −0.4 |
| Multiple gestation | * | * | * | * | – | 110 (6.6) | 689 (5.2) | 110 (6.6) | 854 (6.5) | 0.6 |
| Smoking | * | * | * | * | – | 108 (6.5) | 721 (5.5) | 105 (6.3) | 831 (6.3) | 0.1 |
| Potential indications for β-blockers | ||||||||||
| Migraines | * | * | * | * | – | 147 (8.8) | 622 (4.7) | 146 (8.8) | 1195 (9.0) | −1.0 |
| Anxiety | * | * | * | * | – | 152 (9.1) | 785 (5.9) | 151 (9.1) | 1231 (9.3) | −0.9 |
| Congestive heart failure | * | * | * | * | – | 18 (1.1) | 103 (0.8) | 18 (1.1) | 155 (1.2) | −0.9 |
| Cardiac arrhythmia | * | * | * | * | – | 104 (6.2) | 235 (1.8) | 103 (6.2) | 812 (6.2) | 0.2 |
| Ischemic heart disease | * | * | * | * | – | 27 (1.6) | 121 (0.9) | 27 (1.6) | 211 (1.6) | 0.2 |
| Potential teratogens | ||||||||||
| Danazol | * | * | * | * | – | 0 (0) | 0 (0) | 0 (0) | 0 (0) | |
| Progestins | * | * | * | * | – | 60 (3.6) | 356 (2.7) | 60 (3.6) | 468 (3.5) | 0.4 |
| Methimazole | * | * | * | * | – | 24 (1.4) | 14 (0.1) | 21 (1.3) | 113 (0.9) | 4.0 |
| Propylthiouracil | * | * | * | * | – | 51 (3.1) | 35 (0.3) | 48 (2.9) | 264 (2.0) | 5.7 |
| Corticosteroids | * | * | * | * | – | 239 (14.3) | 1360 (10.3) | 238 (14.3) | 1915 (14.5) | −0.6 |
| Fluconazole | * | * | * | * | – | 55 (3.3) | 373 (2.8) | 55 (3.3) | 425 (3.2) | 0.5 |
| Anticonvulsants | * | * | * | * | – | 15 (0.9) | 103 (0.8) | 15 (0.9) | 123 (0.9) | −0.3 |
| Health care use variables | ||||||||||
| Number of physician visits for any reason (by quartile) | ||||||||||
| 0 to 2 | * | * | * | * | – | 78 (4.7) | 946 (7.2) | 78 (4.7) | 622 (4.7) | −0.1 |
| 3 to 4 | * | * | * | * | – | 182 (10.9) | 1832 (13.9) | 181 (10.9) | 1433 (10.8) | 0.1 |
| 5 to 7 | * | * | * | * | – | 398 (23.9) | 3274 (24.7) | 398 (23.9) | 3177 (24.0) | −0.3 |
| ≥8 | * | * | * | * | – | 1010 (60.6) | 7180 (54.3) | 1008 (60.5) | 7983 (60.4) | 0.3 |
| Number of distinct nonantihypertensive prescription drugs (by quartile) | ||||||||||
| 0 | * | * | * | * | – | 72 (4.3) | 1521 (11.5) | 72 (4.3) | 572 (4.3) | −0.04 |
| 1 to 2 | * | * | * | * | – | 297 (17.8) | 2698 (20.4) | 297 (17.8) | 2377 (18.0) | −0.4 |
| 3 to 4 | * | * | * | * | – | 353 (21.2) | 2812 (21.3) | 353 (21.2) | 2838 (21.5) | −0.67 |
| ≥5 | * | * | * | * | – | 946 (56.7) | 6201 (46.9) | 943 (56.6) | 7426 (56.2) | 0.88 |
PS = propensity score; US MAX = U.S. Medicaid Analytic eXtract.
Not available.
Appendix Table 4.
Specific β-Blockers to Which Women Were Exposed During the First Trimester in the Nordic and US MAX Cohorts
| β-Blocker | US MAX Cohort, n | Nordic Cohort, n |
|---|---|---|
| Atenolol | 395 | 31 |
| Bisoprolol | 51 | ≤5 |
| Labetalol | 677 | 487 |
| Metoprolol | 338 | 92 |
| Pindolol | 11 | 46 |
| Propranolol | 167 | 13 |
US MAX = U.S. Medicaid Analytic eXtract.
Appendix Table 5.
Risk for Congenital Malformations After Exposure to β-Blockers Versus CCBs in the First Trimester of Pregnancy
| Congenital Malformation | Prevalence, (%) | RR (95% CI) | ||
|---|---|---|---|---|
| β-Blocker–Exposed (n = 682 | CCB-Exposed (n = 107) | Unadjusted | PS-Adjusted | |
| Nordic | ||||
| Overall | 48 (7.0) | 9 (8.4) | 0.84 (0.43–1.63) | 1.01 (0.48–2.14) |
| Cardiac | 15 (2.2) | 4 (3.7) | 0.59 (0.20–1.73) | 0.58 (0.17–1.85) |
| CNS | * | * | * | * |
| Cleft lip/palate | * | * | * | * |
| β-Blocker–Exposed (n = 1,668) | CCB-Exposed (n = 700) | |||
| US MAX | ||||
| Overall | 78 (4.7) | 33 (4.7) | 0.99 (0.67–1.48) | 1.22 (0.79–1.90) |
| Cardiac | 37 (2.2) | 16 (2.3) | 0.97 (0.54–1.73) | 1.02 (0.56–1.84) |
| CNS | <11 | <11 | 2.52 (0.3–20.88) | 6.06 (0.20–185.7) |
| Cleft lip/palate | <11 | <11 | 0.42 (0.08–2.07) | 0.40 (0.06–2.77) |
CCB = calcium-channel blocker; CNS = central nervous system; PS = propensity score; RR = relative risk; US MAX = U.S. Medicaid Analytic eXtract.
Cannot be estimated because of restrictions on reporting observations with small cell sizes. Cell sizes <11 cannot be presented because of restrictions on reporting small numbers in the US MAX data.
Footnotes
Publisher's Disclaimer: This is the prepublication, author-produced version of a manuscript accepted for publication in Annals of Internal Medicine. The American College of Physicians, the publisher of Annals of Internal Medicine, is not responsible for the content or presentation of the author-produced accepted version of the manuscript or any version that a third party derives from it. Readers who wish to access the definitive published version of this manuscript and any ancillary material related to this manuscript (e.g., correspondence, corrections, editorials, linked articles) should go to Annals.org or to the print issue in which the article appears. Those who cite this manuscript should cite the published version, as it is the official version of record.
Reproducible Research Statement: Study protocol and statistical code: Parts of the study protocol and statistical code can be made available to interested readers; contact Dr. Bateman (bbateman@bwh.harvard.edu). Data set: Not available because of data use agreements.
Contributor Information
Brian T. Bateman, Brigham and Women’s Hospital and Harvard Medical School, Boston, Massachusetts.
Uffe Heide-Jørgensen, Aarhus University Hospital, Aarhus, Denmark.
Kristjana Einarsdóttir, University of Iceland, Reykjavik, Iceland.
Anders Engeland, Norwegian Institute of Public Health, Oslo, and University of Bergen, Bergen, Norway.
Kari Furu, Norwegian Institute of Public Health, Oslo, Norway.
Mika Gissler, National Institute for Health and Welfare, Helsinki, Finland; University of Turku, Turku, Finland; and Karolinska Institutet, Stockholm, Sweden.
Sonia Hernandez-Diaz, Harvard T.H. Chan School of Public Health, Boston, Massachusetts.
Helle Kieler, Karolinska Institutet, Stockholm, Sweden.
Anna-Maria Lahesmaa-Korpinen, National Institute for Health and Welfare, Helsinki, Finland.
Helen Mogun, Brigham and Women’s Hospital, Boston, Massachusetts.
Mette Nørgaard, Aarhus University Hospital, Aarhus, Denmark.
Johan Reutfors, Karolinska Institutet, Stockholm, Sweden.
Randi Selmer, Norwegian Institute of Public Health, Oslo, Norway.
Krista F. Huybrechts, Brigham and Women’s Hospital and Harvard Medical School, Boston, Massachusetts.
Helga Zoega, University of Iceland, Reykjavik, Iceland, and University of New South Wales, Sydney, New South Wales, Australia.
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