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. Author manuscript; available in PMC: 2019 Dec 18.
Published in final edited form as: JAMA. 2018 Dec 18;320(23):2429–2437. doi: 10.1001/jama.2018.18307

Association of maternal first trimester ondansetron use with cardiac malformations and oral clefts in offspring

Krista F Huybrechts 1, Sonia Hernandez-Diaz 2, Loreen Straub 1, Kathryn J Gray 3, Yanmin Zhu 1, Elisabetta Patorno 1, Rishi J Desai 1, Helen Mogun 1, Brian T Bateman 1,4
PMCID: PMC6669077  NIHMSID: NIHMS1043547  PMID: 30561479

Abstract

Importance:

Evidence for the fetal safety of ondansetron, a 5-HT3 receptor antagonist that is commonly prescribed for nausea and vomiting in pregnancy, is limited and conflicting.

Objective:

To evaluate the association between ondansetron exposure during pregnancy and risk of congenital malformations.

Design, Setting, and Participants:

A retrospective cohort study nested in the 2000–2013 nationwide Medicaid Analytic eXtract. The cohort consisted of 1,816,414 pregnancies contributed by 1,502,895 women enrolled in Medicaid from 3 months before the last menstrual period through ≥1 month after delivery; infants were enrolled in Medicaid for ≥3 months after birth. The final date of follow-up was December 31, 2013. Propensity score stratification was used to control for treatment indication and other confounders.

Exposure:

Ondansetron dispensing during the first trimester, the period of organogenesis.

Main Outcome(s) and Measure(s):

Primary outcomes were cardiac malformations and oral clefts diagnosed during the first 90 days after delivery. Secondary outcomes included congenital malformations overall and subgroups of cardiac malformations and oral clefts.

Results:

Among 1,816,414 pregnancies (mean age, 24.3 [5.8] years), 88,467 (4.9%) were exposed to ondansetron during the first trimester. Overall, 14,577 of 1,727,947 unexposed and 835 of 88,467 exposed infants were diagnosed with a cardiac malformation, for an absolute risk of 84.4 (95% CI, 83.0 – 85.7) and 94.4 (88.0 – 100.8) per 10,000 births respectively. The absolute risks of oral clefts were 11.1 (10.6 – 11.6) and 14.0 (11.6 – 16.5) per 10,000 (1,912 unexposed and 124 exposed cases); and the risks of any congenital malformation were 370.4 (358.0 – 382.9) and 313.5 (310.9 – 316.1) per 10,000 (3,277 exposed and 54,174 unexposed cases). The adjusted relative risk (RR) for cardiac malformations was 0.99 (95% CI, 0.93 – 1.06) and the adjusted risk difference (RD) was −0.8 (−7.3 – 5.7 per 10,000 births). For oral clefts, the adjusted RR was 1.24 (1.03 – 1.48) and the RD was 2.7 (0.2 – 5.2 per 10,000 births). The adjusted estimate for congenital malformations overall was RR = 1.01 (0.98 – 1.05) and RD = 5.4 (−7.3 – 18.2 per 10,000 births).

Conclusions and Relevance:

Among offspring of mothers enrolled in Medicaid, first-trimester exposure to ondansetron was not associated with cardiac malformations or congenital malformations overall after accounting for measured confounders, but was associated with a small increased risk of oral clefts.

Introduction

Nausea and vomiting occur in up to 80% of pregnancies.1 About one-third of women with nausea and vomiting in pregnancy have symptoms that are clinically significant.1 A small percentage will develop hyperemesis gravidarum, which is the most common indication for hospital admission during the first part of pregnancy.1 Early treatment of nausea and vomiting in pregnancy is recommended to provide symptom relief and prevent progression to hyperemesis gravidarum.1 Although not formally approved for the treatment of nausea and vomiting in pregnancy, ondansetron, a 5-HT3 receptor antagonist, has rapidly become the most frequently prescribed drug for nausea and vomiting in pregnancy in the US because of its perceived superior antiemetic effects and improved side-effect profile compared with treatment alternatives.2–5 In 2014, an estimated 22% of pregnant women used ondansetron in the US.6

Nausea and vomiting in pregnancy typically occurs during the first trimester, the most sensitive time for exposure to teratogens because of organogenesis. The available evidence on the fetal safety of ondansetron is limited and conflicting (Table S1). While some studies suggested no increase in birth defects in women who took ondansetron early in pregnancy7–9, others suggested the possibility of a doubling in risk of cleft palate and cardiac malformations.10–12 Conflicting data leave pregnant women and clinicians unsure with respect to the appropriate risk-benefit trade-off for ondansetron use. The American College of Obstetricians and Gynecologists recently concluded that there are insufficient data on fetal safety with ondansetron exposure and further studies are warranted.13

The objective of this study was to investigate the association between exposure to ondansetron during the first trimester of pregnancy and risk of congenital malformations in offspring, focusing on cardiac malformations and oral clefts, in a national cohort of publicly-insured pregnant women.

Methods

The research was approved by the institutional review board of Brigham and Women’s Hospital, which granted a waiver of informed consent.

Data Source and Study Cohort

We used the pregnancy cohort nested in the nationwide Medicaid Analytic eXtract (MAX) for 2000–2013 (the most recent data available at the time of study conduct), the development of which has been described elsewhere14 and which has been used extensively to study the safety of medications in pregnancy.15–18 MAX data include demographic and insurance enrollment information, medical visits and hospitalizations, diagnoses and procedures received as an in- or outpatient, and prescriptions filled on an outpatient basis. Women 12–55 years were required to have Medicaid coverage from three months before the date of the last menstrual period to one month after delivery. Infants were required to have coverage through Medicaid for the first three months of life unless they died sooner. These restrictions did not affect the age or race distribution, but resulted in a decrease in the proportion of low-income women eligible through the occurrence of pregnancy and a corresponding increase in the proportion eligible independently of pregnancy.16 Pregnancies with exposure to a known teratogenic medication (i.e., warfarin, antineoplastic agents, lithium, isotretinoin, misoprostol, thalidomide) during the first trimester (N=3,562) and pregnancies with a chromosomal abnormality (N=3,156) were excluded.

Exposure

Women were considered exposed if they filled at least one ondansetron prescription during the first 3 months of pregnancy. Women who did not fill a prescription for ondansetron during the 3 months before the start of pregnancy until the end of the first trimester were considered unexposed. Women who filled an ondansetron prescription during the 3 months before the start of pregnancy were excluded to avoid contaminating the reference group with women who still have medication available for ingestion after the start of pregnancy. Women who filled a prescription during the first 90 days of pregnancy for (i) pyridoxine (with or without doxylamine), (ii) promethazine, (iii) metoclopramide, or (iv) any of these alternative pharmacological treatments, were used as alternative reference groups in additional analyses. Although these medications are available over-the-counter, Medicaid recipients frequently obtain prescriptions for these medications, allowing their use to be observed in our data and making this analysis possible.19

Outcomes

The presence of congenital malformations was defined using algorithms based on in- or outpatient diagnoses and procedure codes in the maternal (first month after delivery) or infant (first three months after date of birth) record which have been shown to identify congenital malformations with high specificity.20,21 The maternal record was considered because Medicaid claims are sometimes recorded under the mother before the infant’s eligibility has been processed.22 Motivated by findings from earlier studies suggesting a possible increased risk, the primary outcomes were defined as cardiac malformations and oral clefts. In secondary analyses, specific subgroups of cardiac malformations and oral clefts (i.e., palate, lip, lip and palate) were evaluated, along with congenital malformations overall. In exploratory analyses, the risks of other specific malformations were evaluated (Table S2 for the definition of all malformation outcomes).

Covariates

A broad range of potential confounders and proxies for confounders were considered, including treatment indication (nausea and vomiting in pregnancy, hyperemesis gravidarum) and associated conditions (weight loss, electrolyte and laboratory abnormalities, dehydration, gastroesophageal reflux), calendar year, state of residence, age, race, multiple gestation, maternal conditions, concomitant medication use, and general markers of the burden of illness. Race or ethnic group was considered to represent a potentially important confounding factor, and was categorized based on information submitted to the Centers for Medicare & Medicaid Services by individual states, which was based on information that had been collected and coded from Medicaid applications. Maternal morbidity and concomitant medication use were measured from 3 months before the start of pregnancy to the end of the first trimester. General markers of the burden of illness were assessed during the 3 months before but not during pregnancy to avoid these measures being affected by early awareness of possible pregnancy complications. Maternal conditions were assessed based on diagnostic codes and included psychiatric and neurological conditions (anxiety, depression, migraine or other headache) and other chronic conditions (diabetes, hypertension, renal disease, Crohn’s disease, irritable bowel syndrome, ulcerative colitis, overweight or obesity, underweight, illicit drug or alcohol abuse or dependence, smoking). Concomitant medications assessed included psychotropic medications (anticonvulsants, antidepressants, benzodiazepines, triptans), oral hypoglycemics, insulin, anti-hypertensives, progestins, corticosteroids, and suspected teratogens (fluconazole, methimazole, danazol, propylthiouracil, aminoglycosides, folic acid antagonists, potassium iodide, tetracycline). General markers of comorbid illness and disease severity included the Obstetric Comorbidity Index23, number of distinct prescriptions for medications other than anti-emetics, number of distinct diagnoses, number of outpatient visits, hospitalizations (yes/no, number, days), and emergency room visits.24

Analyses

Balance with respect to baseline characteristics between women treated with ondansetron and the untreated reference group was assessed using standardized differences. An absolute standardized difference greater than 0.1 was considered an indicator for substantial imbalances between the two exposure groups.25 Absolute risks for each of the outcomes and unadjusted relative risks and risk differences with their 95% confidence interval (CI) were calculated. Prevalence at birth was used as a proxy for absolute risks under the assumption that only a small proportion of fetuses with non-syndromic defects would die or be terminated in-utero. The actual risks may therefore be slightly larger. In the first adjusted analysis, we accounted for the possible confounding effect of the indication for treatment and its associated factors (nausea and vomiting in pregnancy, hyperemesis gravidarum and the use of other anti-emetics). In the second adjusted analysis, we accounted for all potential confounding variables described above (Table S3 for a comprehensive list of covariates included). Propensity scores, estimated using logistic regression, were used to efficiently adjust for a large number of covariates. After trimming observations from the non-overlapping regions of the propensity score distribution, we created 50 equally sized propensity score-strata based on the distribution among the treated women.26 In the outcome models, the untreated observations were weighted using the distribution of the treated among propensity score-strata. In confirmatory analyses, high-dimensional propensity score analyses that included 200 empirically defined covariates, in addition to the pre-specified covariates, were conducted to account for potential residual confounding.27 Adjusted relative risks and risk differences were estimated using a log-binomial model.

Sensitivity analyses were conducted to test the robustness of the primary results. The reference group was changed to women who filled a prescription for a different anti-emetic medication during the first trimester because they might be more comparable to ondansetron-exposed women than women who were never treated with anti-emetics during pregnancy. To evaluate the potential effect of exposure misclassification, exposure was redefined as having filled two or more prescriptions for ondansetron during the first trimester. Because the period of greatest sensitivity to teratogens for oral clefts is likely the later part of the first trimester, we redefined the exposure window as 6 to 12 weeks after the date of last menstrual period. In a negative control analysis, we assessed the risk of malformations in women who filled their first ondansetron prescription in gestational months 5 to 8, which is after the etiologically-relevant window. A null finding in this analysis provides indirect evidence of no substantial residual confounding. We also used this group with first ondansetron exposure in months 5 to 8 as an alternative reference group. Since the cohort included live births only, the potential effect of differences in the proportion of pregnancy losses among women treated with ondansetron versus those untreated within levels of covariates used in the adjustment were conducted, as previously described (Supplementary Appendix: Quantifying the potential effect of restriction to livebirths).15

All analyses were conducted using SAS 9.4 (SAS Institute, Cary, NC). Precision around the estimates of absolute risk and the measures of association are provided using two-sided 95% confidence intervals. No adjustments were made for multiple comparisons, and results of secondary analyses should be interpreted as exploratory.28–30

Results

Characteristics of the study cohort

The cohort consisted of 1,816,414 pregnancies, of which 88,467 (4.9%) were exposed to ondansetron during the first trimester. Use increased from 0.01% in 2000 to 12% in 2013 (Figure S1). Most women (1,251,216; 83.3%) contributed one pregnancy to the cohort, 203,037 (13.5%) contributed two pregnancies, and 48,642 (3.2%) contributed three or more pregnancies. Exposed women were more likely to be white, to have a diagnosis of psychiatric and neurological conditions, and to smoke. They were more likely to also fill a prescription for other medications used to treat nausea and vomiting in pregnancy (metoclopramide, promethazine, pyridoxine), for psychotropic medications, for corticosteroids, and for suspected teratogens. Most markers of comorbid illness and disease severity were also elevated among ondansetron-exposed compared to unexposed women (Table 1, Table S3).

Table 1.

Selected cohort characteristics of pregnancies with and without exposure to ondansetron during the first trimester

Unadjusted Accounting for Propensity Score Strata
Ondansetron Unexposed Ondansetron Unexposed
N(1) %(2) N(1) %(2) Standardized difference(3) N(1) %(2) N(1) %(2) Standardized difference(3)
No. of Pregnancies 88,467 1,727,947 88,446 1,727,546
No. of Women 84,819 1,418,076 84,802 1,417,839
Age, mean (SD) 25.0 5.4 24.3 5.9 0.12 25.0 5.4 24.9 5.5 0.01
Year of delivery
     2000–2005 10,435 11.8 655,408 37.9 −0.63 10,435 11.8 204,192 11.8 0.00
     2006–2010 35,272 39.9 752,283 43.5 −0.07 35,271 39.9 692,277 40.1 0.00
     2011–2013 42,760 48.3 320,256 18.5 0.67 42,740 48.3 831,078 48.1 0.00
Race(4)
     White 42,387 47.9 679,300 39.3 0.17 42,374 47.9 842,206 48.8 −0.02
     Black 27,714 31.3 567,916 32.9 −0.03 27,707 31.3 534,110 30.9 0.01
     Hispanic 7,979 9.0 257,895 14.9 −0.18 7,978 9.0 152,404 8.8 0.01
     Other or unknown 10,387 11.7 222,836 12.9 −0.04 10,387 11.7 198,826 11.5 0.01
Multiple gestation 2,465 2.8 31,623 1.8 0.06 2,463 2.8 48,298 2.8 0.00
Maternal conditions
     Anxiety 6,315 7.1 61,663 3.6 0.16 6,312 7.1 125,573 7.3 −0.01
     Depression 8,804 10.0 103,221 6.0 0.15 8,796 10.0 174,014 10.1 0.00
     Migraine/headache 11,338 12.8 119,261 6.9 0.20 11,332 12.8 229,286 13.3 −0.01
     Diabetes 1,552 1.8 27,507 1.6 0.01 1,550 1.8 30,758 1.8 0.00
     Hypertension 2,696 3.1 38,176 2.2 0.05 2,696 3.1 53,024 3.1 0.00
     Illicit drug abuse or dependence 578 0.7 19,383 1.1 −0.05 578 0.7 11,367 0.7 0.00
     Alcohol abuse or dependence 182 0.2 7,658 0.4 −0.04 182 0.2 3,654 0.2 0.00
     Smoking 5,707 6.5 57,432 3.3 0.15 5,705 6.5 112,629 6.5 0.00
     Obstetric comorbidity index, mean (SD) 1.0 1.4 0.8 1.4 0.08 1.0 1.4 1.0 1.4 0.00
Concomitant medications
     Antidepressants 13,954 15.8 148,884 8.6 0.22 13,946 15.8 280,753 16.3 −0.01
     Benzodiazepines 5,792 6.6 51,791 3.0 0.17 5,788 6.5 116,717 6.8 −0.01
     Oral hypoglycemics 859 1.0 14,882 0.9 0.01 859 1.0 17,164 1.0 0.00
     Insulin 711 0.8 14,673 0.9 −0.01 710 0.8 14,545 0.8 0.00
     Antihypertensives 3,372 3.8 47,085 2.7 0.06 3,371 3.8 67,933 3.9 −0.01
     Suspected teratogens(5) 7,521 8.5 90,213 5.2 0.13 7,514 8.5 151,128 8.8 −0.01
     Metoclopramide 12,947 14.6 52,818 3.1 0.42 12,928 14.6 261,384 15.1 −0.01
     Promethazine 23,628 26.7 139,932 8.1 0.51 23,607 26.7 489,930 28.4 −0.04
     Pyridoxine 2,389 2.7 9,036 0.5 0.17 2,375 2.7 46,722 2.7 0.00
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(1)

Number of pregnancies, unless otherwise indicated.

(2)

Proportion of pregnancies, unless otherwise indicated. Percentages may not total 100 because of rounding.

(3)

Please refer to Table S3 for details on the estimation of the standardized difference

(4)

Race or ethnic group was determined on the basis of information submitted to the Centers for Medicare & Medicaid Services by individual states, which was based on information that had been collected and coded from Medicaid applications.

(5)

Pregnancies with exposure to known teratogens were removed from the cohort. Suspected teratogens include: fluconazole, methimazole, danazol, propylthiouracil, aminoglycosides, folic acid antagonists, potassium iodide, tetracycline.

Risk of congenital malformations

The risk of cardiac malformations was 94.4 (95% CI, 88.0 – 100.8) per 10,000 among ondansetron-exposed pregnancies and 84.4 (83.0 – 85.7) per 10,000 among unexposed pregnancies (estimated based on 835 exposed and 14,577 unexposed cases), for an unadjusted relative risk (RR) of 1.12 (95% CI, 1.04 – 1.20) and an unadjusted risk difference (RD) of 10.0 (95% CI, 3.5 – 16.5) per 10,000 births. The risks of oral clefts were 14.0 (11.6 – 16.5) and 11.1 (10.6 – 11.6) per 10,000 exposed and unexposed pregnancies respectively (estimated based on 124 exposed and 1,912 unexposed cases), resulting in a RR of 1.26 (1.05 – 1.51) and a RD of 2.9 (0.4 – 5.4) per 10,000 births (Table 2, Figure 1).

Table 2.

Absolute risk of congenital malformations among infants born to mothers with and without exposure to anti-emetic medications

Exposure Group Total No. of Pregnancies PRIMARY OUTCOMES SECONDARY OUTCOME
Cardiac Malformations Oral Clefts Any Congenital Malformation
Events Risk 95% CI Events Risk 95% CI Events Risk 95% CI
no. of affected infants no./10,000 infants no. of affected infants no./10,000 infants no. of affected infants no./10,000 infants
No exposure 1,727,947 14,577 84.4 83.0 85.7 1,921 11.1 10.6 11.6 54,174 313.5 310.9 316.1
Ondansetron 88,467 835 94.4 88.0 100.8 124 14.0 11.6 16.5 3,277 370.4 358.0 382.9
Any other anti-emetic 185,876 1,629 87.6 83.4 91.9 219 11.8 10.2 13.3 6,141 330.4 322.3 338.5
Metoclopramide 52,818 476 90.1 82.1 98.2 63 11.9 9.0 14.9 1,732 327.9 312.7 343.1
Promethazine 139,932 1,213 86.7 81.8 91.5 165 11.8 10.0 13.6 4,634 331.2 321.8 340.5
Pyridoxine 9,036 77 85.2 66.3 104.2 <11 7.6 2.0 13.5 283 313.2 277.3 349.1
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Note: Cell sizes <11 are suppressed in accord with the CMS cell size suppression policy

Figure 1. Risk of congenital malformations in infants following exposure to ondansetron during the first trimester: Main analyses.

Figure 1

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Relative risks and 95% confidence intervals are presented to show the risk of cardiac malformations, oral clefts, and any congenital malformation among infants born to women with exposure to ondansetron during the first trimester, as compared with the risk among infants born to women without such exposure. Propensity score stratified level 1 refers to analyses adjusted for treatment indications and associated factors. Propensity score stratified level 2 refers to analyses adjusted for all potential confounding variables as listed in Table S3. High-dimensional propensity score stratified refers to analyses adjusted for 200 empirically identified covariates, in addition to the pre-specified covariates.

The increase in risk of cardiac malformations associated with ondansetron in the unadjusted analyses was related to ventricular septal defects (400 exposed and 6,826 unexposed cases; RR = 1.14, 95% CI 1.04 – 1.27; RD = 5.7, 1.2 – 10.2 per 10,000) and secundum atrial septal defects (216 exposed and 3,080 unexposed cases; RR = 1.37, 1.19 – 1.57; RD = 6.6, 3.3 – 9.9 per 10,000). The other specific cardiac malformations were less common, ranging between <11 and 113 exposed cases, and none of the unadjusted RR suggested an increased risk. The increased risk for oral clefts was attributable to cleft palate (65 exposed and 988 unexposed cases; RR = 1.29, 1.00 – 1.65; RD = 1.6, −0.2 – 3.5 per 10,000). There was no evidence of an increased risk for cleft lip (33 exposed and 620 unexposed cases; RR = 1.04, 0.73 – 1.48; RD = 0.1, −1.2 – 1.4 per 10,000) or cleft lip with cleft palate (48 exposed and 925 unexposed cases; RR = 1.01, 0.76 – 1.35; RD = 0.1, −1.5 – 1.6 per 10,000) in unadjusted analyses. The risk for any congenital malformation was 370.4 (95% CI, 358.0 – 382.9) per 10,000 ondansetron-exposed pregnancies compared with 313.5 (310.9 – 316.1) per 10,000 unexposed pregnancies (estimated based on 3,277 exposed and 54,174 unexposed cases), corresponding to a RR of 1.18 (1.14 – 1.22) and a RD of 56.9 (44.2 – 69.6) per 10,000 births (Table S4-S5).

After stratification on the propensity score, all measured patient characteristics were balanced between the ondansetron-exposed and unexposed groups judged by an absolute standardized difference <0.1 (Table 1, Table S3). While adjusting for the treatment indications and associated factors (propensity score level 1) did not substantially change the crude risk estimates, accounting for all pre-specified potential confounding variables (propensity score level 2) resulted in a null point estimate for cardiac malformations (RR = 0.99, 95% CI 0.93 – 1.06; RD = −0.8, 95% CI −7.3 – 5.7 per 10,000 births) and for the secondary outcome of congenital malformations overall (RR = 1.01, 0.98 – 1.05; RD = 5.4, −7.3 – 18.2 per 10,000 births), and a statistically significant RR of 1.24 (1.03 – 1.48) for oral clefts corresponding to a RD of 2.7 (0.2 – 5.2) per 10,000 births. These findings were confirmed in analyses that adjusted for proxies for unmeasured confounders through use of high-dimensional propensity scores (Figure 1).

Sensitivity and exploratory analyses

A total of 185,876 women filled a prescription for anti-emetics other than ondansetron during the first trimester: 52,818 for metoclopramide, 139,932 for promethazine, and 9,036 for pyridoxine (Table S6-S9). Results when using women exposed to other anti-emetics as the reference group were consistent with the main analyses (adjusted RR [95% CI] for cardiac malformations 1.01 [0.92–1.12], for oral clefts 1.32 [1.03–1.70], and for any congenital malformation 1.00 [0.95–1.05]; Figure 2). Results were similar across the individual anti-emetics (Figure 2, Table S10-S13).

Figure 2. Risk of congenital malformations in infants following exposure to ondansetron during the first trimester: Sensitivity analyses.

Figure 2

Figure 2

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Relative risks and 95% confidence intervals are presented to show the risk of cardiac malformations, oral clefts, and any congenital malformation among infants born to women with exposure to ondansetron during the first trimester. Results are presented for different sensitivity analyses: (i) changing the reference group to women exposed to any of the three alternative anti-emetics during the first trimester, (ii) changing the reference group to women exposed to metoclopramide during the first trimester, (iii) changing the reference group to women exposed to promethazine during the first trimester, (iv) changing the exposure group to women exposed to pyridoxine during the first trimester, (v) requiring ondansetron-exposed women to have filled at least two prescriptions during the first trimester, (vi) changing the exposure window to 6 to 12 weeks after the date of last menstrual period, (vii) using pregnancies with first exposure to ondansetron 5–8 months after the date of last menstrual period as the reference group, (viii) changing the exposure window to 5–8 months after the date of last menstrual period, which is after the etiologically-relevant window (negative control analysis). Panel A shows the unadjusted associations. Panel B shows the associations adjusted for all potential confounding variables (Propensity Score Level 2). For analyses (iv) and (vii), adjustment was done through 1:1 matching rather than fine stratification on the propensity score due to the relatively small size of the reference group compared to the exposed group.

The findings generally were not sensitive to changes in the exposure definition (≥ 2 prescriptions, 6–12 weeks). Requiring ≥ 2 prescriptions to be filled during the first trimester did not strengthen the association for oral clefts (adjusted RR = 1.15, 0.85 – 1.56). In the negative control analyses, which evaluated the association with exposure to ondansetron after the etiologically relevant time window, no increased risk was observed for oral clefts (RR = 0.98, 0.66 – 1.45). When using these women with exposure to ondansetron after the etiologically relevant window as an alternative reference group, adjusted estimates were consistent with the main analyses (Figure 2, Table S14-S17).

We quantified the potential for selection bias due to the restriction of the cohort to livebirths. The most extreme scenario considered was a probability of livebirth of 80% among unexposed infants without a malformation, 55% among unexposed infants with a cardiac or any type of malformation, 70% among unexposed infants with oral clefts, and a 20% absolute decrease in the probability of livebirth in the ondansetron-exposed compared to the unexposed for both malformed and non-malformed fetuses. Based on these assumptions, the RR estimate would shift from 0.99 to 1.17 for cardiac malformations, 1.24 to 1.30 for oral clefts, and 1.01 to 1.19 for any malformation (see Supplementary Appendix for details).

In exploratory analyses, we screened for associations with other specific malformations (Table S4). In the context of multiple comparisons, the risk of ear malformations was elevated: the absolute risks were 3.8 (2.7 – 5.4) per 10,000 exposed and 2.4 (2.2 – 2.7) per 10,000 unexposed pregnancies, corresponding to an adjusted RR of 1.64 (1.16 – 2.33) and a RD of 1.5 (0.2 – 2.8) per 10,000 births.

Discussion

In this national cohort of publicly-insured pregnant women, there was no association between exposure to ondansetron during the first trimester of pregnancy and increased risk of cardiac malformations or congenital malformations overall, after accounting for potentially confounding conditions. Given the large cohort size, the estimates were precise with an upper bound of the 95% CI of the adjusted relative risk of 1.06 for cardiac malformations and of 1.05 for congenital malformations overall. There was a small increase in the risk of oral clefts associated with ondansetron exposure with the upper bound of the 95% CI of the adjusted relative risk at 1.48, corresponding to 5 additional cases per 10,000 prenatally exposed livebirths. These findings were consistent across a broad range of sensitivity analyses.

This study expands on the evidence available to date. The largest and most recent cohort study published in the peer-reviewed literature included 19 cardiac malformations, 1 oral cleft, and 49 malformations overall among 1,349 exposed women (Table S1).31 Conclusions about the doubling of risk of cleft palate in a case-control study conducted by the Centers for Disease Control & Prevention using data from 1997–2004 were based on 11 exposed cases (odds ratio = 2.4, 95% CI 1.2 – 4.8).10 The risk was attenuated but remained increased in more recent analyses using the same data source for the years 2005–2014 based on 40 exposed cases (odds ratio = 1.6, 95% CI 1.1 – 2.4).12 However, in another case-control study using data from the Slone Epidemiology Center Birth Defects Study for 1997–2014 no increase in risk was observed based on 11 exposed cases (odds ratio = 0.5, 95% CI 0.3 – 1.0).12

In some prior studies suggesting an increase in cardiac malformations or cleft palate, exposure assessment was completely10,12 or partially31 based on women’s self-report, which may have resulted in bias since women who gave birth to an infant with a congenital malformation may be more likely to report all their exposures than women who gave birth to a healthy child (Table S1). Other studies did not account for important potential confounding variables, such as the underlying indication or its severity, maternal comorbidities (e.g., diabetes) and concomitant medication use.8,9,31 In addition, some of the previously reported associations have been found in the context of multiple comparisons10, increasing the likelihood of a chance finding. For this same reason, a cautious interpretation of the finding of a potential increase in the risk of ear deformities that emerged in exploratory analyses is warranted. It should be viewed as a potential safety signal that requires replication in other studies. While awaiting additional data, it is important to recognize that ear deformities are uncommon, with a base prevalence of 2 per 10,000 unexposed pregnancies in our cohort.

Apart from the large cohort size, this study has several strengths. First, information on medication exposures was collected in a prospective manner based on filled prescriptions for medications and is thus free from recall bias. Second, the MAX data contain rich patient-level information for confounding control, including maternal demographic characteristics, medical conditions, and medication exposures. Third, in addition to comparing exposed to unexposed women, ondansetron-exposed women were compared to women exposed to other anti-emetics and consistent results were documented. Aside from reducing the likelihood of confounding, this analysis also addressed the clinically relevant question of which treatment is preferred for women whose symptoms are not self-limiting and do not resolve with non-pharmacologic conservative measures. Fourth, both cardiac malformations and oral clefts have been shown to have a high positive predictive value when identified using computerized records. The positive predictive value for cardiac malformations was previously estimated at 78%20 and some well-known associations were replicated in the MAX dataset using this definition (i.e., maternal diabetes, anticonvulsant use, multifetal pregnancy)15 providing confidence in the specificity of the outcome definition. The positive predictive value for oral clefts has been estimated to be between 93% and 97%.21,32 Such modest non-differential misclassification would minimally bias the RR toward the null.

Limitations

This study has several limitations. First, filling a prescription does not mean that the medication was actually consumed which could bias the results towards the null. To address this limitation, sensitivity analyses in which women were required to have filled at least two prescriptions during the first trimester were conducted based on the notion that if a woman refills a prescription she is likely to have consumed the prescribed medication. This approach did not result in stronger associations. Second, in non-randomized studies, there is always concern about residual confounding due to unmeasured or poorly measured characteristics. For example, absence of a recorded diagnosis is equated with absence of the disease. It is possible, however, that the provider did not record the diagnosis. This may result in some misclassification of the confounder information, and hence affect our ability to control for confounding. While negative residual confounding is typically not a concern for null findings because drug exposure is not expected to be associated with factors protective against congenital malformations (i.e., cardiac malformations or congenital malformations overall), positive residual confounding could be a potential explanation for the increased risk in oral clefts. An attempt was made to refute this alternative explanation using different strategies including adjustment for proxies of unmeasured confounders through high-dimensional propensity scores, use of alternate reference groups, and a negative control analysis. No increased risk of oral clefts was observed using the negative control exposure window, supporting the validity of this association. Third, because the cohort was restricted to livebirths, severe congenital malformations that result in pregnancy losses or terminations will be missed. Nonsyndromic oral clefts do not result in fetal deaths and are rarely a reason for terminations.33 It is therefore highly unlikely that a differential proportion of non-live births among ondansetron users versus the reference groups would have biased our findings. For cardiac malformations and any congenital malformation, formal quantification of the potential for selection bias revealed that the differences in the proportion of non-live births among ondansetron users versus the reference groups within levels of covariates used in the adjustment would have to be unrealistically strong in order to dilute a clinically-meaningful increase in risk.

Fourth, this study was focused on individuals with Medicaid insurance. Medicaid covers the medical expenses for approximately 50% of all pregnancies in the US,34 making this an important population to study. The cohort inclusion criteria resulted in the selection of a more disadvantaged subpopulation within Medicaid, mostly composed of low-income adults, multiparae, and women with disabilities. The characteristics of this Medicaid-eligible population of pregnant women, i.e., young, racially diverse, and vulnerable population with a high burden of disabilities, are not expected to affect the biological relations studied. Therefore, the results should be generalizable to the broader Medicaid population, as well as commercially-insured pregnant women in the US, and pregnant populations in other countries. Fifth, the study was based on data from 2000–2013, but the biological association between ondansetron exposure and malformations should not change over time.

Conclusions

Among offspring of mothers enrolled in Medicaid, first-trimester exposure to ondansetron was not associated with cardiac malformations or congenital malformations overall after accounting for measured confounders, but was associated with a small increased risk of oral clefts.

Supplementary Material

Appendix

Key points.

Question:

Is the use of ondansetron during pregnancy associated with increased risk of cardiac malformations and oral clefts in offspring?

Findings:

In this cohort study including 1.8 million pregnancies, first trimester ondansetron use was associated with an increased risk of oral clefts (3 additional cases per 10,000 women treated; adjusted relative risk, 1.24) but not cardiac malformations.

Meaning:

Maternal use of ondansetron in the first trimester was associated with a small increased risk of oral clefts but no increased risk of cardiac malformations.

Acknowledgements:

Krista F. Huybrechts had full access to all the data in the study and takes responsibility for the integrity of the data and the accuracy of the data analysis.

Study Funding

This study was supported by an R03 grant (R03 HD091699) from the National Institute of Child Health & Human Development. Krista F. Huybrechts was supported by a career development grant K01MH099141 from the National Institute of Mental Health. Brian T. Bateman was supported by a career development grant K08HD075831 from the National Institute of Child Health & Human Development. Kathryn J. Gray was supported by the NIH K12 BIRCWH career development grant to Harvard Medical School. Elisabetta Patorno was supported by a career development grant K08AG055670 from the National Institute on Aging. The funders had no role in the design and conduct of the study; collection, management, analysis, and interpretation of the data; preparation, review, or approval of the manuscript; and decision to submit the manuscript for publication.

KFH, EP, RJD, and BTB report research funding to Brigham and Women’s Hospital from Eli Lilly, Pfizer, GlaxoSmithKline, Boehringer-Ingelheim, Merck, Bayer, Vertex, Pacira, and Baxalta outside the submitted work.

SHD reports research funding to Harvard T.H. Chan School of Public Health from Eli Lilly, Pfizer, and GlaxoSmithKline outside the submitted work.

SHD has consulted for Roche for unrelated topics and has worked with the North American antiepileptic drug pregnancy registry, which is funded by multiple companies. KJG has consulted for Quest Diagnostics, Inc. for unrelated topics. BTB has consulted for Aetion for unrelated topics, and served on an expert panel for a postpartum hemorrhage quality improvement project that was conducted by the Association of Women’s Health, Obstetric and Neonatal Nurses and funded by a grant from Merck for Mothers.

Footnotes

Conflict of Interest Disclosures:

LS, YZ, and HM report no conflicts of interest.

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Supplementary Materials

Appendix
NIHMS1043547-supplement-Appendix.docx (461.2KB, docx)

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