Objective
To assess the efficacy and safety of vaginal progesterone to prevent recurrent preterm birth and adverse perinatal outcomes in singleton gestations with a history of spontaneous preterm birth.
Data Sources
MEDLINE, Embase, LILACS, and CINAHL (from their inception to February 28, 2022), Cochrane databases, Google Scholar, bibliographies, and conference proceedings.
Study Eligibility Criteria
Randomized controlled trials that compared vaginal progesterone to placebo or no treatment in asymptomatic women with a singleton gestation and a history of spontaneous preterm birth.
Methods
The primary outcomes were preterm birth <37 and <34 weeks of gestation. The secondary outcomes included adverse maternal and perinatal outcomes. Pooled relative risks with 95% confidence intervals were calculated. We assessed the risk of bias in the included studies, heterogeneity ( I 2 test), small-study effects, publication bias, and quality of evidence; performed subgroup and sensitivity analyses; and calculated 95% prediction intervals and adjusted relative risks.
Results
Ten studies (2958 women) met the inclusion criteria: 7 with a sample size <150 (small studies) and 3 with a sample size >600 (large studies). Among the 7 small studies, 4 were at high risk of bias, 2 were at some concerns of bias, and only 1 was at low risk of bias. All the large studies were at low risk of bias. Vaginal progesterone significantly decreased the risk of preterm birth <37 weeks (relative risk, 0.64; 95% confidence interval, 0.50–0.81; I 2 =75%; 95% prediction interval, 0.31–1.32; very low-quality evidence) and <34 weeks (relative risk, 0.62; 95% confidence interval, 0.42–0.92; I 2 =66%; 95% prediction interval, 0.23–1.68; very low-quality evidence), and the risk of admission to the neonatal intensive care unit (relative risk, 0.53; 95% confidence interval, 0.33–0.85; I 2 =67%; 95% prediction interval, 0.16–1.79; low-quality evidence). There were no significant differences between the vaginal progesterone and the placebo or no treatment groups in other adverse perinatal and maternal outcomes. Subgroup analyses revealed that vaginal progesterone decreased the risk of preterm birth <37 weeks (relative risk, 0.43; 95% confidence interval, 0.33–0.55; I 2 =0%) and <34 weeks (relative risk, 0.27; 95% confidence interval, 0.15–0.49; I 2 =0%) in the small but not in the large studies (relative risk, 0.98; 95% confidence interval, 0.88–1.09; I 2 =0% for preterm birth <37 weeks; and relative risk, 0.94; 95% confidence interval, 0.78–1.13; I 2 =0% for preterm birth <34 weeks). Sensitivity analyses restricted to studies at low risk of bias indicated that vaginal progesterone did not reduce the risk of preterm birth <37 weeks (relative risk, 0.96; 95% confidence interval, 0.84–1.09) and <34 weeks (relative risk, 0.90; 95% confidence interval, 0.71–1.15). There was clear evidence of substantial small-study effects in the meta-analyses of preterm birth <37 and <34 weeks of gestation because of funnel plot asymmetry and the marked differences in the pooled relative risks obtained from fixed-effect and random-effects models. The adjustment for small-study effects resulted in a markedly reduced and nonsignificant effect of vaginal progesterone on preterm birth <37 weeks (relative risk, 0.86; 95% confidence interval, 0.68–1.10) and <34 weeks (relative risk, 0.92; 95% confidence interval, 0.60–1.42).
Conclusion
There is no convincing evidence supporting the use of vaginal progesterone to prevent recurrent preterm birth or to improve perinatal outcomes in singleton gestations with a history of spontaneous preterm birth.
Why was this study conducted?
Some professional organizations recommend offering vaginal progesterone to women with a singleton gestation and a history of spontaneous preterm birth despite conflicting evidence about its efficacy.
Key findings
Meta-analyses including data from 10 trials showed that vaginal progesterone reduced the risk of preterm birth <37 and <34 weeks of gestation and the risk of admission to the neonatal intensive care unit. However, subgroup analyses revealed that this intervention was beneficial only in small (N=7) but not in large (N=3) trials. Sensitivity analyses restricted to studies at low risk of bias and adjustment for small-study effects both resulted in a markedly reduced and nonsignificant effect of vaginal progesterone on preterm birth <37 and <34 weeks of gestation.
What does this add to what is known?
There is no convincing evidence supporting the use of vaginal progesterone to prevent recurrent preterm birth in singleton gestations with a history of spontaneous preterm birth.
Introduction
In the United States, the rate of preterm birth declined steadily from 2007 to 2014 but then rose continuously up to 2019. In 2020, the rate of preterm birth declined to 10.09%—the first decline in the rate since 2014. The most recent global estimates of preterm birth showed that 14.8 million live births (10.6% of all live births) were born preterm in 2014. In 2019, preterm birth complications were the leading cause of death among children younger than 5 years worldwide, accounting for 17.7% of all deaths and 36.1% of neonatal deaths. In addition, surviving preterm neonates are at an increased risk of short-term complications, long-term neurodevelopmental disabilities, chronic diseases in adulthood, and mortality in early to mid-adulthood.
Preterm labor is a syndrome associated with multiple mechanisms of disease, including infection and/or inflammation, decidual hemorrhage and vascular disease, uterine overdistention, cervical disease, disruption of maternal-fetal tolerance, decidual senescence, immunologically mediated processes, maternal stress, and decline in progesterone action, among others. The syndromic nature of preterm labor explains why a single method of intervention does not prevent all, or even predict most, cases of preterm birth.
A history of spontaneous preterm birth (following preterm labor, preterm prelabor rupture of membranes, or cervical insufficiency) is a major risk factor for recurrent preterm birth. Women with a history of spontaneous preterm birth have a 2.5- to 4-fold increased risk of spontaneous preterm birth in a subsequent pregnancy than women with no history. In 2017, a meta-analysis of 32 studies involving just over 55,000 women with at least 1 previous singleton spontaneous preterm birth reported that the pooled rate of recurrent spontaneous preterm birth was 30% (95% confidence interval [CI], 27%–34%). The risk of recurrent preterm birth increases as the gestational age of the previous preterm birth declines and as the number of previous preterm births increases. Recurrences often occur at the same gestational age as that of the previous preterm birth. , Women born spontaneously preterm have an increased risk of spontaneous preterm delivery in their own pregnancies. Genetic, environmental, and behavioral risk factors shared with 2 pregnancies could contribute to the increased risk of recurrent preterm birth. , ,
Since 2003, the American College of Obstetricians and Gynecologists (ACOG) has recommended the administration of 17α-hydroxyprogesterone caproate (17-OHPC) to patients with a singleton gestation and a history of spontaneous preterm birth, aiming to prevent preterm birth. In 2021, the ACOG updated its guidelines on the prediction and prevention of spontaneous preterm birth and recommended offering either vaginal progesterone or 17-OHPC. These guidelines were endorsed by the Society for Maternal-Fetal Medicine (SMFM). The evidence base for making this recommendation was an individual patient data (IPD) meta-analysis that included 31 trials of vaginal progesterone, 17-OHPC, and oral progesterone in asymptomatic women at an increased risk of preterm birth. This study reported that vaginal progesterone reduced the risk of preterm birth <34 weeks of gestation in high-risk singleton gestations but did not report results separately for the subgroup of women with a singleton gestation and a history of spontaneous preterm birth. Therefore, an assessment of the efficacy and safety of vaginal progesterone in such women is needed.
The objective of this systematic review and meta-analysis was to evaluate the efficacy and safety of vaginal progesterone to prevent recurrent preterm birth and adverse perinatal outcomes in asymptomatic women with a singleton gestation and a history of spontaneous preterm birth.
Materials and methods
This systematic review and meta-analysis followed a predefined protocol registered with the International Prospective Register of Systematic Reviews (PROSPERO; CRD42021275154) and was performed and reported according to the methods recommended in the Cochrane Handbook for Systematic Reviews of Intervention and the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines, respectively. Both the authors independently retrieved and reviewed studies for eligibility, assessed their risk of bias, and extracted data. Any discrepancies were resolved through discussion between the review authors.
Data sources and searches
Eligible trials were identified through searches in MEDLINE, EMBASE, LILACS, CINAHL, the Cochrane Central Register of Controlled Trials, and clinical trial registries (all from their inception to February 28, 2022) using the keywords progesterone and preterm birth to be as inclusive as possible. Google Scholar, proceedings of congresses and scientific meetings on obstetrics and maternal-fetal medicine, reference lists of identified studies, previously published systematic reviews, and review articles were also searched. We did not use any language restrictions, and we translated a non-English study..
Eligibility criteria
We included randomized controlled trials in which asymptomatic women with a singleton gestation and a history of at least 1 spontaneous preterm birth in any of their previous pregnancies were randomly allocated to receive either vaginal progesterone or placebo or no treatment for the prevention of preterm birth. Quasirandomized trials, trials that assessed vaginal progesterone in women with threatened or arrested preterm labor or second-trimester bleeding, and trials in which vaginal progesterone was administered in the first trimester to prevent miscarriage were excluded from the review. The results of subgroup analyses for women with a history of spontaneous preterm birth in randomized controlled trials whose primary aim was to prevent preterm birth in singleton gestations with a midtrimester sonographic short cervix were not included in the meta-analysis.
Outcome measures
The primary outcomes were preterm birth <37 and <34 weeks of gestation. The secondary outcomes included preterm birth <28 weeks of gestation, threatened preterm labor or need for tocolysis, use of antenatal corticosteroids, cesarean delivery, any maternal adverse event, discontinuation of treatment because of adverse events, preterm prelabor rupture of membranes, preeclampsia, gestational hypertension, gestational diabetes mellitus, respiratory distress syndrome (RDS), necrotizing enterocolitis (NEC), intraventricular hemorrhage (IVH), grade III or IV IVH, neonatal sepsis, retinopathy of prematurity, bronchopulmonary dysplasia, periventricular leukomalacia, fetal death, neonatal death, perinatal death, birthweight <1500 g and <2500 g, admission to the neonatal intensive care unit (NICU), use of mechanical ventilation, patent ductus arteriosus, and long-term neurodevelopmental and health outcomes.
Assessment of risk of bias
We assessed the risk of bias in each included study for the primary and secondary outcomes by using the Cochrane risk of bias tool 2 (RoB2), which considers the following domains: (1) bias arising from the randomization process; (2) bias due to deviations from intended interventions; (3) bias due to missing outcome data; (4) bias in measurement of the outcome; and (5) bias in the selection of the reported result. For each domain, the tool comprises a series of signaling questions aiming to elicit information about the features of the trial relevant to the risk of bias. Once the signaling questions had been answered, the next steps were to reach a risk-of-bias judgment and to assign 1 of 3 levels to each domain as follows: low risk of bias, some concerns of bias, or high risk of bias. Finally, an overall risk of bias judgment was reached for each study as follows: low risk of bias (the study is judged to be at low risk of bias for all domains), some concerns of bias (the study is judged to raise some concerns in at least 1 domain, but not to be at high risk of bias for any domain), and high risk of bias (the study is judged to be at high risk of bias in at least 1 domain or to have some concerns for multiple domains in a way that substantially lowers confidence in the result).
Data extraction
A data extraction form was used to collect information on the authors, title, publication date, language, duplicate publications, trial registration, funding sources, study characteristics (trial design, setting, follow-up period, attrition and exclusions from the analysis, and intention-to-treat analysis), participants (inclusion and exclusion criteria, number of women randomized, baseline characteristics, and the country and date of recruitment), interventions (gestational age at trial entry, daily dose of vaginal progesterone, duration, compliance, use of cointerventions, and characteristics of interventions used in the control group), and outcomes (prespecified primary and secondary outcomes, definition of outcomes, number of outcome events and/or mean±standard deviation for each outcome, and the total number of participants in each group). Relevant additional data of included studies that were provided to previous meta-analyses were included in our meta-analysis.
Statistical analysis
Statistical analysis was performed according to the recommendations of the Cochrane Handbook for Systematic Reviews of Interventions. The data were analyzed according to the intention-to-treat principle. We calculated the pooled relative risk (RR) for dichotomous outcomes with the associated 95% CI by using a random-effects model. This approach was chosen in anticipation of significant heterogeneity among the included studies. The number needed to treat (NNT) for an additional beneficial or harmful outcome with its 95% CI was calculated for the outcomes for which there was a statistically significant reduction or increase in the absolute risk difference. ,
Heterogeneity of the results among studies was firstly assessed with the visual inspection of forest plots for any lack of overlap of CIs. Then, we quantified statistical heterogeneity by using the I 2 statistical test. In the presence of statistical heterogeneity ( I 2 ≥30%), we investigated the potential causes through subgroup analyses. We also addressed heterogeneity by calculating the 95% prediction intervals for meta-analyses that contained at least 5 studies. The prediction interval shows the range of the true effect size in future studies similar to those in the meta-analysis.
If there were at least 10 studies included in a meta-analysis, we constructed funnel plots to investigate small-study effects and publication biases. Funnel plot asymmetry was assessed visually and with Egger’s and Harbord’s tests. A P value of <.10 indicated significant asymmetry of the funnel plot and evidence of small-study effects. In the presence of both heterogeneity and funnel plot asymmetry, we compared the fixed-effect and random-effects estimates of the intervention effect, as the random-effects model weights small studies higher. Different effect sizes strongly suggest small-study effects bias. If the results were inconsistent between the fixed-effect and random-effects models meta-analyses, we presented both results. In case of funnel plot asymmetry, a contour-enhanced funnel plot was constructed to differentiate asymmetry attributed to publication bias from that which owes to other factors. , , On a contour-enhanced funnel plot, contour lines separating areas of statistical significance from nonsignificance are superimposed on the funnel plot. If studies appear to be missing in areas of statistical nonsignificance of the plot, then it is possible that the asymmetry is due to publication bias. Conversely, if studies appear to be missing in areas of high statistical significance, this reduces the plausibility that publication bias is the underlying cause of funnel plot asymmetry. If small-study effects were suspected, we planned to use the iterative nonparametric Trim and Fill method for adjusting treatment effect estimates as a sensitivity analysis. The basic idea of this method is to add studies to the funnel plot until it becomes symmetrical.
We performed subgroup analyses according to the study sample size (<150 vs ≥150), study setting (low- and middle-income countries vs high-income countries vs both low- and middle- and high-income countries), study center status (single center vs multicenter), trial registration status (registered vs not registered), mean gestational age at treatment initiation (<24 weeks vs ≥24 weeks of gestation), and daily dose of vaginal progesterone (90–100 mg vs ≥200 mg). We assessed the subgroup differences by an interaction test in which a P value ≥.05 was considered to indicate that the effect of treatment did not differ significantly between the subgroups. To test the robustness of the meta-analyses, we carried out sensitivity analyses by including only the studies at overall low risk of bias. Subgroup and sensitivity analyses were restricted to the primary outcomes of preterm birth <37 and <34 weeks of gestation.
Assessment of quality of evidence
The quality (certainty) of the body of evidence for each individual outcome was assessed by using the 5 Grading of Recommendations Assessment, Development and Evaluation (GRADE) criteria (overall risk of bias, consistency of effect, imprecision, indirectness, and publication bias). The GRADE approach categorizes the certainty of the evidence into 4 levels as follows: (1) high: we are very confident that the true effect lies close to that of the estimate of the effect; (2) moderate: we are moderately confident in the effect estimate, and the true effect is likely to be close to the estimate of the effect. However, there is a possibility that it is substantially different; (3) low: our confidence in the effect estimate is limited, and the true effect may be substantially different from the estimate of the effect; and (4) very low: we have very little confidence in the effect estimate, and the true effect is likely to be substantially different from the estimate of effect. We estimated an overall GRADE quality rating by taking the lowest quality of evidence from all of the outcomes critical to decision-making. As all studies included in the review were randomized controlled trials, the starting level for all assessments was high certainty. We downgraded the level of certainty in the presence of a high risk of bias in included trials, indirectness of the evidence, unexplained heterogeneity or inconsistency, imprecision of results, and high probability of publication bias. We downgraded the evidence by 1 level if we considered the limitation to be serious and by 2 levels if we considered it to be very serious.
Statistical analyses were performed by using Review Manager (RevMan [Computer program], version 5.4.1; The Cochrane Collaboration, 2020), StatsDirect (version 3.3.5; StatsDirect Ltd, Wirral, United Kingdom), and Comprehensive Meta-Analysis (version 3; Biostat, Englewood, NJ). GRADEpro Guideline Development Tool (GRADEpro GDT [software], McMaster University and Evidence Prime, 2021) was used to assess the certainty of the evidence and to create the summary of findings table.
Results
Selection, characteristics, and risk of bias of studies
Our search strategy identified 14 studies for possible inclusion, of which we excluded 4 (2 quasirandomized trials, , 1 nonrandomized trial, and 1 trial that included women presenting with symptoms or signs of threatened preterm labor ) ( Supplemental Figure 1 ). Ten studies including 2958 women with a singleton gestation and a history of spontaneous preterm birth fulfilled the inclusion criteria. Nine studies were published in English , and 1 in Persian.
The main characteristics of the included trials are presented in Table 1 . Four studies included only women with a singleton gestation and a history of spontaneous preterm birth. , , , The remaining 6 studies comprised women with a singleton gestation and a history of spontaneous preterm birth and women with other risk factors for preterm birth , , , , or women with a twin gestation with a history of spontaneous preterm birth. We obtained data separately for singleton gestations with a history of spontaneous preterm birth from 4 of these studies. , Seven studies had a sample size <150 (“small studies”), , , and 3 had a sample size >600 (611 in the study by O’Brien et al, 912 in the study by Norman et al, and 775 in the study by Crowther et al ; “large studies”). All large studies were double-blind, placebo-controlled, registered (2 prospectively , and 1 retrospectively ), multicenter trials (2 conducted in high-income countries , and 1 conducted in both high-income [67% of recruited women] and low- and middle-income [33% of recruited women] countries ). All small studies were conducted in single centers located in low- and middle-income countries. Four small studies evaluated vaginal progesterone vs placebo, , and 3 evaluated vaginal progesterone vs no treatment. , , Only 2 small trials were registered, both retrospectively. , The remaining 5 small trials , , were not registered in a clinical trials registry.
| First author, year | Trial enrollment | Main inclusion and exclusion criteria | Interventions (number of women with a singleton gestation and a history of spontaneous preterm birth) | Compliance | Trial registration | Primary outcome |
|---|---|---|---|---|---|---|
| Da Fonseca, 2003 | Single center in Brazil |
|
| Unreported | Not registered | Preterm birth <37 wk |
| O’Brien, 2007 | 53 centers in United States, India, South Africa, Czech Republic, Chile, and El Salvador |
|
| 96.2% for vaginal progesterone group and 96.4% for placebo group | NCT00086177 | Preterm birth ≤32 wk |
| Majhi, 2009 | Single center in India |
|
| 100% for vaginal progesterone group | Not registered | Preterm birth <37 and <34 wk |
| Akbari, 2009 | Single center in Iran |
|
| Unreported | Not registered | Mean gestational age at delivery |
| Cetingoz, 2011 | Single center in Turkey |
|
| 100% for both study groups | Not registered | Preterm birth <37 wk |
| Modi, 2014 | Single center in India |
|
| 41.5% for vaginal progesterone group and 70.0% for placebo group | CTRI/2008/091/000218 | Preterm birth <37 wk |
| Azargoon, 2016 | Single center in Iran |
|
| Unreported | IRCT201012273386N2 | Preterm birth <37 wk |
| Norman, 2016 | 65 centers in United Kingdom and 1 in Sweden |
|
| 66.3% for vaginal progesterone group and 70.9% for placebo group | ISRCTN14568373 | Fetal death or delivery occurring before 34 wk of gestation; a composite of neonatal death, brain injury, or bronchopulmonary dysplasia; and the Bayley-III cognitive composite score at 2 y of age |
| Crowther, 2017 | 32 centers in Australia, 5 in New Zealand, and 2 in Canada |
|
| 91.6% for vaginal progesterone group and 90.8% for placebo group | ISRCTN20269066 | Respiratory distress syndrome and severity of respiratory disease |
| Abdou, 2018 | Single center in Egypt |
|
| Unreported | Not registered | Preterm birth <37 wk |
The daily dose of vaginal progesterone used in the trials was 90 mg in 1 study, 100 mg in 6 studies, , , 200 mg in 2 studies, , and 400 mg in 1 study. Most studies administered the treatment from 20–24 weeks to 34–36 weeks of gestation. Compliance >90% was reported in 4 studies. , , , In the studies by Modi et al and Norman et al, compliance in the vaginal progesterone group was 42% and 66%, respectively, whereas in the placebo group, it was 70% and 71%, respectively. Compliance was not reported in 4 trials. , , ,
Figure 1 shows the risk of bias in each included study. The 3 large studies , , were at overall low risk of bias for the primary outcomes and most secondary outcomes. Among the 7 small studies, 4 had an overall high risk of bias, , , , 2 had some concerns of bias, , and only 1 had an overall low risk of bias. One study was deemed to be at high risk of bias, given the deviations from the intended interventions: 13 women with preterm birth because of preterm prelabor rupture of membranes or medically indicated delivery (8 in the vaginal progesterone group and 5 in the placebo group) were inappropriately excluded from analyses after randomization. The inclusion of these women in the analyses (intention-to-treat effect) would become the effect of vaginal progesterone on preterm birth <37 weeks of gestation into a non-statistically significant result. The studies by Akbari et al, Modi et al, and Abdou et al were judged to have an overall high risk of bias because they had some concerns of bias in multiple domains. The studies by Akbari et al and Abdou et al did not provide information on the methods used to generate the random allocation sequence and to conceal allocation, on the blinding of outcome assessors to intervention status, and on the selection of the reported result. In addition, the study by Akbari et al provided insufficient information on deviations from intended interventions. The study by Modi et al did not report information on the number of women with missing outcome data and on the selection of the reported result. The study by Majhi et al was judged as having some concerns of bias because participants and personnel were aware of intervention, and there was no information on blinding of outcome assessors to the intervention status. The study by Azargoon et al was judged to have some concerns of bias arising from the randomization process because the allocation concealment method was not reported.
Primary outcomes
Vaginal progesterone was associated with a significant decrease in the risk of preterm birth <37 weeks of gestation (32.0% vs 37.8%; pooled RR from random-effects model, 0.64; 95% CI, 0.50–0.81; P =.0003; pooled RR from fixed-effect model, 0.85; 95% CI, 0.77–0.93; P =.0008; I 2 =75%; 95% prediction interval of the RR, 0.31–1.32; NNT, 17; 95% CI, 11–42) ( Figure 2 ) and preterm birth <34 weeks of gestation (13.5% vs 17.0%; pooled RR from random-effects model, 0.62; 95% CI, 0.42–0.92; P =.02; pooled RR from fixed-effect model, 0.79; 95% CI, 0.67–0.94; P =.008; I 2 =66%; 95% prediction interval of the RR, 0.23–1.68; NNT, 28; 95% CI, 16–109) ( Figure 3 ).
Secondary outcomes
Table 2 shows the effect of vaginal progesterone on pregnancy, maternal, and perinatal outcomes. There was no significant difference between the vaginal progesterone and placebo or no treatment groups in the frequency of preterm birth <28 weeks of gestation (4.3% vs 4.0%; pooled RR, 1.12; 95% CI, 0.70–1.78; I 2 =18%). The frequencies of other pregnancy and maternal outcomes did not significantly differ between the study groups. Infants whose mothers received vaginal progesterone had a significantly lower risk of NICU admission (14.4% vs 20.7%; pooled RR from random-effects model, 0.53; 95% CI, 0.33–0.85; P =.008; pooled RR from fixed-effect model, 0.70; 95% CI, 0.57–0.85; P =.0005; I 2 =67%; 95% prediction interval of the RR, 0.16–1.79; NNT, 16; 95% CI, 10–36). There was no evidence of an effect of vaginal progesterone on RDS, NEC, IVH, grade III or IV IVH, neonatal sepsis, retinopathy of prematurity, bronchopulmonary dysplasia, periventricular leukomalacia, fetal death, neonatal death, perinatal death, birthweight <1500 g and <2500 g, use of mechanical ventilation, and patent ductus arteriosus.
| Outcome | Number of trials | Vaginal progesterone | Placebo or no treatment | Relative risk (95% CI) | P value | I 2 , % | 95% prediction interval for the RR | ||
|---|---|---|---|---|---|---|---|---|---|
| Preterm birth <28 wk | 9 , | 62/1429 | (4.3%) | 56/1406 | (4.0%) | 1.12 (0.70–1.78) | 0.64 | 18 | 0.51–2.49 |
| Threatened preterm labor or need for tocolysis | 5 , , | 176/866 | (20.3%) | 196/845 | (23.2%) | 0.82 (0.63–1.06) | 0.12 | 37 | 0.48–1.42 |
| Use of antenatal corticosteroids | 2 , | 211/699 | (30.2%) | 213/687 | (31.0) | 0.98 (0.83–1.14) | 0.76 | 0 | NA |
| Cesarean delivery | 3 , , | 215/749 | (28.7%) | 191/737 | (25.9%) | 1.11 (0.94–1.31) | 0.21 | 0 | NA |
| Any maternal adverse event | 3 , , | 385/740 | (52.0%) | 369/718 | (51.4%) | 1.01 (0.89–1.15) | 0.88 | 41 | NA |
| Discontinuation of treatment because of adverse events | 4 , , , | 44/790 | (5.6%) | 31/768 | (4.0%) | 1.38 (0.88–2.14) | 0.16 | 0 | NA |
| Preterm prelabor rupture of membranes | 4 , , , | 88/794 | (11.1%) | 85/775 | (11.0%) | 1.02 (0.77–1.35) | 0.87 | 0 | NA |
| Preeclampsia | 2 , | 12/440 | (2.7%) | 8/435 | (1.8%) | 1.48 (0.61–3.58) | 0.38 | NA | NA |
| Gestational hypertension | 2 , | 6/440 | (1.4%) | 5/435 | (1.2) | 1.19 (0.37–3.85) | 0.78 | 0 | NA |
| Gestational diabetes mellitus | 2 , | 45/440 | (10.2%) | 40/435 | (9.2%) | 1.12 (0.75–1.67) | 0.59 | 0 | NA |
| Respiratory distress syndrome | 6 , , | 84/896 | (9.4%) | 114/883 | (12.9%) | 0.62 (0.37–1.04) | 0.07 | 57 | 0.18–2.13 |
| Necrotizing enterocolitis | 4 , , , | 5/782 | (0.6%) | 8/766 | (1.0%) | 0.64 (0.22–1.90) | 0.42 | 0 | NA |
| Intraventricular hemorrhage | 4 , , , | 17/801 | (2.1%) | 15/788 | (1.9%) | 1.11 (0.56–2.22) | 0.76 | 0 | NA |
| Grade III or IV intraventricular hemorrhage | 4 , , , | 2/782 | (0.3) | 2/766 | (0.3) | 0.98 (0.14–6.94) | 0.98 | 0 | NA |
| Neonatal sepsis | 5 , | 15/827 | (1.8%) | 22/825 | (2.7%) | 0.69 (0.29–1.68) | 0.42 | 26 | 0.14–3.32 |
| Retinopathy of prematurity | 2 , | 12/422 | (2.8%) | 9/414 | (2.2%) | 1.32 (0.56–3.09) | 0.53 | NA | NA |
| Bronchopulmonary dysplasia | 3 , , | 10/473 | (2.1%) | 5/464 | (1.1%) | 1.97 (0.68–5.71) | 0.21 | NA | NA |
| Periventricular leukomalacia | 2 , | 0/423 | (0.0) | 1/414 | (0.2) | 0.33 (0.01–8.03) | 0.49 | NA | NA |
| Fetal death | 6 , , , | 16/1271 | (1.3%) | 15/1251 | (1.2%) | 1.05 (0.52–2.13) | 0.89 | 0 | 0.52–2.13 |
| Neonatal death | 7 , | 20/1340 | (1.5%) | 32/1323 | (2.4%) | 0.65 (0.36–1.15) | 0.14 | 0 | 0.36–1.15 |
| Perinatal death | 6 , , , | 33/1271 | (2.6%) | 37/1251 | (3.0%) | 0.90 (0.56–1.45) | 0.67 | 0 | 0.56–1.45 |
| Birthweight <1500 g | 4 , , , | 55/781 | (7.0%) | 42/762 | (5.5%) | 1.28 (0.87– 1.89) | 0.21 | 0 | NA |
| Birthweight <2500 g | 5 , | 206/850 | (24.2%) | 229/834 | (27.5%) | 0.77 (0.54–1.10) | 0.15 | 64 | 0.33–1.82 |
| Admission to NICU | 6 , , | 129/896 | (14.4%) | 183/883 | (20.7%) | 0.53 (0.33–0.85) | 0.01 | 67 | 0.16–1.79 |
| Use of mechanical ventilation | 5 , | 78/825 | (9.5%) | 104/825 | (12.6%) | 0.65 (0.39–1.08) | 0.10 | 44 | 0.21–2.00 |
| Patent ductus arteriosus | 3 , , | 18/721 | (2.5%) | 15/718 | (2.1%) | 1.19 (0.61–2.36) | 0.61 | 0 | NA |
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