ABSTRACT

Objective:

Gestation weight (GW), body mass index (BMI), and blood 25-hydroxyvitamin D [25(OH)D] level during pregnancy are important determinants of the gestational outcomes. This study aimed to study how these parameters vary between antenatal vitamin D recipients and non-recipients in gestational diabetes mellitus (GDM) patients.

Material and Methods:

The randomized controlled trials comparing these outcomes between vitamin D recipient and non-recipient GDM patients were searched in electronic databases (PubMed, Embase, and Scopus). The reviewed studies’ data were abstracted and critically appraised using the Cochrane tool. The estimation of the weighted mean difference for GW and BMI and standardized mean difference (SMD) for 25(OH)D levels occurred by juxtaposing the interventions meta-analytically (random-effect model). The statistical inconsistency was determined by Chi² and I² method. The statistical significance was estimated at p<0.05 and 95% confidence interval (CI).

Results:

Eleven eligible trials (all Iran-based, except one), sourcing data from about 875 GDM patients, were reviewed. Overall, the risk of bias was low, except for selection and performance bias. On random-effect model meta-analysis, the 25(OH)D levels of the GDM patients favored the vitamin D recipients when compared to non-vitamin D (SMD 1.97, 95% CI: 1.06-2.88, p<0.001; I² 96.2%, p of Chi² <0.001) and placebo (SMD 1.86, 95% CI: 0.95-2.77, p<0.001; I² 95.3%, p of Chi² <0.001) recipients, respectively. On meta-regression, sample size was a predictor of the observed heterogeneity. For GW and BMI the interventions did not differ statistically significantly.

Conclusion:

In GDM patients, antenatal use of vitamin D aids in the rise of blood 25(OH)D levels. However, vitamin D supplementation did not affect change in GW or BMI.

Keywords:

Gestational diabetes, vitamin D, dietary supplement

Inclusion criteria

1. Study design: Parallel arm randomized controlled trials of any number of intervention arms.

2. Population: Pregnant females of any age were eligible, irrespective of their pre-pregnancy BMI and 25(OH)D levels. They must be diagnosed with GDM during their concurrent pregnancy.

3. Intervention arm: The treatment arm/s should have received vitamin D as a sole or co-supplement.

4. Comparator arm: The comparator arm/s may have received a placebo or any other supplement except vitamin D. Comparator arm/s not receiving any intervention were also eligible.

5. Outcomes: The trials must report the GW (kg), BMI (kg/m²), and 25(OH)D (in ng/mL or mmol/L) in the above GDM patients before and after receiving these interventions and before childbirth.

We accepted the diagnosis and management of GDM and the dosage and regimen of interventions received by the participants in the respective treatment arms as per the trialists.

Exclusion criteria

1. Study design other than those described above, e.g., observational studies and crossover studies.

2. Participants with diabetes types besides GDM, like type 1 or type 2 diabetes.

3. Studies conducted on animals.

4. Editorials, abstracts from conference presentations (where a full published manuscript is not available), letters, or any other brief communications.

Statistical analysis

Using random-effect and fixed-effect models, all outcomes were compared meta-analytically between vitamin D and placebo-receiving GDM patients.

We estimated the statistical significance of meta-analysis derived effect sizes at p<0.05 and 95% confidence interval (CI). Stata statistical software v16 (StataCorp, College Station, TX) was used for analysis.

Scope of the review

The database search retrieved 271 citations. After eliminating the duplicates, 188 citations underwent skimming against the eligibility criteria. Out of the 22 articles needing full-text reading, 11 trials sourcing data from about 875 participants published between 2014-19, were included in this review (Figure 1) (15,16,17,18,33,34,35,37,49,50,51). All trials except the Chinese one (37) were Iran-based, and the average age of participants in the respective intervention arms was approximately 28-32 years. The intervention period of Iranian (16,17,18,33,49,50,51) and Chinese (37) trials were 6-8 and 16 weeks, respectively. In most trials (15,16,17,18,34,35,37,49,50,51), GDM was diagnosed primarily using the American Diabetes Association criteria (52,53). Insulin was not used during the intervention period, except in the trial by Yazdchi et al. (33). Eight trials (15,16,18,33,34,37,49,50) used the D3 form of the vitamin while this was not clear among the remaining trials (17,35,51). In most of the trials (81.8%), a co-supplement (e.g., calcium, magnesium, zinc, omega-3 fatty acid, evening primrose oil, probiotic) accompanied the vitamin D supplementation (15,16,17,18,34,35,37,50,51). The intervention was given between 24-28 weeks of gestation in nine trials (15,16,17,18,33,34,35,49,50), at 16 weeks of gestation in one trial (37), and in the remaining one, this was unclear (51). Table 1 depicts the salient features of the trials.

RoB assessment

In most studies, the allocation concealment component of the selection bias and performance bias was unclear (Table 2 and Figure 2). Otherwise, the RoB was low.

Meta-analysis findings

Eleven trials comparing GW (15,16,51,17,18,33,34,35,37,49,50), and 10 trials contrasting BMI (15,16,17,18,33,34,35,49,50,51) with one study (37) excluded as it did not report the follow up BMI, and 25(OH)D with one trial (33) excluded for reporting follow up value in median, were included in the meta-analytic juxtaposition between vitamin D recipients and its non-recipients.

The antenatal vitamin D use in GDM patients favored plasma 25(OH)D level attainment compared to its non-supplementation (random-effect model: SMD 1.97, 95% CI: 1.06-2.88, p<0.001; I² 96.2%, p of Chi² <0.001).

The post-intervention GW (random-effect model: WMD 0.18, 95% CI, -1.10-1.47, p=0.773; I² 0%, p of Chi² 0.559) and BMI (random-effect model: WMD 0.27, 95% CI, -0.28-0.82, p=0.331; I² 0%, p of Chi² 0.838) were not statistically significantly different between the juxtaposed interventions (Figure 3).

Meta-regression and subgroup analysis

The univariate meta-regression suggested that sample size was a statistically significant predictor of the observed heterogeneity in the effect size of 25(OH)D level (Supplemental Table 2). Upon subgroup analysis by the sample size, heterogeneity was moderate when sample size was ≥100, and the effect size increased (random-effect model: SMD 3.81, p<0.001; 95% CI, 3.03-4.59; I² 72.5%) (Supplemental Figure 1).

Publications bias

For 25(OH)D, a small study effect was suggested by the asymmetric funnel plots (Supplemental Figure 2) and Egger’s test (p=0.005). On trim-and-fill analysis, no additional study was imputed. Funnel plots for the rest of the outcomes were approximately symmetric.

Sensitivity analysis

On using a fixed-effect model meta-analysis, the summary estimate of 25(OH)D level, reduced slightly (SMD 1.74, 95% CI, 1.57-1.92, p<0.001). The fixed-effect meta-analysis results for the rest of the outcomes were identical to the preliminary analysis. The meta-analysis findings for all outcomes remained unchanged on dropping a study each time and repeating the meta-analysis.

Supplementary meta-analysis

Between vitamin D and placebo, ten trials (15,16,17,18,33,34,35,49,50,51) compared GW and BMI, and nine trials (15,16,17,18,34,35,49,50,51) juxtaposed 25(OH)D levels, with one study (33) excluded because the study reported 25(OH)D values as medians. Vitamin D recipients achieved a favorable blood 25(OH)D level compared to the placebo recipients (random-effect model: SMD 1.86, 95% CI, 0.95-2.77, p<0.001; I², 95.36%, p of Chi² <0.001) (Figure 4). The effect size of 25(OH)D levels reduced slightly on using a fixed-effect meta-analysis model (SMD 1.45, 95% CI, 1.25-1.64). GW and BMI, when contrasted among the intervention arms, were not statistically significantly different. Since <10 studies were available for the 25(OH)D levels, we did not explore heterogeneity or assess the publication bias for it.

Evidence quality

Utilizing the GRADE Working Group’s (2004) (54) approach of grading evidence quality we graded the evidence concerning the 25(OH)D level as of moderate-quality, due to the unclear RoB components and heterogeneity.

Comparison with what is known

As the context remains underexplored in contemporary literature, a direct juxtaposition of our findings to existing reviews is not possible. However, clinical trials studying the effect of vitamin D supplementation on 25(OH)D level in pregnant females with no glucose intolerance are available for a contrast. Two such trials found that vitamin D supplementation in the third trimester increased maternal plasma 25(OH)D levels compared to the control group (55,56). Another randomized trial found that vitamin D supplementation caused a statistically significantly greater increase in the 25(OH)D level than the placebo (57). Mirroring these trials’ findings (55,56,57), we observed that vitamin D3 supplementation in GDM patients increased the maternal 25(OH)D level.

Implications and strengths

The chief inference of this paper is that it informs about the rigor of the current evidence of the maternal benefits of prenatal vitamin D supplementation in GDM patients. From the perspective of maternal health, this study may help health authorities to determine if large scale supplementation for all GDM pregnancies will be an appropriate public health initiative or not, given the current evidence. Moreover, as the reviewed trials were primarily Iran-based, this paper might encourage future trialists to conduct identical trials globally to generate generalizable evidence. Concerning the strength, this systematic review is one of the preliminary efforts to investigate the maternal effects of antenatal vitamin D supplementation in GDM patients. A further strength was the unbound nature of our electronic database searche to include any date, language, or geographical boundary, thus adding comprehensiveness to our review. Finally, evidence generated from this systematic review and meta-analysis is likely to be rigorous, as it's grounded on the highest level of epidemiological evidence, randomized controlled trials.

Despite these strengths, this paper has a few weaknesses. As most trials were conducted in Iran, the external validity of this review is likely to be compromised. The heterogeneity observed for the 25(OH)D levels might have increased the risk of bias in our estimates, and this can be because one of the studies was not from Iran. Besides, the maternal health effects of vitamin D supplementation remains inseparable from other supplements that were simultaneously given to the participants in most trials.

Introduction

Gestational diabetes mellitus (GDM) refers to any degree of glucose intolerance that develops or is identified initially during gestation (1). Its global prevalence is about 7-10% (2,3,4,5). GDM diagnosis is made using glucose challenge tests between 24-28 weeks of gestation (1). Initial GDM management encompasses diet and exercise therapy, but if these fail to achieve glycemic control, physicians start insulin therapy (1).

GDM is a crucial health burden since it can affect both the GDM patient and her neonate. Gestational weight (GW) and body mass index (BMI) in pregnancy are two important anthropometric determinants of GDM-related outcomes. Studies in GDM patients suggest that an excessive GW accumulation increases the risk of maternal complications, such as increased likelihood of cesarean delivery, large for gestational age, and gestational hypertension and fetal problems, such as macrosomia, large for gestational age, hypoglycemia in newborns, and poor APGAR score (6,7,8,9,10,11). It remains unclear if the Institute of Medicine’s guideline (2009) for recommended GW gain for respective BMI categories can be applied to the GDM subpopulation or not (8,12). However, studies on overweight and obese GDM patients found that gaining GW less than that recommended for their respective BMI categories resulted in favorable obstetric and neonatal outcomes (8,13,14). Maintaining an optimum weight before and during pregnancy, therefore, can decrease the complications of pregnancy (9). Nevertheless, it remains unclear if any antenatal intervention in GDM patients may be beneficial in achieving an acceptable GW and BMI.

In this respect, vitamin D has emerged as a potentially useful agent that has attracted attention. In GDM patients, various clinical trials (15,16,17,18) have tested the maternal health effect of antenatal vitamin D supplementation, and due to the different relationships between GDM and vitamin D status in the body, such testing appears pertinent. For instance, inadequate vitamin D levels in the body are associated with an increased risk of developing GDM (19,20,21,22,23). Vitamin D deficiency (<20 ng/mL) has a nearly fourfold increased risk of GDM development than women with sufficient vitamin D level (>30 ng/mL) after adjusting for the age of the mother, race, ethnicity, and family history of type 2 diabetes among first-degree family members (24). Moreover, studies showed a decreased GDM prevalence in prenatal vitamin D recipients (25,26). Given this evidence it is important to understand how vitamin D supplementation in GDM mothers can affect GW and their BMI. Additionally, as the fetus entirely depends on maternal 25-hydroxyvitamin D [25(OH)D] levels, maternal levels in GDM mothers after vitamin D supplementation also requires evaluation (27).

Intervention description

The inactive forms of the fat-soluble vitamin D are D2 (ergocalciferol) and D3 (cholecalciferol) (28,29). Both forms are available from diet and supplements and vitamin D3 is also produced in the skin on exposure to sunlight (28,29). On hydroxylation of pre-vitamin D in the liver, the main circulating form, 25(OH)D, of vitamin D is produced (27). In blood, 25(OH)D is either present in the bound form (to albumin) or free form (27). For its physiologic role, it is converted to the active form, calcitriol 1,25-dihydroxyvitamin D (28,30). The physiological effect of Vitamin D in pregnancy is mediated via calcitriol’s action on the vitamin D receptors in uteroplacental tissue (28,30). Compared to calcitriol, which has a half-life of 4-6 hours (27), the relatively longer half-life of 25(OH)D of between two and three weeks (31) makes the latter an ideal marker for vitamin D status (32).

In GDM patients, contemporary trials have supplemented vitamin D at various dosages. For oral preparations, while some trials used it at a dose of 50,000 IU, two to three weeks apart for three to eight weeks (33,34,35,36), other trials used it twice daily at 200-500 IU for six to sixteen weeks (17,37). One trial used a single intramuscular injection of vitamin D at a dose of 300,000 IU (38). Furthermore, while few trials used the vitamin as a single supplement (33,37,38), others co-supplemented it with various micronutrients, including zinc, magnesium, and calcium (17,34).

What this review adds?

In contemporary medicine, several clinical trials have tested the changes in GW, BMI, and plasma 25(OH)D level in GDM mothers, after antenatal vitamin D supplementation (15,34,35,36). Recent reviews have studied the effect of antenatal vitamin D supplementation on certain maternal complications such as cesarean section rate, pre-eclampsia, preterm delivery, macrosomia, and polyhydramnios and/or on neonatal complications including hyperbilirubinemia, hypoglycemia, and hospitalization (39,40,41). However, to the best of our knowledge, there is no systematic review and meta-analysis that studied how maternal GW, BMI, and 25(OH)D levels change in the blood on vitamin D supplementation in GDM patients. Therefore, this study explores this under-reviewed area of modern medicine by a systematic literature search, critical appraisal, and meta-analysis.

Aims

This study compared the GW, BMI, and 25(OH)D levels among vitamin D supplemented and not-supplemented GDM patients.

Material and Methods

This systematic review is registered in the PROSPERO database (CRD42020149613) and has a pre-published protocol (42,43). This report adheres to the PRISMA 2009 reporting guideline (Supplemental Table 1) (44).

Database search

We searched the title and abstract of prospective trials matching the above eligibility criteria in PubMed, Embase, and Scopus databases, irrespective of the date and language of publication and geographical boundary. The following search terms were used “vitamin D” OR “calciferol” OR “vitamin D2” OR “ergocalciferol” OR “vitamin D3” OR “cholecalciferol” AND “GDM” OR “gestational diabetes” along with these MeSH terms- “Cholecalciferol”, “ergocalciferols”, and “diabetes, gestational”. To identify the clinical trials in PubMed [(Clinical Trial) and (Randomized Controlled Trial)] and Embase [(controlled clinical trial) and (randomized controlled trial)], we used filters. In Scopus, instead of filters, the following search terms were used: “trial,” “randomised,” “randomized,” and “controlled.” The last date of the search was 17 September, 2020. Additionally, we reviewed the references of the papers included in this review.

We uploaded the retrieved citations (from database search) in the Rayyan systematic review software (45) and eliminated the duplicate articles. Successively, skimming of the remaining citations’ titles and abstracts against the eligibility criteria commenced. Articles were read in full-text when it seemed to meet the inclusion criteria, or if the suitability for incorporation in this review was doubtful.

Data abstraction and risk of bias assessment

We extracted data about the study design, consent, ethics, registration number of the trial, participant features, interventions contrasted, and the outcomes of interest in a pre-piloted form. With the Cochrane collaboration tool, individual trial's risk of selection bias, performance bias, detection bias, attrition bias, reporting bias, and any other bias was determined, and each of these risk of bias (RoB) components was categorized as low, high, or unclear (46). To assess selection bias, the random allocation sequence generation method, and its concealment method from participants, were judged. The blinding mechanism of study participants and personnel and that of outcome assessors were used to evaluate the performance and detection bias, respectively. By evaluating missing outcome data, and its reason among the intervention arms, the risk of attrition bias, was evaluated. Any additional bias, besides the above, comprised the other bias type. For a visual presentation of the RoB, we prepared an RoB graph and an RoB summary using the Review Manager (RevMan) software (46,47).

The review authors independently performed study selection, data abstraction, and RoB assessment, and resolved any disagreement in an opinion by discourse.

Meta-analysis

The juxtaposed interventions’ effect on each of the outcomes was contrasted by random-effects meta-analysis (using DerSimonian and Laird method) since we assumed clinical heterogeneity among the trials attributable to the different types of vitamin D co-supplements used in these. The use of endpoint means of the respective outcomes and their SDs ensued to conduct the meta-analysis. We estimated the meta-analytic effect sizes of GW and BMI in weighted mean differences (WMD) and that of 25(OH)D levels in standardized mean differences (SMD) due to the identical and non-identical types of measuring units used in the trials, respectively. A decrease in the summary effect of GW and BMI, and its increase in 25(OH)D levels, denoted a favorable meta-analytic finding. For any outcome, when multiple treatment arms tested an intervention, the post-intervention means and their SDs of those intervention groups were combined for meta-analysis (46). Outcome reported in the median were not considered for meta-analysis.

Heterogeneity and meta-regression

The statistical heterogeneity was determined by Chi² (statistically significant at p<0.1) and I² (categorized as low, moderate, and high at I² values of 25, 50, and 75%, respectively) statistics (48). To account for any substantial heterogeneity, we performed univariate meta-regression by presence or absence of missing outcome data and sample size (categorized as <100 and ≥100). Using the predictor identified by meta-regression, we did a subgroup analysis to see how heterogeneity changed across the different categories of the predictor.

Publication bias and sensitivity analysis

The publication bias assessment incorporated visual inspection of funnel plots and Egger’s test. For each outcome, a sensitivity analysis included iteration of the meta-analysis using a fixed-effect model and by dropping a trial each time.

Results

Discussion

Overall, 11 trials, mostly Iranian, tested the effect of antenatal vitamin D complementation (as a co-supplement primarily) on GW, BMI, and 25(OH)D in 875 GDM patients, were retrieved. The intervention favored a rise in blood 25(OH)D levels, and the sample size was the plausible predictor of the observed heterogeneity.

Conclusion

Using vitamin D as the chief ingredient of antenatal supplements favors in blood 25(OH)D level rise in GDM patients. However, the effect of these supplements on GW and BMI was not distinguishable from those subjects who did not receive supplementation.

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