Background {#Sec1}
==========

Labour induction refers to the process of artificially stimulating the uterus to begin labour \[[@CR1]\], which is an increasingly common procedure. Cervical status, measured by the Bishop score \[[@CR2]\], is a good predictor for the outcome of labour induction. If the cervix is unfavourable, no method is highly successful, and a ripening process is generally employed to obtain cervical effacement and dilatation prior to induction \[[@CR3]--[@CR5]\]. Methods used for cervical ripening can be broadly divided into mechanical devices and pharmacologic options \[[@CR6], [@CR7]\]. Compared with pharmacologic agents, mechanical methods, which were the first methods developed to ripen the cervix or induce labour \[[@CR8]\], have similar levels of effectiveness but incur fewer episodes of adverse events (such as uterine tachysystole), have lower costs and are easier to preserve \[[@CR6]\].

The balloon catheter, including both double- and single-balloon catheters, appears to be a widely accepted mechanical method and is recommended by the WHO for the induction process \[[@CR9]\]. The original version of the Foley (single-balloon) catheter was initially described by Barnes in 1863 but was not described again until 1967, by Embrey and Mollison \[[@CR10]\]. In 1991, Atad described the first double-balloon variation \[[@CR11]\]. The Cook Cervical Ripening Balloon (CCRB), which uses an identical mechanism to that of the Atad catheter, was approved by the United States Food and Drug Administration (USFDA) in 2013 \[[@CR12]\]. Only the double-balloon catheter (either Atad or Cook) is specifically designed and licensed for labour induction, while the Foley catheter is used beyond instructions.

Mechanical ripening devices apply pressure to the internal face of the cervix, directly overstretching the lower uterine segment and indirectly increasing the localised secretion of prostaglandin \[[@CR13]\]. In addition to the local effect, mechanisms that involve neuroendocrine reflexes (such as the Ferguson reflex) may promote the onset of contractions \[[@CR14]\]. Purportedly, the double-balloon (either Atad or Cook) option has an additional cervico-vaginal balloon, which applies greater pressure to both sides of the cervix and avoids the need for traction \[[@CR11]\].

Given the increasing induction rate, the knowledge of even small differences between methods could be useful, not only to guide clinical practices but also to further explore the mechanism underlying the mechanical induction of labour and may promote a better understanding of the optimal methods for labour induction. However, studies examining the superiority of the double-balloon catheter reveal mixed results \[[@CR15]--[@CR23]\]. We conducted this meta-analysis and systematic review using the best available evidence to assess the efficacy, efficiency, safety and patient satisfaction of double-balloon catheters in comparison with those of single-balloon devices among women who underwent labour induction with unfavourable cervixes.

Methods {#Sec2}
=======

Search strategy {#Sec3}
---------------

Together with a clinical librarian (R.O.), an electronic literature search was conducted with the PubMed, EMBASE, OVID, SCI (via WOS), CENTRAL (The Cochrane Central Register of Controlled Trials) and [ClinicalTrial.gov](http://clinicaltrial.gov) databases from inception through ebruary 14, 2018. The searching strategy was based on the PICOS principle, utilising medical subject headings and Boolean logic-based free-text combinations of the following search items: "labour induction", "cervical ripening", "balloon", "Foley", "Cook" and "Atad". In addition, we used sensitivity-maximising search filters to identify randomised controlled trials \[[@CR24]\]. With the abovementioned databases, several meta-analyses and systematic reviews were identified. Aiming to identify more pertinent meta-analyses or systematic reviews, an additional search was performed in the CDSR (Cochrane Database of Systematic Reviews) database. All of the reference lists from the relevant reviews were manually retrieved to locate further eligible trials. There were no language restrictions. Differences of opinion were resolved by team discussion.

Study selection and data collection {#Sec4}
-----------------------------------

All related RCTs that directly compared the double-balloon catheter with the single-balloon catheter for the purposes of labour induction or cervical ripening were included in the analysis. There were no restrictions with regards to settings, demographics, obstetrics characteristics (e.g., race, maternal age, and gestational weeks) and outcome measures. We excluded the following types of studies: (1) studies of balloon catheters used for outpatient purposes; and (2) protocols, observational studies, and secondary analyses of previous studies and guidelines. Prior to the formal review process, we performed pretesting with the kappa statistic to calculate the level of agreement between the inclusion/exclusion decisions of different reviewers and adjusted our criterion until kappa ≧ 0.75.

To improve the precision of the collected data, two reviewers (X.Y.L., Y.W.), one who majored in obstetrics and one who did not, screened each record for eligibility and independently extracted and tabulated the following information from the text, tables, and graphs: lead author; publication year; country of origin; study design; participants and intervention characteristics; outcomes; and sponsor. Prior to determining the categories for the data collection forms, a pilot test was performed using representative samples of the studies to be reviewed. All of the collected data are available upon request.

Due to the uncertain benefits of blinded assessments and the large workload, we did not conceal the general contents of the studies during this process. Any disagreements were resolved through discussion, or if necessary, through consultation with a third reviewer (F.Z.) who specialises in evidence-based medicine. When information regarding any of the extracted data points listed above was unclear, an attempt was made to access further details by contacting the authors of the original reports.

Selection of outcomes {#Sec5}
---------------------

The primary and secondary outcomes were defined before trial retrieval was performed. The primary outcome was the caesarean delivery rate. The secondary outcomes included: (1) catheter placement (placement difficulty/failure, spontaneous expulsion); (2) intervals (insertion to delivery, insertion to expulsion/removal, expulsion to delivery); (3) Bishop score increment; (4) vaginal delivery (vaginal delivery within 24 h, normal vaginal delivery, assisted vaginal delivery); (5) analgesia usage; (6) maternal adverse events (death, infection, postpartum haemorrhage); (7) neonatal adverse events (death, low Apgar score, NICU admission); (8) length of hospitalisation; and (9) satisfaction **(**pain during the process, maternal total satisfaction). While we attempted to collect all of the above datapoints from all of the analysed studies, only those that provided all of the data appear in the analysis tables.

Quality assessment {#Sec6}
------------------

Two independent investigators (X.Y.L., Y.W.) openly (not blinded) assessed the methodological quality of the included RCTs based on Cochrane risk-of-bias tool. Quality was graded based on the following criteria \[[@CR25]\]: (1) high quality: both randomisation and allocation concealment were assessed as having low risks of bias, and all other items were assessed as having low or unclear levels of risk; (2) low quality: either randomisation or allocation concealment was assessed as having a high risk of bias, regardless of the risk levels of other items; and (3) moderate quality: trials did not meet the criteria for high or low quality. Discrepancies were resolved by consensus.

Statistical analysis {#Sec7}
--------------------

All statistical analyses were performed with RevMan version 5.3, with the help of a statistician (X.N.Z.). The relative risks (RRs) and mean differences (MDs), with corresponding 95% confidence intervals (CIs), were used to describe the intervention effects for dichotomous and continuous variables, respectively. All potential data conversions utilised standard formulae recommended by the Cochrane Handbook \[[@CR24]\].

Heterogeneity was identified by Cochrane's Q test and the I^2^-statistic test, in which a Q test *p*-value \< 0.1 and an I^2^ value ≥50% indicated significant heterogeneity. When both the *p*-value and the I^2^ value displayed no heterogeneity, we chose the fixed-effect model. Else, a random-effect model was used.

Subgroup analysis was pre-specified and performed on parity. A sensitivity analysis was conducted to identify studies involving data conversions that may have exerted a disproportionate influence on the pooled estimates. We assessed publication bias by examining funnel plots for the primary outcome only.

Results {#Sec8}
=======

Study characteristics {#Sec9}
---------------------

The literature search and screening process is shown in Fig. [1](#Fig1){ref-type="fig"}. Initially, 1326 potentially relevant records were identified. The titles and abstracts were reviewed, and 12 relevant trials were further screened. After thorough investigation, 7 RCTs, containing 1159 women and available data (577 and 582 in the double- and single-balloon groups, respectively), were determined to be eligible for inclusion \[[@CR15]--[@CR21]\]. The characteristics of the included trials are summarised in Table [1](#Tab1){ref-type="table"}. Table [2](#Tab2){ref-type="table"} shows the risk of bias and the corresponding quality of each individual trial, which is illustrated in Fig. [2](#Fig2){ref-type="fig"}a and b. Basic demographic and obstetric variables are presented in Table [3](#Tab3){ref-type="table"}. Except for postdates, which only two studies reported and which show slight heterogeneity, all other variables were comparable. Fig. 1Literature search and screening process Table 1Characteristics of the included trialsStudyPeriodCountryMethodSample size total (double/single)Parity total (double/single)Balloon CatheterDouble^x^Single^y^Ahmed 20162013.03--2014.04EgyptRCT78 (39/39)all nulliparousCookFoley (50 ml)Haugland 20122010.03--2011.01NorwayRCT178 (88/90)NACookFoley (NA)Hoppe 20152010.01--2013.11USARCT98 (50/48)nulliparous: 50 (25/25)multiparous: 48 (25/23)CookFoley (30 ml)Pennell 2009\*2001.07--2003.12AustraliaRCT217 (107/110)all nulliparousAtadFoley (30 ml)Rab 20152011.01--2013.12EgyptRCT200 (100/100)nulliparous: 113 (55/58)multiparous: 87 (45/42)CookFoley (30 ml)Salim 20112008.06--2010.12IsraelRCT293 (148/145)nulliparous: 155 (78/77)multiparous: 138 (70/68)CookFoley (60 ml)Solt 20092006.01--2008.05IsraelRCT95 (45/50)^§^nulliparous: 95 (45/50)multiparous: 85 (NA)^§^CookFoley (NA)Except for two studies \[[@CR16], [@CR21]\], in which we could not find detailed information, all studies offered similar standard instructions for how to use the balloon cathetersx: COOK/Atad: 80 ml + 80 ml, without tensiony: All Foley catheters were applied with light tensionNA: Data not found: unable to contact the authors of the original reports\*: Pennell 2009 \[[@CR18]\] was a multi-arm study, and we only extracted data for the double-balloon catheter and single-balloon catheter comparison arms§: Solt 2009 \[[@CR21]\] only reported the results of nulliparous women; therefore, we eliminated the multiparous subgroup and extracted nulliparous data only. Table 2Risk of bias and corresponding qualityStudyRisk of bias for the included studiesQualityRandom sequence generation (selection bias)Allocation concealment (selection bias)Blinding of participants and personnel (performance bias)^x^Blinding of outcome assessment (detection bias)Incomplete outcome data (attrition bias)Selective reporting (reporting bias)^y^Other biasAhmed 2016LLUULLLHHaugland 2012LULLLLUMHoppe 2015^A^LLHHLLLMPennell 2009LLHLLLLHRab 2014LUULLLLMSalim 2011^A^LLHULLLHSolt 2009^B^LLHLLHUMOther bias: trials sponsored by drug companies or trials in which baseline characteristics were not similar between different intervention groupsL: Low risk or low quality. For the risk of bias, L means appropriate methods were adequately describedH: High risk or high quality. For the risk of bias, H means high risk was found (unable to avoid bias)U: Unclear risk, no descriptionM: Moderate qualityA: Hoppe et al. \[[@CR17]\] and Salim et al. \[[@CR20]\] reported a few lost follow-ups and did not perform intention-to-treat (ITT) analyses. Fortunately, they were balanced in numbers with similar reasons across intervention groups and had little influence on the following analysis. Therefore, we assessed the attrition bias as lowB: Solt et al. \[[@CR21]\] only described a single-blind method. Considering the nature of the study, we evaluated performance bias as H, while detection bias was evaluated as L. Additionally, they selectively reported outcomes for the nulliparous group, and we could not obtain supplemental data for the multiparous group by contacting the authorx: Blinding of participants and personnel, though graded, was excluded from the quality assessment because it was impractical for these trialsy: Some studies did not offer their protocols; therefore, it is difficult hard to determine whether the outcomes were not measured or not reported. Unless selective reporting was obvious, we evaluated this situation as being low risk Fig. 2**a** Risk of bias graph. **b**. Risk of bias summary Table 3Basic demographic and obstetric variablesVariables\*Number of studies analysedInterventionsPooled effect (95% CI)^x^Q *p*-value^y^I^2^- statistic^y^*p*-value^z^DoubleSingleMaternal age (years)^A^5 \[15,17--20\]4444420.73 \[−0.08, 1.53\]0.8900.08Gestational weeks (weeks)^A^5 \[15,17--20\]444442−0.07 \[− 0.31, 0.17\]0.7100.56Bishop score at catheter insertion^B^4 \[15,17,19--20\]3373320.02 \[−0.19, 0.23\]0.4200.85Induction indications Postdates^C^2 \[17--18\]1571581.10 \[0.51, 2.37\]0.15**51%**0.81 Diabetes mellitus3 \[17--18,20\]3053031.35 \[0.80, 2.26\]0.2039%0.26 Hypertensive disease3 \[17--18,20\]3053030.98 \[0.66, 1.44\]0.2135%0.91 Intrauterine growth restriction2 \[17--18\]1571580.92 \[0.39, 2.21\]0.2329%0.86\*: Only those variables contained in more than one study are displayed in this analysis tableOnly Hoppe 2015 \[[@CR17]\] described oligohydramnios and abnormal foetal monitoring and did not reveal significant differencesRab 2014 \[[@CR19]\] mentioned data on body mass index (BMI) at insertion, while Salim, 2011 \[[@CR20]\] reported BMI data before pregnancyNot all of the baseline data was described in Haugland, 2012 \[[@CR16]\] and Solt, 2009 \[[@CR21]\]A: Pennell 2009 \[[@CR18]\] described the maternal age and gestational weeks by median and typical range. The results remained the same after sensitivity analysisB: Pennell 2009 \[[@CR18]\] described the Bishop score using ordinal data, making synthesis impossibleC: Postdates, as one of the induction indications, shows heterogeneity between the two studies \[[@CR17], [@CR18]\]x: The odds ratio (OR) was the pooled effect for dichotomous variables. The mean difference (MD) was the pooled effect for continuous outcomesy: Study heterogeneity was explored using Cochrane's Q test and the I^2^-statistic test. Random- and fixed-effect models were used as appropriatez: The *p*-value ≥0.05 for the test indicated that the pooled effect was insignificant; for dichotomous outcomes, the test was OR = 1; for continuous outcomes, the test was MD = 0

Of the 7 RCTs, 3 trials \[[@CR15], [@CR18], [@CR21]\] focused on nulliparous women, while 2 trials \[[@CR17], [@CR20]\] conducted subgroup analysis by parity. These 5 trials, which included 781 women (595 nulliparous and 186 multiparous), were suitable for parity subgroup analysis.

Effects of interventions {#Sec10}
------------------------

All trials reported the rates of caesarean section. There were no significant differences in the rates of caesarean delivery (RR, 0.88 \[0.65, 1.2\]; *p*-value, 0.43) among trials, but heterogeneity existed (Q *p*-value, 0.04; I^2^, 55%) (Fig. [3](#Fig3){ref-type="fig"}a). A corresponding funnel plot is shown in Fig. [3](#Fig3){ref-type="fig"}b. During sensitivity analysis, heterogeneity disappeared only when Salim 2011 \[[@CR20]\] was excluded (Q *p*-value, 0.11; I^2^, 45%), while the pooled effect was always robust (no significant differences). The secondary outcomes, shown in Table [4](#Tab4){ref-type="table"}, did not differ obviously between the two types of catheter, except for the Bishop score increment (MD, 0.57 \[0.28, 0.86\]; *p*-value, 0.0001). Fig. 3**a** Forest plot of cesarean delivery. **b**. Funnel plot of cesarean delivery Table 4Secondary outcomesOutcomesNumber of studies analysedInterventionsEffect measurePooled effect\
(95% CI)Q *p*-valueI^2^- statistic*p*-valueSensitivity analysisDoubleSingleplacement difficulty/failure4 \[15--16,18,20\]382384RR1.34 \[0.66, 2.71\]0.5500.42stablespontaneous expulsion4 \[15,17,19--20\]337332RR0.86 \[0.60, 1.23\]**0.000285%**0.40stableinsertion to delivery interval^A^6 \[15--20\]532532MD0.98 \[− 0.03, 2.00\]0.1636%0.06**Unstable** ^I^1.36 \[−0.63, 3.34\]**0.0261%**0.18**Unstable** ^II^insertion to expulsion/removal interval^B^4 \[15,17,19--20\]278267MD1.72 \[−0.35, 3.79\]**\< 0.0000193%**0.10**Unstable** ^III^expulsion to delivery interval2 \[19,21\]145150MD−2.81 \[−10.82, 5.19\]**0.0672%**0.49--Bishop score increment^C^5 \[15,17,19--21\]382382MD0.57 \[0.28, 0.86\]0.3018%**0.0001**stablevaginal delivery within 24 h3 \[17--18,20\]305303RR0.95 \[0.72, 1.26\]0.11**54%**0.74**Unstable** ^IV^normal vaginal delivery^D^6 \[15--20\]532532RR1.02 \[0.86, 1.20\]**0.0359%**0.84**Unstable** ^V^assisted vaginal delivery^D^4 \[16,18--20\]443445RR1.08 \[0.84, 1.41\]0.4800.54stableanalgesia usage4 \[17--20\]405403RR1.07 \[0.99, 1.16\]0.7200.10stablematernal adverse events maternal infection5 \[17--21\]450453RR1.04 \[0.66, 1.66\]0.385%0.85stable postpartum haemorrhage3 \[15,18--19\]246249RR1.03 \[0.74, 1.42\]0.7300.88stableneonatal adverse events low Apgar score (\< 7 at 5 min)^E^3 \[17--18,20\]305303RR0.53 \[0.15, 1.88\]0.4600.32-- NICU admission3 \[17--18,20\]305303RR0.70 \[0.45, 1.070.9000.10stablelength of hospitalisation2 \[19--20\]248245MD0.16 \[−0.10, 0.41\]0.2815%0.22--satisfaction^F^ pain during the process2 \[15,19\]139139MD0.07 \[−0.53, 0.67\]0.4200.82-- maternal total satisfaction2 \[15,19\]139139MD−0.10 \[−1.25, 1.04\]**0.0868%**0.86--A: No studies, except for those of Hoppe 2015 \[[@CR17]\] and Salim 2011 \[[@CR20]\], specifically defined delivery as being either total delivery or vaginal delivery. Hoppe 2015 \[[@CR17]\] offered data on the time from insertion to vaginal delivery, while Salim 2011 \[[@CR20]\] reported both measurements. We synthesised these data by involving Hoppe 2015 \[[@CR17]\]. The upper and lower data shows the effects when we added the total and vaginal delivery data from Salim 2011 \[[@CR20]\]B: Salim 2011 \[[@CR20]\] excluded 124 women (70 in the double-balloon catheter group and 54 in the single-balloon catheter group) with spontaneous expulsion during this processC: We depended primarily on the Bishop score increment. For those studies that included only a second Bishop score, we included these data and conducted sensitivity analysesD: Hoppe 2015 \[[@CR17]\] reported only vaginal deliveries but did not define whether assisted vaginal deliveries were included; we treated these data as though it did not include assisted vaginal deliveriesE: Salim 2011 \[[@CR20]\] reported no events on this outcome for either arm, which was inestimableF: All measured by VASI: When we eliminated Salim 2011 \[[@CR20]\], the MD pooled effect changed to 2.16 \[0.76, 3.57\] (*p*-value, 0.003), in favour of the single-balloon catheter. The results remained comparable after all other sensitivity analyses were performedII: Excluding Rab 2014 \[[@CR19]\], though heterogeneity disappeared, the effect remained comparable (Q *p*-value, 0.17; I^2^, 38%; *p*-value, 0.58). Excluding Salim 2011 \[[@CR20]\], the result was shown in superscript note IIII: Significant heterogeneity existed regardless of which study we excluded; however, when we repeated the analysis after excluding Salim 2011 \[[@CR20]\], the result changed (MD, 2.40 \[0.32, 4.48\]; supporting the single-balloon catheter)IV: Stable effect but became homogeneous only when we excluded Hoppe 2015 \[[@CR17]\]V: Stable effect but became homogeneous only when we excluded Rab 2014 \[[@CR19]\] or Salim, 2011 \[[@CR20]\]

Subgroup analysis results by parity are shown in Tables [5](#Tab5){ref-type="table"} and Table [6](#Tab6){ref-type="table"}. Only the Bishop score increment in nulliparous women exhibited a statistically significant difference; however, heterogeneity was demonstrated among studies (MD, 1.08 \[0.38, 1.78\]; Q *p*-value, 0.11; I^2^, 56%; *p*-value, 0.002), suggesting that the double-balloon catheter may have a greater ability to increase the Bishop score. Unless otherwise **highlighted**, studies were homogeneous, and sensitivity analysis displayed no meaningful changes. Table 5Outcomes by parity (nulliparous)Outcomes\*Number of studies analysedInterventionsEffect measurePooled effect\
(95% CI)Q *p*-valueI^2^- statistic*p*-valueSensitivity analysisDoubleSinglecaesarean delivery5 \[15,17-18,20--21\]294301RR0.86 \[0.56, 1.33\]**0.0265%**0.50stableplacement difficulty/failure2 \[15,18\]146149RR0.72 \[0.15, 3.55\]0.5700.69--spontaneous expulsion1 \[15\]3939RR0.88 \[0.70, 1.10\]----0.27--insertion to delivery interval^A^4 \[15,17-18,20\]249251MD0.88 \[− 0.43, 2.18\]0.5900.19stable0.43 \[−0.84, 1.71\]0.2626%0.50stableinsertion to expulsion/removal interval2 \[15,17\]6464MD0.88 \[−0.00, 1.76\]0.3800.05--expulsion to delivery interval1 \[21\]4550MD−8.00 \[−16.35, 0.35\]----0.06--Bishop score increment^C^3 \[15,17,21\]109114MD1.08 \[0.38, 1.78\]0.11**56%0.002Unstable** ^**I**^vaginal delivery within 24 h3 \[17--18,20\]210212RR0.91 \[0.75, 1.10\]0.1940%0.33stablenormal vaginal delivery^D^4 \[15,17-18,20\]249251RR1.00 \[0.78, 1.29\]0.17**58%**0.98**Unstable** ^**II**^assisted vaginal delivery^D^2 \[18,20\]185187RR1.02 \[0.65, 1.59\]0.5600.94--analgesia usage2 \[18,20\]185187RR1.06 \[0.95, 1.19\]0.321%0.28--maternal adverse events maternal infection3 \[18,20--21\]230237RR1.16 \[0.69, 1.95\]0.2724%0.58stable postpartum haemorrhage2 \[15,18\]146149RR1.00 \[0.72, 1.40\]0.4900.98--neonatal adverse events low Apgar score (\< 7 at 5 min)^E^2 \[18,20\]185187RR0.21 \[0.01, 4.23\]----0.31-- NICU admission1 \[18\]107110RR0.74 \[0.45, 1.22\]----0.24--satisfaction^F^ pain during the process1 \[15\]3939MD−0.40 \[−1.69, 0.89\]----0.54-- maternal total satisfaction1 \[15\]3939MD0.47 \[−0.42, 1.36\]----0.30--\* Superscript notes (A-F) are the same as those for Table [4](#Tab4){ref-type="table"}I: Excluding Hoppe 2015 \[[@CR17]\], although heterogeneity disappeared, the effect remained (Q *p*-value, 0.46; I^2^, 0; *p*-value, 0.002). Excluding Solt 2009 \[[@CR21]\], heterogeneity still existed, but the effect significance disappeared (Q *p*-value, 0.03; I^2^, 78%; *p*-value, 0.07)II: Results remained comparable, but heterogeneity disappeared after we excluded Hoppe 2015 \[[@CR17]\] (Q *p*-value, 0.27; I^2^, 24%; *p*-value, 0.21) Table 6Outcomes by parity (multiparous)Outcomes\*Number of studies analysedInterventionsEffect measurePooled effect (95% CI)Q *p*-valueI^2^- statistic*p*-valueSensitivity analysisDoubleSinglecaesarean delivery2 \[17,20\]9591RR1.35 \[0.54, 3.36\]0.4300.52--insertion to delivery interval^A^2 \[17,20\]9591MD−0.48 \[−2.24, 1.28\]0.6200.60--−1.05 \[− 2.68, 0.58\]0.7000.21--insertion to expulsion/removal interval1 \[17\]2523MD2.30 \[−0.97, 5.57\]----0.17--Bishop score increment^C^1 \[17\]2523MD0.20 \[−0.88, 1.28\]----0.72--vaginal delivery within 24 h2 \[17,20\]9591RR1.05 \[0.93, 1.18\]0.2136%0.46--normal vaginal delivery^D^2 \[17,20\]9591RR0.92 \[0.83, 1.02\]0.3600.13--assisted vaginal delivery^D^1 \[20\]7068RR8.75 \[0.48, 159.42\]----0.14--analgesia usage1 \[20\]7068RR1.20 \[0.70, 2.07\]----0.51--maternal adverse events (infection)1 \[20\]7068RR2.92 \[0.12, 70.35\]----0.51--neonatal adverse events (low Apgar score)^E^1 \[20\]7068RRNot estimable^§^--------\* Superscript notes (A-F) are the same as those for Table [4](#Tab4){ref-type="table"}§: Reporting no events in both groups, which were inestimable

Discussion {#Sec11}
==========

Summary of main results {#Sec12}
-----------------------

### Efficacy and efficiency {#Sec13}

Balloon catheters were initially designed for cervical dilatation and ripening during labour induction. The best indicator of efficacy is the Bishop score increment. However, when correlated with baseline data, the Bishop score served only as a secondary outcome. No significant differences were observed for obstetric characteristics (including the Bishop score before catheter insertion) between women treated with the single-balloon catheter and those treated with the double-balloon catheter. Therefore, we could use the Bishop score after catheter removal (the second Bishop score) to roughly calculate this effect size, and it was not necessary to perform covariance analyses to adjust the baseline data. According to our analysis, the double-balloon catheter increases the Bishop score more significantly, especially for nulliparous women. However, this result was not observed for the multiparous subgroup. In support of this finding, one study \[[@CR17]\] reported a Bishop score \> 6 at balloon removal, and a similar trend in was observed for both general and subgroup subjects. Additionally, the ripening success rates (defined by the individual articles) appeared to be higher in the double-balloon groups, but without enough statistical power to determine significance \[[@CR16], [@CR19], [@CR22], [@CR23]\]. Atad et al. also reported similarly large average increments in the Bishop scores for both nulliparous and multiparous women for the double-balloon catheter, without a single-balloon catheter comparison group \[[@CR11]\]. Later, the researchers reported that the Bishop score increment when employing the single-balloon catheter was lower than that achieved by the double-balloon catheter, with a higher failure rate \[[@CR26]\].

Efficiency, best evaluated by the interval length and the 24 h delivery rate, is comparable regardless of parity. In the double-balloon catheter group, the interval from insertion to delivery appears to be longer, while the interval from expulsion to delivery appears to be shorter, though neither measure achieves significance. Ahmed, et al. \[[@CR15]\] stated that women treated with a single-balloon catheter had a shorter insertion to amniotomy time (*p* = 0.02) than women treated with a double-balloon catheter, while Pennell, et al. \[[@CR18]\] found that the length of labour did not significantly differ (*p* = 0.152) between the two groups; there is little consensus on the time from insertion to active labour, with Pennell, et al. \[[@CR18]\] preferring the single-balloon catheter (*p* = 0.014), while Rab, et al., \[[@CR19]\] demonstrated no obvious differences. Ahmed and Mei-Dan \[[@CR15], [@CR22]\] suggested that the shorter interval between insertion and expulsion for the single-balloon catheter likely resulted in the observed shorter induction to delivery interval, although the second Bishop score was lower in this group.

The frequency of placement difficulty or failure and spontaneous expulsion are similar between the two groups. In addition, Salim, et al. \[[@CR20]\] found that women who spontaneously expelled their catheter demonstrated favourable outcomes with regards to shorter times from induction to delivery (1.10 \[1.06--1.15\]; *p* = 0.001) and a significantly lower proportion of operative deliveries (2.15 \[1.26--3.69\]; *p* = 0.003).

### Safety {#Sec14}

Both maternal and neonatal adverse events are of great concern. Although we hoped to consider mortality data, no study provided this information. Other measurements were also equivalent, including maternal infection, postpartum haemorrhage, low Apgar scores and NICU admissions. Some studies also \[[@CR18], [@CR20]\] reported placental abruptions, uterine hyperstimulation, cord prolapse, malpresentation, and Apgar \< 4 at 1 min, with no significant differences between groups.

### Satisfaction {#Sec15}

Patient-reported outcomes (PROs), such as maternal satisfaction, represent what is most important to patients about a condition and its treatment \[[@CR24]\]. However, few reports related to PROs were found. Here, we can report patient satisfaction based on two original reports \[[@CR15], [@CR19]\], both evaluated by the visual analogue scale (VAS) \[[@CR27]\], with identical measurement times and protocols. The pooled results of these two studies suggests similar satisfaction levels for the two catheter types.

### Comprehensive outcomes {#Sec16}

Delivery modes, which are of particular clinical concern, represent a comprehensive measurement of the effectiveness and safety of labour induction protocols and can incorporate economic evidence. Caesarean section delivery is the most frequently used outcome pre-specified by trials. According to our analysis, no strong evidence exists to demonstrate which mechanical device is more effective, and heterogeneity exists among studies. Similarly, both normal and assisted vaginal delivery rates were comparable between groups, regardless of parity, as were the rates of analgesia usage during the ripening process and the lengths of hospitalisation.

### Heterogeneity {#Sec17}

Heterogeneity exists in many results, which may be the result of differences in study design or quality, participants, interventions, demographic feature or local policies. During our heterogeneity test, three studies \[[@CR17], [@CR19], [@CR20]\] were potential candidates for being the sources of heterogeneity. Unlike other studies, Rab et al. \[[@CR19]\] enrolled women who had experienced a stillbirth and had scarred uteri, which could be responsible increasing the general heterogeneity. Additional differences among these studies involved parity and balloon volumes (discussed below).

Applicability of evidence {#Sec18}
-------------------------

### Guide clinical Practice {#Sec19}

Despite the fact that the double-balloon makes results in more favourable Bishop scores, it appears to result in prolonged intervals. No differences were observed for delivery mode, which is the most meaningful obstetrical outcome. As for the economical consideration, it is mainly related to hospitalization length, delivery mode and device itself. What is noteworthy is that the single balloon (Foley catheter) is approximately 30--40 times cheaper than the double-balloon catheter at different institutions, and the difference of price varies from countries to countries. As the producers offered that Foley catheters cost approximately \$1.12, while Cook catheters cost approximately \$39.33. As for our hospital in China, the single balloon catheter costs 20--30 RMB while the double one costs 600 RMB, and the price for placing balloon catheters is about 600 RMB in both situations. Considering the fact that caesarean section and hospitalization length were similar in the two groups, and when coupled with a substantial price differences in the devices, the single balloon catheter seems like to be more cost-effective for labour induction, particularly in low resource settings.

### Exploring the mechanisms {#Sec20}

Practically, in our hospital, we prefer to place a balloon catheter at night, avoiding expulsion due to daily activity. Thus far, no studies have focused on this issue as a potential mechanism for labour induction. Theoretically, the insertion of a foreign object could increase the risk of intrauterine infections; however, the limited data from our analysis and previous studies did not show any evidence that the cervical ripening balloon catheter contributes to increased infection occurrences \[[@CR6], [@CR18], [@CR20], [@CR22], [@CR28]--[@CR31]\]. More studies are required to address the effects of the balloon-catheters on the rupturing of membranes and infection. In addition, physiologic differences in the mechanism through which balloon catheters induce labour according to parity also must be assessed.

Prior research demonstrated that a Bishop score \> 5 was associated with a greater likelihood of vaginal delivery \[[@CR32], [@CR33]\]. Although a higher Bishop score was achieved in the double-balloon group in our analysis, there were no differences in the vaginal delivery rates between the two groups. This result interested us, and we hypothesise that there may exist a threshold for the Bishop score that, once achieved, no further effects will be generated; after this threshold is met, the level of hormone secretions takes precedence over cervical conditions. Similar what is observed in our practice, favourable outcomes are rarely observed with balloon usage alone, unless augmentations (e.g., prostaglandin or oxytocin) are utilised.

The larger volume, the application of pressure on two sides (harder expulsion), and the ability to abandon traction when using the double-balloon catheter may explain the observed outcomes. The larger volume balloon may increase the separation between the amniotic membranes and the uterine decidua, resulting in an increase in the local secretion of prostaglandins and enhancing the cervical ripening process. Though 60 ml and 80 ml Foley catheters are more effective than 30 ml catheters \[[@CR34]--[@CR36]\], 80 ml + 80 ml Atad or COOK balloons do not demonstrate superiority to smaller Foley catheters, which may be due to other factors (e.g., traction). We hypothesise that traction may have a greater effect on the induction of labour and that the one-sided application of pressure may interfere with the labour pattern less than two-sided pressure. In theory, traction may cause discomfort for patients. However, this finding has not been confirmed by our analysis. Instead, speculum application prior to catheter insertion, which followed the same procedure in both groups, appears to be the main source of discomfort \[[@CR15]\].

Further studies are required to investigate the possible biological mechanisms on cervical ripening and the sources of discomfort, to provide practice guidelines and instrument improvement.

### Identifying the optimal methods for various populations {#Sec21}

Although there were no restrictions on settings, demographics or obstetrics characteristics, the participants from all of the included studies, except for Rab \[[@CR19]\], were women with viable singletons and without scarred uteri, making the applicability of our evidence limited. Vaginal birth after caesarean delivery (VBAC) has received increasing attention \[[@CR37]\], but identifying the optimal method for labour induction in this specific population remains controversial. Pharmacological methods are often rejected in VBAC women because of greater risks of complications. However, whether balloon catheters can and should be utilised in women with scarred uteri, which manufacturers do not recommend, requires further studies. In addition, twins and other multiple pregnancies are contraindications for the use of balloon catheters, despite the increased frequency of multiple pregnancies. Whether balloon catheters can be used in situations with multiple pregnancies also deserves further study.

Strengths and limitations {#Sec22}
-------------------------

In the current meta-analysis, no demographic or obstetric characteristics were restricted, which increases the applicability of the evidence. We performed evaluations examining evidence of bias and applied quality grades strictly based on the original reports and the Cochrane handbook. The 7 included trials are all rigorous in design, enabling the appraisal and interpretation of their results. Additionally, because bias is more important for studies with subjective events and positive results than for studies with negative results and objective outcomes, such as our analysis, it was acceptable to assume that bias would not practically undermine the results of our analysis.

When extracting data, some outcomes with various forms required data conversions, which likely led to analytical bias. Although we conducted sensitivity analysis specifically to test this possibility, it cannot be clearly determined whether these conversions influenced our outcomes. In addition, the outcomes we chose for this analysis are widely used in practice to avoid potential inconsistencies, and appropriate subgroup analyses were performed to identify potential sources of heterogeneity; however, heterogeneity remained too comprehensive to analyse fully.

The sample size of the current analysis had adequate power for the evaluation of the primary outcome. For some secondary outcomes, fewer data points were available, which may result in insufficient power and higher risks of publication bias. To minimise this bias and to involve more relevant studies, we have done our best to search databases using a wide range of publication years, to consider potentially eligible reviews and to fully utilise trial registration databases, with sensitivity-maximising search filters. Unfortunately, we are still incapable of accessing conference abstracts or proceedings and grey literature. Thus, publication bias cannot be excluded completely, and caution should be taken.

The procedures performed during our analysis to reduce bias and assess risks can provide direction for further research, although not all of these are necessary.

Conclusions {#Sec23}
===========

Both kinds of balloon catheter perform similarly with regards to efficacy, efficiency, safety and patient satisfaction. The single-balloon device appears to be more economical and practical, particularly in low resource settings.

CCRB

:   Cook Cervical Ripening Balloon

CDSR

:   Cochrane Database of Systematic Reviews

CENTRAL

:   Cochrane Central Register of Controlled Trials

CIs

:   Confidence intervals

MDs

:   Mean differences

PICOS

:   Population, intervention, comparison, outcomes and study designs

PROs

:   Patient-reported outcomes

RCTs

:   Randomised controlled trials

RRs

:   Relative risks

USFDA

:   United States Food and Drug Administration

VAS

:   Visual analogue scale

VBAC

:   Vaginal birth after caesarean delivery

**Publisher's Note**

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Xiyao Liu and Yu Wang contributed equally to this work.

Not applicable.

HBQ, XL and XYL contributed to the protocol design. The literature search was conducted by XYL and YW, with the help of RO. Study screening and data extraction were independently performed by XYL and YW. FZ contributed to resolving disagreements during these processes. All statistical analyses were completed with the aid of XNZ. XYL and YW reviewed and prepared the manuscript. HBQ and XL are guarantors of this manuscript. All authors have read and approved the final version of the manuscript.

The National Natural Science Foundation of China (No. 81771614 and No. 81771613).

The National Key Research and Development Program of China (No.2016YFC1000407).

The funders had no involvement in the study design, data collection and analysis, interpretation of data and preparation of the manuscript.

All data generated or analysed during the current study are available from the corresponding author on reasonable request.

Not applicable.

Not applicable.

We declare that none of the authors have any financial relationships or personal relationships with people, pharmaceutical companies or other commercial entities that have an interest in the subject matter or materials discussed in the article.
