Factors predisposing to perinatal death related to uterine rupture during attempted vaginal birth after caesarean section: retrospective cohort study

Introduction

Uterine rupture during attempted vaginal birth after a previous caesarean section is a rare event that affects about one in 200 women,1 and consequent death of the infant is even rarer, affecting about one per 2000.2 Analysis of the factors determining perinatal death due to uterine rupture therefore requires data from large numbers of women. Most large scale databases of births lack detailed information on the cause of perinatal death and the obstetric characteristics of the population. Consequently, we know of no reports of the risk factors for perinatal death due to uterine rupture during attempted vaginal birth after previous caesarean section. We linked national registries of pregnancy discharge data and perinatal death to determine the factors associated with this event.

Methods

The inclusion criteria for the study were that a woman had previously had one caesarean section and was delivered in her current pregnancy by a means other than planned caesarean section. We excluded women with more than one previous caesarean, those with multiple gestations, those delivering before 37 weeks' gestation, and those delivering beyond 43 weeks' gestation. We also excluded cases in which the infant died from causes other than intrapartum uterine rupture.

Data sources—The Scottish Morbidity Record (SMR2) collects information on clinical and demographic characteristics and outcomes for all patients discharged from Scottish maternity hospitals. The register is subjected to regular quality assurance checks and has been > 99% complete since the late 1970s.3 The register collects both specific obstetric data and up to six ICD-9 (international classification of diseases, ninth revision) or ICD-10 (10th revision) diagnostic codes relating to the admission. In 1996-7 a quality assurance analysis comparing 1414 records with the clinical notes showed that the register was free from significant errors in > 98% of records in all the specific fields used in the present analysis. Exceptions were postcode (94.0%), height (96.2%), estimated gestation (94.4%), and method of induction of labour (93.6%). The previous caesarean section field was 99.7% accurate. ICD diagnostic codes were found to be 80-90% accurate for the first four diagnoses and 70-80% accurate for the remainder (Jim Chalmers, personal communication). SMR2 records were linked to records from the Scottish Stillbirth and Infant Death Survey. This national register routinely classifies all perinatal deaths in Scotland. Coding of the cause of death is performed by a single medically qualified individual, and the survey is described in detail elsewhere.4

Definitions—We defined trial of labour as a singleton delivery at term by a means other than planned caesarean section in women with only one previous caesarean delivery. The definition of perinatal death due to uterine rupture was that the obstetric cause of death was coded as “mechanical” under the modified Wigglesworth classification5 and that the ICD-9 diagnostic code for intrapartum uterine rupture (665.1) was listed under the specific diagnoses. Intrapartum uterine ruptures that did not result in perinatal death were identified with ICD-9 and ICD-10 diagnostic codes 665.1 and O711, respectively, from the diagnostic fields in the SMR2 record related to hospital discharge after delivery. Hospital throughput was defined as the total number of births recorded in the SMR2 database for the given hospital over the given year. Hospital throughput was categorised into above or below the median (3000 births). Other maternal characteristics were defined as previously described.2

Statistical analyses—We summarised continuous variables with medians and interquartile ranges and used the Mann-Whitney U test for comparisons between groups and Fisher's exact test for univariate comparisons of dichotomous data. P values for all hypothesis tests were two sided. Multivariate analysis was performed using logistic regression analysis. The significance of interaction terms was assessed with the likelihood ratio test. The goodness of fit of models was assessed with the Hosmer and Lemeshow test based on tenths of probability.6 Because of the rarity of the event we used exact logistic regression to model the risk of perinatal death due to uterine rupture.7 When we treated annual number of births as a continuous variable in the exact model, we rounded it to the nearest 50 to make the model computationally feasible. All statistical analyses were performed with the Stata software package version 8.2 (StataCorp, College Station, TX), except for exact logistic regression, which was performed with LogXact version 5.0 (Cytel Software Corporation, Cambridge, MA).

Results

There were 871 283 SMR2 birth records for Scotland for 1985-98. In total 39 729 (4.6%) women had had one previous caesarean delivery and were delivered by a means other than planned caesarean section. There were 452 (1.1%) multiple births, 3462 (8.7%) births outside the range of 37-43 weeks' gestation, and 543 (1.4%) perinatal deaths due to causes other than intrapartum uterine rupture. We excluded a further 32 (< 0.1%) records because the women were documented as being primigravid, despite having had a previous caesarean delivery. We therefore excluded 3875 (9.8%) records (some cases were excluded in more than one category), leaving a study group of 35 854. We compared the demographic and obstetric characteristics of the study group according to whether there was a perinatal death due to uterine rupture (table 1).

There was no association between the annual number of deliveries and the risk of emergency caesarean delivery (odds ratio 1.00, 95% confidence interval 0.98 to 1.01, P = 0.58) or uterine rupture overall (0.98, 0.86 to 1.11, P = 0.70), but there was a significant negative association with the risk of perinatal death due to uterine rupture (0.68, 0.46 to 0.99, P = 0.04) (figure).

Proportions of emergency caesarean section, all uterine rupture, and perinatal death due to uterine rupture, in relation to size of hospital

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On univariate analysis, with the 35 854 women who attempted vaginal birth as the denominator, the risk of uterine rupture was higher in women who had not previously given birth vaginally and in women who had been induced with prostaglandin but not with other methods of induction (table 2). Though delivery in a hospital with < 3000 births a year was not associated with the risk of uterine rupture overall, the other associations remained highly significant in multivariate analysis and the point estimates were similar (table 3). There were enough uterine ruptures for us to test the goodness of fit of the model and to examine interactions between the variables. The goodness of fit was adequate (P > 0.05), and there were no significant first order interactions between any of the variables with each other or with the year of birth.

The risk of perinatal death due to uterine rupture was also higher in women who had not previously given birth vaginally and in women who had been induced with prostaglandins but not with other methods of induction (table 2). However, in addition, delivery in a hospital with < 3000 births a year was associated with a significantly increased risk of perinatal death due to uterine rupture. The risk of perinatal death was about one in 1300 in hospitals with < 3000 births a year and one in 4700 in hospitals with ≥ 3000 births a year. Because of the small number of deaths caused by uterine rupture, significance was generally attenuated in multivariate analysis (table 3). However, the point estimates were similar to those from the univariate analysis, indicating that the associations seen in univariate analysis were not due to confounding by the factors included in the model. There were too few events for us to assess goodness of fit or first order interactions.

Among the 124 cases of uterine rupture, there were 17 (13.7%) intrapartum stillbirths or neonatal deaths. There were 63 uterine ruptures in hospitals delivering < 3000 women per year and 13 (20.6%) resulted in perinatal death. In hospitals delivering ≥ 3000 women a year there were 61 uterine ruptures and four (6.6%) resulted in perinatal death (P=0.03). Among women with uterine rupture, the relative risk of perinatal death in a hospital with < 3000 births a year was about threefold (table 4).

When we confined the analysis to births ≥ 40 weeks' gestation, the risk of uterine rupture was significantly associated with no previous vaginal birth (odds ratio 2.0, 95% confidence interval 1.2 to 3.4, P = 0.009) and induction of labour with prostaglandin (2.2, 1.4 to 3.5, P = 0.001). Formal tests of interaction between each of these variables and gestation ≥ 40 weeks showed that the strength of the associations did not significantly differ before and after 40 weeks (P = 0.2 and 0.3, respectively). Among the 22 170 births ≥ 40 weeks' gestation, there were seven deaths out of 10 602 births in hospitals with < 3000 births a year and one death out of 11 568 births in hospitals with ≥ 3000 births a year (P=0.02).

Of the 12 633 women who had previously given birth vaginally, 1499 (11.8%) were induced with prostaglandin compared with 2976 of the 20 215 (12.8%) women who had not done so (P = 0.006). We used a logistic regression model to estimate the absolute risk of uterine rupture (including cases in which the infant survived and cases in which the infant died) in relation to different combinations of parity and induction of labour with prostaglandin in relation to 1998 rates. Among women who had not previously given birth vaginally, the risk of uterine rupture without induction of labour with prostaglandin was one in 210 and with induction of labour with prostaglandin was one in 71. Among women with a previous vaginal birth, the risk of uterine rupture without induction of labour with prostaglandin was one in 514 and with induction of labour with prostaglandin was one in 175.

Discussion

We found that after a previous caesarean section women who had not previously given birth vaginally and those who had labour induced with prostaglandin were at increased risk of uterine rupture. The same two factors were associated with the risk of perinatal death due to uterine rupture. In contrast, delivering in a hospital with low throughput was not associated with uterine rupture overall but was associated with an increased risk of perinatal death due to uterine rupture. We found that uterine rupture was three times more likely to result in death of the infant if the delivery took place in a hospital with < 3000 births a year. Confining trials of labour to larger obstetric units may therefore reduce the risk of perinatal death associated with uterine rupture during a trial of labour.

The finding that units with high throughput had lower rates of perinatal death due to uterine rupture is plausible. Hospitals with greater throughput are more likely to have resident obstetric, anaesthetic, and neonatal services as well as a dedicated obstetric operating theatre. These factors would allow a faster response to fetal distress due to uterine rupture, which in turn would allow more rapid delivery and resuscitation of the neonate. We did not have information on the structure of services at each unit over the period of study. However, the factors are likely to be highly correlated and interdependent. A unit with no resident obstetric or anaesthetic cover is unlikely to have a dedicated obstetric operating theatre or a resident experienced neonatologist. Therefore, even if these data were available, it would be extremely difficult to determine the independent contributions of each of these factors in reducing the risk of death. Therefore, the total number of births is a useful composite measure of the level of support and has the pragmatic advantage of being easy to define.