Date Published: October 16, 2018
Publisher: Public Library of Science
Author(s): E. Vonesh, K. L. Gooch, V. Khangulov, C. R. Schermer, K. M. Johnston, S. M. Szabo, J. S. Rumsfeld, Antonia Vlahou.
For managing overactive bladder (OAB), mirabegron, a β3 adrenergic receptor agonist, is typically used as second-line pharmacotherapy after antimuscarinics. Therefore, patients initiating treatment with mirabegron and antimuscarinics may differ, potentially impacting associated clinical outcomes. When using observational data to evaluate real-world safety and effectiveness of OAB treatments, residual bias due to unmeasured confounding and/or confounding by indication are important considerations. Falsification analysis, in which clinically irrelevant endpoints are tested as a reference, can be used to assess residual bias. The objective in this study was to compare baseline cardiovascular risk among OAB patients by treatment, and assess the presence of residual bias via falsification analysis of OAB patients treated with mirabegron or antimuscarinics, to determine whether clinically relevant comparisons across groups would be feasible. Linked electronic health record and claims data (Optum/Humedica) for OAB patients in the United States from 2011–2015 were available, with index defined as first date of OAB treatment during this period. Unadjusted characteristics were compared across groups at index and propensity-matching conducted. Falsification endpoints (hepatitis C, shingles, community-acquired pneumonia) were compared between groups using odds ratios (ORs) and 95% confidence intervals (CI). The study identified 10,311 antimuscarinic- and 408 mirabegron-treated patients. Mirabegron patients were predominantly older males, with more comorbidities. The analytic sample included 1,188 antimuscarinic patients propensity-matched to 396 mirabegron patients; after matching, no significant baseline differences remained. Estimates of falsification ORs were 0.7 (CI:0.3–1.7) for shingles, 1.5 (CI:0.3–8.2) for hepatitis C, 0.8 (CI:0.4–1.8) and 0.9 (CI:0.6–1.4) for pneumonia. While propensity matching successfully balanced observed covariates, wide CIs prevented definitive conclusions regarding residual bias. Accordingly, further observational comparisons by treatment group were not pursued. In real-world analysis, bias-detection methods could not confirm that differences in cardiovascular risk in patients receiving mirabegron versus antimuscarinics were fully adjusted for, precluding clinically relevant comparisons across treatment groups.
Overactive bladder (OAB) is characterized by urge urinary incontinence and urgency, nocturia, and high urinary frequency. The prevalence of OAB has been estimated to be 11.8% in the United States (US), with higher rates in older individuals. While behavioral modifications including bladder training, pelvic floor training, and limitation of fluids are intended as the first line of treatment for OAB, pharmacologic intervention is a mainstay of OAB management. To date, antimuscarinic therapies–including oxybutynin, tolterodine, solifenacin, flavoxate, fesoterodine, trospium, or darifenacin–have been the most common first-line pharmacologic treatment for OAB. Mirabegron (Myrbetriq/Betmiga; Astellas Pharma) is a β3 adrenergic receptor agonist with demonstrated efficacy and safety in managing the symptoms of OAB. In clinical practice, mirabegron is typically given as second-line pharmacotherapy after discontinuation or failure of therapy with antimuscarinics. While clinical trials have shown mirabegron to be both efficacious and safe, in two randomized, placebo-controlled studies of healthy volunteers, mirabegron was associated with dose-related increases in supine blood pressure with the currently marketed and maximum recommended dose of 50 mg. The mean increase in systolic blood pressure (SBP)/diastolic blood pressure (DBP) was approximately 3.5/1.5 mm Hg greater than placebo. In three, phase 3, 12-week, double-blind, placebo-controlled, safety and efficacy studies of OAB patients receiving mirabegron 25 mg, 50 mg or 100 mg once daily, mean increases in SBP/DBP of approximately 0.5–1.0 mm Hg were observed compared to placebo. Both SBP and DBP increases were reversible upon discontinuation of treatment.
To derive the study sample, 34,243 individuals (1,417 ever treated with mirabegron and 32,836 ever treated with an antimuscarinic) were initially identified for potential inclusion (Fig 1), based on a diagnosis of OAB and a billing record for a dispensed prescription at any point in time (e.g. not necessarily during the study period). It is of interest to note that more than twice as many patients had a record of a written prescription for mirabegron or an antimuscarinic in the EHR (data not shown) without a record of a prescription being dispensed in the claims data, perhaps indicating concerns with primary adherence and potential for bias in inducing differences across treatment groups. More than half of antimuscarinic patients (n = 17,426) and approximately one third of mirabegron patients (n = 847) were excluded because they were not dispensed a prescription during the study follow-up period (i.e. while these individuals received the medication of interest at some point during data coverage, they did not have a filled prescription during the study period). Of the resulting 15,980 patients (15,410 who received an antimuscarinic during the study period and 570 receiving mirabegron), most had 12 months’ continuous data available prior to the index date in at least one of the EHR or claims databases, and were at least 18 years of age, with only a small number of exclusions related to these criteria (n = 16 exclusions for mirabegron and n = 740 exclusions for antimuscarinics). The requirement of a blood pressure measure being available within 90 days of index date resulted in 137 exclusions in the mirabegron arm and 4,163 exclusions in the antimuscarinic arm. A small number of exclusions were made due to recent pregnancy and/or cardiovascular events. After applying all inclusion and exclusion criteria, the final sample was reduced to 408 mirabegron patients and 10,311 antimuscarinic patients. The antimuscarinic group was then further reduced to create a 3:1 propensity-matched sample to mirabegron. After 3:1 propensity matching, the final sample size was 396 in the mirabegron group and 1,188 in the antimuscarinic group. Thus, of the 15,980 OAB patients who received a prescription during the study period, approximately ten percent were eligible for the final analytic study population.
This study aimed to characterize the cardiovascular risk profile of untreated, mirabegron-treated, and antimuscarinic-treated OAB patients, using an integrated claims/EHR dataset. An incremental analytical approach was implemented to ensure rigor and accuracy within the constraints of observational data. In a retrospective cohort of individuals with OAB, mirabegron patients were found to be older, with more comorbidities and more prior cardiovascular events relative to antimuscarinic patients. Mirabegron is typically prescribed as a second-line agent to antimuscarinics, due to either inadequate response or poor tolerability of antimuscarinics, or due to formulary rules e.g. stepped therapy conditions. As such, unadjusted comparisons of treatment effectiveness or safety between mirabegron and antimuscarinics based on these data may be biased by differences in the distribution of baseline risk factors between treatment cohorts. Indeed, despite what appeared to be adequate propensity score matching, when residual bias was assessed through falsification analyses, the resulting confidence intervals were sufficiently wide that associations with falsification outcomes (as evidenced by wide intervals) could not be ruled out. The small sample size ultimately eligible for study inclusion–due to the significant attrition caused by lack of overlap of subjects between the claims and EHR datasets, and application of the relatively limited exclusion criteria–contributed to limited power and to confidence interval width, preventing a definitive interpretation of results and conclusions.
Using a repeated measures analysis of variance (RM-ANOVA) carried out over 4 time points (0, 3, 6, and 12 months), sample sizes needed to detect clinically relevant differences in SBP/DBP were calculated on the basis of a global hypothesis that tests for equality between the two treatment groups at every time point. This global test is appropriate whenever there are a wide variety of hypotheses of interest, including, for example, treatment differences at one or more time points, or comparisons between the change scores from one set of time points to another, etc. In this test, protection against a Type I error is maintained at the prescribed alpha-level for all such comparisons. To err on the conservative side, the Type I error for comparing mean change from baseline at 3, 6 and 12 months for the co-primary endpoints of change in systolic and diastolic blood pressures was initially set at 0.05/3 = 0.0167 so as to yield an overall Bonferroni-adjusted Type I error of 0.05. Power was set at 0.90. Under conservative assumptions regarding correlation and assuming 1:1 matching of patients, sample size calculations showed that detecting a systolic blood pressure difference of 2.5 mmHg requires 1,117 individuals in each treatment group assuming a 14 mm Hg standard deviation, and 2,277 individuals in each treatment group assuming a 20 mm Hg standard deviation. However, based on a lower than anticipated number of eligible mirabegron patients relative to antimuscarinic patients, revised sample size calculations were made based on a 3:1 matching of antimuscarinic to mirabegron patients with the Type I error set at 0.05. At power of 0.80, a sample of 500 mirabegron patients and 1500 antimuscarinic patients would be required to detect a systolic blood pressure difference of 2.5 mmHg assuming a 14 mmHg standard deviation. When based on a power of 0.90, the sample size increased to 645 mirabegron patients and 1935 antimuscarinic patients.