Short- and long-term clinical outcomes of use of beta-interferon or glatiramer acetate for people with clinically isolated syndrome: a systematic review of randomised controlled trials and network meta-analysis

The current work follows a systematic review (PROSPERO identifier: CRD42016043278) that was undertaken as part of a larger project assessing the clinical and cost effectiveness of IFN-β and GA acetate for MS within the UK [8]. The following electronic databases were searched for RCTs in CIS in February 2016: Cochrane Multiple Sclerosis and Rare Diseases of the CNS group specialised register; MEDLINE (Ovid); MEDLINE In-Process and Other Non-Indexed Citations (Ovid); Embase (Ovid); Cochrane Library (Wiley), including Cochrane Database of Systematic Reviews, Cochrane Central Register of Controlled Trials (CENTRAL), DARE, NHS EED, and HTA databases); Science Citation Index and Conference Proceedings—Science (Web of Science); UKCRN Portfolio Database; WHO ICTRP; Current Controlled Trials; ClinicalTrials.gov. No date restriction was applied. The included and excluded study lists from previous relevant Cochrane systematic reviews were also checked [9‐11]. Additionally, any CIS RCTs identified during the broader MS systematic review were also considered for inclusion. These additional searches are described in the report of this larger project [8] (Online resource 1). During our original work, we also identified long-term follow-up studies of immediate versus delayed treatment at title/abstract level, but excluded them at full-text review. For the purpose of the current review, we retrieved these records from our initial list of excluded references and undertook targeted searches using more than one source to find other potential subsequent reports from the original CIS studies.

Our original systematic review included RCTs conducted in people with RRMS, SPMS, or CIS, evaluating all forms of beta-interferon (IFN β-1a; pegylated IFN β-1a; IFN β-1b) and GA compared against each other or placebo/best supportive care, and where clinical outcomes were reported such as relapse rates, progression to multiple sclerosis, or disability progression as measured by the Expanded Disability Status Scale (EDSS). We consider here a subset of studies reporting RCTs conducted in people with a single clinical attack diagnosed as having CIS and where the interventions were used at the authorised dose regimen (IFN β-1b 250 µg subcutaneously (SC) every other day for Betaferon^® (Bayer, Leverkusen, Germany) and Extavia^® (Novartis, Basel, Switzerland); IFN β-1a 30 µg intramuscularly (IM) once weekly for Avonex^® (Biogen Idec Ltd, Cambridge, MA, USA); GA 20 mg SC daily for Copaxone^® (Teva Pharmaceutical Industries, Petah Tikva, Israel); IFN β-1a 44 µg SC three times weekly for Rebif^® (Merck, Darmstadt, Germany)).

Our main outcomes of interest were: (1) time to CDMS using Poser criteria [12] and involving a second relapse or neurological deterioration, measured using hazard ratios (HR); we also undertook a sensitivity analysis examining time to ‘McDonald MS’ instead of time to CDMS, in which MRI findings could be used with clinical findings to arrive at a diagnosis; (2) discontinuation due to AE, measured using risk ratios (RR). Although these were not included in our original systematic review, we also included long-term open-label extension studies following the double-blind period of RCTs that had enrolled patients with CIS. Open-label extension studies were defined as studies that compared all those randomised to active treatment who continued on treatment in extension phase (early DMT) versus all those randomised initially to placebo who then crossed over to active treatment in extension phase (delayed DMT). We used these studies to explore the impact of DMTs on disability progression, which was measured using the Expanded Disability Status Scale (EDSS). We also examined long-term annualised relapse rates (ARR) from open-label studies.

In our initial work, we first examined relevant past systematic reviews from the Cochrane MS group [7, 10, 11] for studies meeting the inclusion criteria. For updated and new searches, we collected all retrieved records in a specialised database and duplicate records were identified and removed. Subsequently, two reviewers applied the inclusion/exclusion criteria and screened all identified bibliographic records for title/abstract and then for full text. Any disagreements were resolved through discussion consensus or by recourse to a third-party reviewer.

Data extraction was performed by two reviewers and was cross-checked. Any disagreement was resolved through discussion or recourse to a third reviewer. We undertook the quality assessment of included studies using the Cochrane risk of Bias tool for randomised-controlled trials [13]. The quality assessment was cross-checked within the reviewer team. Any disagreements were resolved as in earlier stages. The included papers were organised in text and tables. The summary tables tabulated characteristics and results for included primary studies. Tables for primary studies present summary data on participants, interventions, and outcomes.

For RCTs reported up to the end of double-blind periods, we performed a conventional pairwise meta-analysis using a method-of-moment random effects model (inverse-variance weighted) to directly compare each pair of interventions for which direct evidence exists, and to compare included DMTs as a class against placebo. We examined these pairwise meta-analyses for heterogeneity, measured as Cochran’s Q and I². We then estimated random effects network meta-analyses in the frequentist framework, using package network in Stata v.14. For each outcome of interest, a network diagram of pairwise direct comparisons between multiple treatments was created. Tests for inconsistency were inapplicable because the network was star shaped. The summary estimates (with 95% CIs) for all pair-wise comparisons were provided in a league table [14]. We aimed to determine the rankings of all treatments (e.g. based on the relative probability of an intervention being superior) with respect to effectiveness and safety outcomes in reference to placebo/standard care and we used the surface under the cumulative ranking curve (SUCRA) [14].

For open-label extension studies, we calculated a pooled estimate of the effect of early DMT, including all DMTs as a class, compared to delayed DMT on time to CDMS (effect again measured as HR) through a method-of-moments random effects meta-analysis. When HR were reported at different follow-up durations, we used the most mature data, i.e. those with the longest follow-up. In these studies, the outcomes were only reported for patients originally randomised to active treatment or placebo, who entered extension phase. Our exploratory work measuring the impact of early DMT on disability progression and examining long-term ARR was summarised narratively.

Our original searches identified 6420 records. Of these, 6146 were excluded based on title and/or abstract, leaving 274 to be examined at full-text (Fig. 1).

Among these, 14 records [15‐28] (including two additional ones from our later targeted search) met our inclusion criteria and corresponded to five trials: BENEFIT comparing IFN β-1b 250 µg SC every other day to placebo [15]; CHAMPS [16] and Pakdaman et al. [17] both evaluating IFN β-1a 30 µg IM once a week against placebo; PRECISE comparing the effect of GA 20 mg SC once daily to placebo [18]; and REFLEX comparing IFN β-1a 44 µg SC three times a week against placebo [19]. In these trials, all included patients had a single clinical event and also had evidence of clinically silent lesions based on MRI. The double-blind period in these trials was 2 years after randomisation in BENEFIT and REFLEX, and 3 years in PRECISE, Pakdaman et al. [17], and CHAMPS (Table 1).

Nine of the 14 selected records were open-label extension studies of four of the five above-listed trials (Online resource 2): BENEFIT with long-term follow-up at 3, 5, 8, and 11 years [20‐23]; CHAMPS with results at 5 and 10 years (CHAMPIONS study) [24, 25]; PRECISE with a prolonged follow-up at 5 years [27]; and REFLEX with a 5-year follow-up (REFLEXION study) [26, 28]. We found no subsequent long-term follow-up for the Pakdaman et al. [17] RCT. In open-label extension studies, patients initially allocated to placebo were offered active treatment from the time of a second clinical attack or after the end of the double-blind period (delayed DMT group). Conversely, people initially allocated to active treatment were offered to continue on treatment during an extension study (early DMT group). Overall, the proportion of patients entering open-label extension studies ranged from 53 to 86% after completion of the double-blind phase with no significant differences between the two groups (early or delayed DMTs) within each study (Online resource 2).

We judged there was a substantial proportion of studies with insufficient information to rate the risk of selection bias (60% of studies rated as unclear risk on random sequence generation and 40% of studies rated as unclear risk on allocation concealment) (Online resource 3 and 4). Although studies were placebo controlled, we found that four of the five (80%) studies were at high risk of performance bias (due to loss of blinding of participant and personnel). This was because of the much increased rates of side effects such as injection-site reactions in patients allocated to active treatment which led to high risk of participant unblinding.

Overall, we found no major issue of detection (blinding of outcome assessment) or reporting bias across studies. Only one study [17] was rated as at high risk at attrition bias (incomplete outcome data). Understandably, the open-label extension studies were at much higher risk of attrition bias with a range of complete follow-up rates after entering the extension phase of 44–96%. Finally, all studies funded by drug manufacturers were designated as high risk of bias under the ‘other’ category.

Short-term clinical outcomes were informed by the results from the double-blind period of RCTs.