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Erschienen in: BMC Medicine 1/2021

Open Access 01.12.2021 | Research article

Comparative effectiveness and safety of pharmaceuticals assessed in observational studies compared with randomized controlled trials

verfasst von: Yoon Duk Hong, Jeroen P. Jansen, John Guerino, Marc L. Berger, William Crown, Wim G. Goettsch, C. Daniel Mullins, Richard J. Willke, Lucinda S. Orsini

Erschienen in: BMC Medicine | Ausgabe 1/2021

Abstract

Background

There have been ongoing efforts to understand when and how data from observational studies can be applied to clinical and regulatory decision making. The objective of this review was to assess the comparability of relative treatment effects of pharmaceuticals from observational studies and randomized controlled trials (RCTs).

Methods

We searched PubMed and Embase for systematic literature reviews published between January 1, 1990, and January 31, 2020, that reported relative treatment effects of pharmaceuticals from both observational studies and RCTs. We extracted pooled relative effect estimates from observational studies and RCTs for each outcome, intervention-comparator, or indication assessed in the reviews. We calculated the ratio of the relative effect estimate from observational studies over that from RCTs, along with the corresponding 95% confidence interval (CI) for each pair of pooled RCT and observational study estimates, and we evaluated the consistency in relative treatment effects.

Results

Thirty systematic reviews across 7 therapeutic areas were identified from the literature. We analyzed 74 pairs of pooled relative effect estimates from RCTs and observational studies from 29 reviews. There was no statistically significant difference (based on the 95% CI) in relative effect estimates between RCTs and observational studies in 79.7% of pairs. There was an extreme difference (ratio < 0.7 or > 1.43) in 43.2% of pairs, and, in 17.6% of pairs, there was a significant difference and the estimates pointed in opposite directions.

Conclusions

Overall, our review shows that while there is no significant difference in the relative risk ratios between the majority of RCTs and observational studies compared, there is significant variation in about 20% of comparisons. The source of this variation should be the subject of further inquiry to elucidate how much of the variation is due to differences in patient populations versus biased estimates arising from issues with study design or analytical/statistical methods.
Hinweise

Supplementary Information

The online version contains supplementary material available at https://​doi.​org/​10.​1186/​s12916-021-02176-1.

Publisher’s Note

Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.
Abkürzungen
CI
Confidence interval
DAPT
Dual-antiplatelet therapy
EHR
Electronic health record
FDA
Food and Drug Administration
HR
Hazard ratio
MAPT
Mono-antiplatelet therapy
OPERAND
Observational Patient Evidence for Regulatory Approval and Understanding Disease
OR
Odds ratio
RCT
Randomized controlled trial
RWD
Real-world data
RWE
Real-world evidence
RR
Relative risk

Background

Health care decision makers, particularly regulators but also health technology assessment agencies, have depended upon evidence from randomized clinical trials (RCTs) to assess drug effectiveness and to make comparisons among treatment options. Widespread adoption of the RCT was the hallmark of progress in clinical research in the twentieth century and accelerated the development and approval of new therapeutics; confidence in RCTs derived from their experimental nature, designs to minimize bias, rigorous data quality, and analytic approaches that supported causal inference.
In the last 30 years, we have witnessed an explosion of observational real-world data (RWD) and evidence (RWE) derived from RWD that has supplemented our understanding of the benefits and risks of treatments in broader populations of patients. RWE has been largely leveraged by regulators to assess the safety of marketed products and for new drug approvals when RCTs are infeasible, such as in rare diseases, oncology, or for long-term adverse effects. RCTs often do not have sufficient sample size to detect rare adverse events or long enough follow-up to detect long-term adverse effects. In such cases, regulatory decisions are often supplemented by RWE. However, leveraging of RWE has been much more slowly embraced in comparison to the adoption of RCTs for a variety of reasons. Imputation of causality is less certain in the absence of randomization and RWD can be much sparser and often requires extensive curation before it can be analyzed. Thus, skepticism about the robustness of observational RWD studies has made decision makers cautious in relying solely upon it to render judgments about the availability and appropriate use of new therapeutics, particularly by regulatory bodies.
Moreover, observational studies examining the effectiveness of treatments in similar populations have not always provided results consistent with RCTs. Despite many studies finding similar treatment effect estimates from RCTs and RWD analyses [13], other analyses have documented wide variation in results from RWD analyses within the same therapeutic areas [4], including analyses using propensity score-based methods [5]. Nonetheless, public interest has grown in the routine leveraging of RWD to promote the creation of a learning healthcare system, and regulatory bodies and other decision makers are exploring ways to expand their use of RWE. This is partly due to increasing acknowledgement of the value of RWE, such as its ability to better reflect actual environments in which the interventions are used.
One promising approach to understanding the sources of variability between RCT and observational study results is to compare estimates obtained from RWD analyses that attempt to emulate the eligibility criteria, endpoints, and other features of trials as closely as possible. A small number of RWD analyses have generated findings similar to previous RCTs [6, 7], and the findings of other RWD analyses have been consistent with subsequent RCTs [8]. In a small number of cases, RCTs and RWD studies have been published simultaneously [9]. This has the advantage of not knowing the RCT estimate when conducting the RWD study. There have been disagreements between observational RWD analyses and RCTs that were based upon avoidable errors in the RWD analysis design [7, 10]. This has led to a focus on the importance of research design in observational RWD analyses attempting to draw causal inferences regarding treatment effects [1113].Emulation studies can improve understanding of when observational studies may reliably generate results consistent with RCTs; however, not all RCTs can be feasibly emulated using RWD due to limitations in observational datasets. Existing sources of observational data, such as health insurance claims and electronic health records (EHRs), may not routinely capture the intervention, indication, inclusion and exclusion criteria, and/or endpoints used in RCTs [14].
The objective of this paper is to provide further evidence on the comparability of RCTs and observational studies when the latter use a range of study designs and were not designed to emulate RCTs. We aim to quantify the extent of the difference in treatment effect estimates between RCTs and observational studies. We go beyond previous comparisons of RCTs and observational studies, with a focus purely on pharmaceuticals, and provide a systematic landscape review of the (in)consistency between RCT and observational study treatment effect estimates. The reasons for the variation in relative treatment effects are not assessed in this review but should be the subject of further study.

Methods

Eligibility criteria

Inclusion criteria

  • Study design:
    • Published systematic literature reviews designed to compare relative treatment effects from observational studies with the corresponding effects from RCTs; or
    • Published systematic literature reviews that reported subgroup analyses stratified by RCT and observational study design; and
    • Observational studies included in these reviews have to be retrospective or prospective cohort studies, or case-control studies
  • Population: Human subjects
  • Intervention(s) and comparator(s): Any active or placebo-controlled pharmaceutical or biopharmaceutical intervention
  • Outcome(s):
    • Efficacy/effectiveness or safety outcomes
    • Pooled relative treatment effect estimates for both observational studies and RCTs

Exclusion criteria

  • Systematic reviews that compared absolute outcomes, such as event rates, between non-comparative observational studies and RCTs
  • Non-pharmaceutical-based studies, e.g., surgical procedures, traditional medicine, vitamin/herbal supplements, etc.
  • Non-English language
  • Abstracts or conference proceedings

Search strategy

We searched PubMed and Embase to identify relevant systematic literature reviews published between January 1, 1990, and January 31, 2020. Anglemeyer et al.’s search strategy [1] was used as a template to develop the search strategy, which included a wide range of MeSH terms and relevant keywords. We updated Anglemeyer et al.’s systematic review hedge and used the more recent CADTH systematic review/meta-analysis hedge, created in 2016, in both PubMed and Embase [15]. We restricted our search to focus on pharmaceuticals only. PubMed and Embase were searched for the following concepts: pharmaceuticals, study methodology, and comparisons (filters: Humans and English language). The PubMed search strategy which was adapted for use in Embase can be found in Additional File 1.

Study selection

After removing duplicate references, three authors (JG, YH and LO) screened the titles and abstracts to identify relevant reviews. Once complete, LO verified the screening for accuracy. Following the title and abstract screen, full text articles were obtained for all potentially relevant reviews. Full text articles were then assessed to determine if they meet the selection criteria for final inclusion in the review.

Data extraction

A pilot extraction was first done by two authors (JG and YH) on a sample of three articles using a standardized extraction table. This was done to test the standardized extraction table and to ensure consistency between the authors performing the data extraction. JG and YH then independently extracted information from each review using the standardized extraction table. A third author (LO) verified the extraction for accuracy and identified any discrepancies. These discrepancies were discussed until resolved.
We focused on primary outcomes reported in the reviews and extracted information summarizing the scope of each of the identified systematic reviews. Extracted information included the following: review objective, population, disease/therapeutic area, interventions, outcome(s), number of included RCTs and observational studies, pooled relative treatment effect estimates for RCTs and observational studies along with the 95% confidence intervals (95% CI), and measures of heterogeneity.

Analysis

Based on the extracted information, we calculated the ratio of the relative treatment effect estimate from observational studies over the relative treatment effect estimate from RCTs (e.g., RRobs/RRrct), along with the corresponding 95% CI obtained via a Monte Carlo simulation for each pair of pooled RCT and observational study estimates. Outcomes for which the relative treatment effect was not expressed with a relative risk (RR), odds ratio (OR), or hazard ratio (HR) were excluded from our analysis.
We expressed differences in pooled effect estimates with the following measures: ratios that were < 1, > 1, or = 1, ratios indicating an “extreme difference” (< 0.70 or > 1.43) [16] and absence of an extreme difference. We evaluated (in)consistency between pooled RCT and observational study estimates with the following measures: presence of opposite direction of effect, RCT effect estimate outside the 95% CI of the observational study estimate, and vice versa, statistically significant difference between RCT and observational study estimates, and statistically significant difference along with opposite direction of effect. Statistically significant difference was determined by examining the 95% CI of the ratio of the relative treatment effect estimates from observational studies and RCTs derived from the Monte Carlo simulation. We examined differences in relative effect measures from observational studies and RCTs by outcome type and therapeutic area.
To test the robustness of our findings, we conducted two sensitivity analyses. As some reviews assessed more than one endpoint and contributed more than one pair of pooled relative treatment effects from RCTs and observational studies to our analysis, we repeated the analysis with one endpoint per review, i.e., a single pair of pooled relative treatment effects from RCTs and observational studies from each review, selecting the most frequently used endpoints for inclusion whenever possible. Additionally, as some studies were included in more than one review, we repeated the analysis ensuring that there is no overlap of data between the included reviews, i.e., ensuring that each study was included in only one review included in our analysis. Details on the sensitivity analyses are included in Additional File 2. All analyses were conducted using RStudio, version 1.3.1073 (©2009-2020 RStudio, PBC).

Results

Our search on PubMed and Embase yielded 3798 unique citations after removing duplicates. After screening titles and abstracts, we identified 93 full text articles for further review. Of these, 30 reviews met our inclusion criteria (Fig. 1).

Included systematic reviews

The characteristics of the included reviews and the pairs of pooled relative treatment effects from RCTs and observational studies reported in the reviews are summarized in Table 1. Thirty systematic reviews across 7 therapeutic areas (cardiovascular disease [15/30], infectious disease [6/30], oncology [3/30], mental health [2/30], immune-inflammatory [1/30], metabolic disease [1/30], and other [2/30]) were identified from the literature. These reviews included 519 RCTs and observational studies and provided 79 pairs of pooled relative treatment effects from RCTs and observational studies across multiple interventions, comparators, and outcomes. Five pairs were excluded from our assessment because they concerned continuous outcomes (n = 1) or no pooled effect estimate was reported for observational studies (n = 4). As a result, 74 pairs of pooled relative treatment effects from RCTs and observational studies from 29 reviews were available for assessment of consistency.
Table 1
Characteristics of included reviews
Review
Disease
Treatment
Comparator
Endpoint
Number of RCTs
Number of Observational Studies
RCT pooled effect estimate (95% CI)
Observational pooled effect estimate (95% CI)
Abuzaid (2018) [17]
Severe aortic stenosis
Dual anti-platelet therapy (DAPT)
Single anti-platelet therapy (SAPT)
30-day all-cause mortality
3
7
RR 1.2 (0.5–2.89)
RR 1.19 (0.74–1.91)
Abuzaid (2018) [17]
Severe aortic stenosis
DAPT
SAPT
Longest reported all-cause mortality
3
7
RR 1.14 (0.54–2.42)
RR 1.06 (0.52–2.18)
Abuzaid (2018) [17]
Severe aortic stenosis
DAPT
SAPT
Major bleeding
3
7
RR 1.74 (0.52–5.82)
RR 2.23 (1.36–3.65)
Agarwal (2019) [18]
Acute coronary syndrome (ACS), coronary artery disease (CAD)
Dual therapy
Triple therapy
Major bleeding
3
3
RR 0.53 (0.38–0.76)
RR 0.88 (0.46–1.67)
Agarwal (2018) [19]
CAD
DAPT
Aspirin
Primary outcome: mid- to long-term (> 30 days) composite of myocardial infarction (MI), stroke, or death
8
4
RR 0.43 (0.17–1.11)
RR 0.85 (0.72–1.01)
An (2019) [20]
Severe aortic stenosis
Antiplatelet
Anticoagulation
Mortality
2
5
RR 0.82 (0.33–2.03)
RR 0.47 (0.18–1.22)
An (2019) [20]
Severe aortic stenosis
Antiplatelet
Anticoagulation
Stroke/transient ischemic attack (TIA)
2
5
RR 0.9 (0.35–2.33)
RR 0.57 (0.31–1.03)
An (2019) [20]
Severe aortic stenosis
Antiplatelet
Anticoagulation
Thromboembolic events
2
5
RR 1.13 (0.51–2.49)
RR 0.71 (0.38–1.32)
An (2019) [20]
Severe aortic stenosis
Antiplatelet
Anticoagulation
Bleeding
2
5
RR 0.34 (0.11–1.04)
RR 0.34 (0.2–0.58)
Chien (2020)* [21]
Multi-drug resistant gram-negative bacteria (MDR-GNB) infections
Colistin
Other antibiotics
Colistin-associated acute kidney injury (CA-AKI)
1
19
OR 2.75 (0.43–17.49)
Not reported
Chien (2020) [21]
MDR-GNB infections
Colistin monotherapy
Colistin combination therapy
Acute kidney injury (AKI)
3
6
OR 1.77 (1.17–2.66)
OR 1.15 (0.76–1.76)
Chopra (2012)* [22]
Pneumonia
Statin therapy
No statin therapy
Unadjusted all-cause mortality following an episode of pneumonia
1
9
OR 0.84 (0.32–2.18)
Not reported
Chopra (2012) [22]
Pneumonia
Statin therapy
No statin therapy
Adjusted all-cause mortality following an episode of pneumonia
1
11
OR 0.84 (0.32–2.18)
OR 0.66 (0.55–0.79)
Desai (2016) [23]
Ankylosing spondylitis, inflammatory bowel diseases, juvenile idiopathic arthritis, plaque psoriasis, psoriatic arthritis, and rheumatoid arthritis
Adalimumab
Etanercept
Discontinuation due to adverse events
1
3
RR 0.83 (0.39–1.78)
Adjusted HR 1.67 (1.26–2.22)
Desai (2016) [23]
Ankylosing spondylitis, inflammatory bowel diseases, juvenile idiopathic arthritis, plaque psoriasis, psoriatic arthritis, rheumatoid arthritis
Adalimumab
Infliximab
Discontinuation due to adverse events
1
5
RR 6.17 (0.78–48.71)
Adjusted HR 0.57 (0.46–0.7)
Gandhi (2015) [24]
Aortic stenosis
DAPT
Mono-antiplatelet therapy (MAPT).
Combined end point of 30-day major stroke, spontaneous MI, all-cause mortality, and combined lethal and major bleeding
2
2
OR 0.98 (0.46–2.11)
OR 3.02 (1.91–4.76)
Ge (2018) [25]
Atrial fibrillation
Novel oral anticoagulants (NOACs)
Vitamin K antagonists (VKAs)
Major bleeding events (Fixed effects model)
4
25
OR 0.3 (0.14–0.62)
OR 0.68 (0.48–0.95)
Ge (2018) [25]
Atrial fibrillation
NOACs
VKAs
Thromboembolic events (Fixed effects model)
4
25
OR 0.14 (0.01–1.3)
OR 0.91 (0.49–1.67)
Heffernan (2020) [26]
Serious infections
β-lactam/aminoglycoside combination therapy
β-lactam monotherapy
All-cause mortality
2
4
OR 3.18 (0.79–12.73)
OR 0.79 (0.64–0.99)
Ho (2013) [27]
Kidney disease (kidney transplant)
Once daily tacrolimus
Twice daily tacrolimus
Biopsy-proven acute rejection (RCT: at 6 months; Observational: mean follow-up ranges from 3 months to 672 months)
4
5
RR 1.18 (0.82–1.68)
RR 0.83 (0.39–1.78)
Ho (2013) [27]
Kidney disease (kidney transplant)
Once daily tacrolimus
Twice daily tacrolimus
Biopsy-proven acute rejection (RCT: at 12 months
Observational: mean follow-up ranges from 3 months to 672 months)
2
5
RR 1.24 (0.93–1.65)
RR 0.83 (0.39–1.78)
Ho (2013) [27]
Kidney disease (kidney transplant)
RCT: twice daily tacrolimus
Observational: once daily tacrolimus
RCT: once daily tacrolimus
Observational: twice daily tacrolimus
Patient survival (RCT: at 6 months
Observational: mean follow-up ranges from 3.5 months to 12 months)
2
2
RR 1.03 (1–1.06)
RR 1.02 (0.94–1.1)
Ho (2013) [27]
Kidney disease (kidney transplant)
RCT: twice daily tacrolimus
Observational: once daily tacrolimus
RCT: once daily tacrolimus
Observational: twice daily tacrolimus
Patient survival (RCT: at 12 months
Observational: mean follow-up ranges from 3.5 months to 12 months)
3
2
RR 0.99 (0.97–1.02)
RR 1.02 (0.94–1.1)
Khan (2019) [28]
CAD
Proton pump inhibitor (PPI)
No PPI
All-cause mortality
3
24
RR 1.35 (0.56–3.23)
RR 1.25 (1.11–1.41)
Kirson (2013)* [29]
Schizophrenia
Depot antipsychotics
Oral antipsychotics
Varies across studies: hospitalization, relapse, discontinuation
5
8
RR 0.89 (0.64–1.22)
Not reported
Land (2017) [30]
Psychiatric illnesses
Clozapine
Control drugs (other antipsychotics)
Hospitalization
3
19
RR 0.62 (0.41–0.94)
RR 0.75 (0.69–0.81)
Li (2016) [31]
Diabetes
Dipeptidyl peptidase-4 (DPP-4) inhibitors
RCT: control
Observational: sulfonylurea (SU)
Heart failure
38
1
OR 0.97 (0.61–1.56)
Unadjusted OR 0.88 (0.22–3.48)
Li (2016) [31]
Diabetes
DPP-4 inhibitors
RCT: control
Observational: SU
Heart failure
38
1
OR 0.97 (0.61–1.56)
Adjusted HR 1.10 (1.04–1.17)
Li (2016) [31]
Diabetes
RCT: DPP-4 inhibitors
Observational: sitagliptin
RCT: control
Observational: SU
Heart failure
38
1
OR 0.97 (0.61–1.56)
Unadjusted OR 0.39 (0.02–6.26)
Li (2016) [31]
Diabetes
RCT: DPP-4 inhibitors
Observational: sitagliptin
RCT: control
Observational: no sitagliptin use
Heart failure
38
1
OR 0.97 (0.61–1.56)
Adjusted OR 0.75 (0.38–1.46)
Li (2016) [31]
Diabetes
DPP-4 inhibitors
RCT: control
Observational: active control
Hospital admission for heart failure
5
6
OR 1.13 (1.00–1.26)
Adjusted OR 0.85 (0.74–0.97)
Li (2016) [31]
Diabetes
DPP-4 inhibitors
RCT: control
Observational: SU
Hospital admission for heart failure
5
3
OR 1.13 (1.00–1.26)
Adjusted HR 0.84 (0.74–0.96)
Li (2016) [31]
Diabetes
DPP-4 inhibitors
RCT: control
Observational: pioglitazone
Hospital admission for heart failure
5
2
OR 1.13 (1.00–1.26)
Adjusted HR 0.67 (0.57–0.78)
Li (2016) [31]
Diabetes
DPP-4 inhibitors
RCT: control
Observational: other oral antidiabetics
Hospital admission for heart failure
5
1
OR 1.13 (1.00–1.26)
Adjusted OR 0.88 (0.63–1.22)
Li (2016) [31]
Diabetes
DPP-4 inhibitors
Control
Hospital admission for heart failure
5
1
OR 1.13 (1.00–1.26)
Adjusted HR 0.58 (0.38–0.88)
Li (2016) [31]
Diabetes
RCT: DPP-4 inhibitors
Observational: sitagliptin
RCT: control
Observational: no sitagliptin use
Hospital admission for heart failure
5
2
OR 1.13 (1.00–1.26)
Adjusted OR 1.41 (0.95–2.09)
Li (2016) [31]
Diabetes
RCT: DPP-4 inhibitors
Observational: sitagliptin
RCT: control
Observational: no sitagliptin use
Hospital admission for heart failure
5
1
OR 1.13 (1.00–1.26)
Adjusted HR 1.21 (1.04–1.42)
Li (2016) [31]
Diabetes
RCT: DPP-4 inhibitors
Observational: sitagliptin
RCT: control
Observational: no sitagliptin use
Hospital admission for heart failure
5
1
OR 1.13 (1.00–1.26)
Adjusted OR 1.84 (1.16–2.92)
Melloni (2015) [32]
Unstable angina/non–ST-segment–elevation myocardial infarction (UA/NSTEMI)
RCT: omeprazole
Observational: any PPI
RCT: placebo
Observational: no PPI
Composite ischemic endpoint at ≈ 1 year
1
20
HR 0.99 (0.68–1.44)
Adjusted HR 1.35 (1.18–1.54)
Melloni (2015) [32]
UA/NSTEMI
RCT: omeprazole
Observational: any PPI
RCT: placebo
Observational: no PPI
Nonfatal MI at ≈ 1 year
1
10
HR 0.92 (0.44–1.9)
HR 1.331 (1.146–1.547)
Miles (2019) [33]
Heart failure
Furosemide
Torsemide
All-cause mortality
5
3
OR 1.12 (0.7–1.8)
OR 0.97 (0.44–2.13)
Miles (2019) [33]
Heart failure
Furosemide
Torsemide
Heart failure readmissions
4
1
OR 2.04 (1.16–3.60)
OR 2.91 (0.78–10.91)
Miles (2019) [33]
Heart failure
Furosemide
Torsemide
New York Heart Association class improvement
7
2
OR 0.91 (0.61–1.35)
OR 0.65 (0.50–0.85)
Mongkhon (2019) [34]
Atrial fibrillation
OAC
Non-OAC
Risk of dementia
1
4
RR 1.31 (0.79–2.18)
RR 0.75 (0.67–0.83)
Mongkhon (2019) [34]
Atrial fibrillation
VKA
Non-VKA
Risk of dementia
1
4
RR 1.31 (0.79–2.18)
RR 0.71 (0.68–0.74)
Raheja (2018) [35]
Aortic stenosis
DAPT
SAPT
All-cause mortality
3
2
RR 1.07 (0.48–2.41)
RR 1.34 (0.51–3.48)
Raheja (2018) [35]
Aortic stenosis
DAPT
SAPT
Stroke or TIA
3
2
RR 0.93 (0.28–3.06)
RR 1.25 (0.32–4.92)
Raheja (2018) [35]
Aortic stenosis
DAPT
SAPT
MI
3
2
RR 3.62 (0.60–21.76)
RR 1.18 (0.14–9.98)
Raheja (2018) [35]
Aortic stenosis
DAPT
SAPT
Major/life-threatening bleeding
3
3
RR 1.75 (0.88–3.50)
RR 3.24 (1.82–5.75)
Ramjan (2014) [36]
HIV
Fixed-dose combination (FDC) antiretroviral therapy (ART)
Separate tablet regimens
Virological suppression
4
2
RR 1.04 (0.98–1.10)
RR 1.07 (0.97–1.18)
Ramjan (2014) [36]
HIV
FDC ART
Separate tablet regimens
Adherence to ART
5
2
RR 1.1 (0.98–1.22)
RR 1.17 (1.07–1.28)
Shi (2014) [37]
Liver cancer
Statins
Placebo/non-use
Liver cancer
1
11
RR 1.06 (0.66–1.71)
RR 0.57 (0.50–0.64)
Teo (2014) [38]
Acute infections
Prolonged infusion, which was defined as administration of either extended infusion or continuous infusion of beta-lactam antibiotics
Identical beta-lactams that were administered as intermittent boluses (20–60 min infusion) according to the manufacturer’s package insert
All-cause in-hospital mortality
10
9
RR 0.83 (0.57–1.21)
RR 0.57 (0.43–0.76)
Teo (2014) [38]
Acute infections
Prolonged infusion, which was defined as administration of either extended infusion or continuous infusion of beta-lactam antibiotics.
Identical beta-lactams that were administered as intermittent boluses (20–60 min infusion) according to the manufacturer’s package insert
Clinical success (cure or improvement)
14
5
RR 1.05 (0.99–1.12)
RR 1.34 (1.02–1.76)
Vinceti (2018) [39]
Cancer
Highest selenium exposure
Lowest selenium exposure
Total (any) cancer incidence
3
7
RR 1.01 (0.93–1.10)
OR 0.72 (0.55–0.93)
Vinceti (2018) [39]
Cancer
Highest selenium exposure
Lowest selenium exposure
Cancer mortality
1
7
RR 1.02 (0.80–1.30)
OR 0.76 (0.59–0.97)
Vinceti (2018) [39]
Colorectal cancer
Highest selenium exposure
Lowest selenium exposure
Colorectal cancer risk
2
1
RR 0.99 (0.69–1.43)
OR 0.80 (0.68–0.94)
Vinceti (2018) [39]
Lung cancer
Highest selenium exposure
Lowest selenium exposure
Lung cancer risk
2
5
RR 1.16 (0.89–1.50)
OR 0.74 (0.43–1.28)
Vinceti (2018) [39]
Breast cancer
Highest selenium exposure
Lowest selenium exposure
Breast cancer risk
1
8
RR 2.04 (0.44–9.55)
OR 1.09 (0.87–1.37)
Vinceti (2018) [39]
Bladder cancer
Highest selenium exposure
Lowest selenium exposure
Bladder cancer risk
2
2
RR 1.07 (0.76–1.52)
OR 0.65 (0.46–0.92)
Vinceti (2018) [39]
Prostate cancer
Highest selenium exposure
Lowest selenium exposure
Prostate cancer risk
4
21
RR 1.01 (0.90–1.14)
OR 0.84 (0.75–0.95)
Wang (2019) [40]
Pneumonia
PPI
No PPI
Pneumonia
10
48
OR 1.13 (0.71–1.78)
OR 1.45 (1.32–1.59)
Wat (2019) [41]
Traumatic brain injury (TBI)
Antiepileptic drugs
Placebo/no treatment
Early seizures after TBI
3
6
RR 0.58 (0.20–1.72)
RR 0.42 (0.29–0.62)
Wong (2017)* [42]
Coronary heart disease/CAD
Macrolides
Placebo/no treatment
Short-term primary outcome (defined as cardiac mortality, cardiovascular mortality, sudden death, cardiac arrest, all-cause mortality, or composite outcomes including death and/or other cardiovascular events or procedures)
5
15
RR 0.99 (0.74–1.34)
Not reported
Wong (2017) [42]
Coronary heart disease/CAD
Macrolides
Placebo/no treatment
Long term primary outcome (defined as cardiac mortality, cardiovascular mortality, sudden death, cardiac arrest, all-cause mortality, or composite outcomes including death and/or other cardiovascular events or procedures)
14
8
RR 1.03 (0.96–1.10)
RR 1.05 (0.91–1.22)
Yang (2019)* [43]
Cancer
Epoetin alfa biosimilar drugs
Epoetin alfa drugs
Mean of hemoglobin increase
1
4
WMD -0.02 (− 0.38–0.34)
WMD 0.07 (− 0.12–0.25)
Yang (2019) [43]
Cancer
Epoetin alfa biosimilar drugs
Epoetin alfa drugs
Hemoglobin response
1
1
RR 1.09 (0.86–1.38)
RR 1.18 (0.87–1.60)
Yang (2019) [43]
Breast cancer
Granulocyte colony-stimulating factor (G-CSF) biosimilar drugs
G-CSF drugs
Febrile neutropenia in cycle 1
5
3
RR 1.14 (0.80–1.63)
RR 1.36 (0.84–2.23)
Yang (2019) [43]
NHL
G-CSF biosimilar drugs
G-CSF drugs
Febrile neutropenia in cycle 1
1
1
RR 0.54 (0.20–1.46)
RR 0.87 (0.20–3.85)
Yang (2019) [43]
Cancer
G-CSF biosimilar drugs (filgrastim biosimilars)
G-CSF drugs
Bone pain
4
4
RR 0.90 (0.78–1.05)
RR 0.86 (0.59–1.24)
Yu (2018) [44]
Non-cardiac vascular disease
Statins
Placebo/no statin treatment
All-cause mortality
3
6
OR 0.62 (0.41–0.92)
OR 0.65 (0.48–0.88)
Yu (2018) [44]
Non-cardiac vascular disease
Statins
Placebo/no statin treatment
Primary patency
1
10
OR 0.39 (0.09–1.65)
OR 0.77 (0.59–0.99)
Yu (2018) [44]
Non-cardiac vascular disease
Statins
Placebo/no statin treatment
Amputation
1
10
OR 0.47 (0.07–2.94)
OR 0.64 (0.50–0.83)
Yu (2018) [44]
Non-cardiac vascular disease
Statins
Placebo/no statin treatment
Cardiovascular events
3
2
OR 0.55 (0.36–0.83)
OR 0.87 (0.16–4.60)
Zhang (2019) [45]
Atrial fibrillation
NOAC
Non-NOAC therapy
Renal impairment
11
3
HR 0.82 (0.71–0.93)
HR 0.64 (0.58–0.69)
Zhao (2018) [46]
CAD
DAPT
SAPT
Any bleeding events
5
8
RR 1.25 (0.98–1.59)
RR 0.87 (0.76–1.01)
Zhao (2018) [46]
CAD
DAPT
SAPT
Minor bleeding events
4
3
RR 1.15 (0.73–1.81)
RR 0.84 (0.37–1.93)
Zhao (2018) [46]
CAD
DAPT
SAPT
Major bleeding events
5
6
RR 1.28 (0.95–1.71)
RR 0.99 (0.66–1.51)
Zhao (2018) [46]
CAD
DAPT
SAPT
Major bleeding events during hospitalization (random effects model)
3
3
RR 1.27 (0.91–1.78)
RR 0.50 (0.12–2.09)
*Not included in the analysis

Ratio of relative effect measures from observational studies and RCTs

Figure 2 presents the scatterplot of relative effect measures from observational studies and RCTs across the 74 pairs of pooled relative treatment effects with the 95% CI bars. The ratio of the relative effect measure from observational studies over the corresponding relative effect measure from RCTs ranged from 0.09 to 6.50 (median = 0.92, interquartile range = 0.69–1.27). The ratio was greater than 1, i.e., the relative effect was larger in observational studies in 31 of the 74 pairs (41.9%). The ratio was less than 1, i.e., the relative effect was larger in RCTs in 42 of the 74 pairs (56.8%), and the ratio was equal to 1 in one of the 74 pairs (1.4%). The ratio was greater than 1.43 in 12 of the 74 pairs (16.2%) and less than 0.7 in 20 of the 74 pairs (27.0%) indicating an extreme difference. There was an absence of an extreme difference (0.7 ≤ ratio ≤ 1.43) in 42 of the 74 pairs (56.8%; Table 2). Sensitivity analyses including only one endpoint from each review and ensuring no overlap of data between the included reviews resulted in similar findings (Table 2). Scatterplots of relative effect measures from observational studies and RCTs by outcome type and therapeutic area can be found in Additional File 3: Figures S1 and S2.
Table 2
Ratio of relative effect measures from observational studies and relative effect measures from RCTs (e.g., RRobs/RRrct): (a) among 74 pairs of pooled estimates, (b) with only one endpoint per review included, and (c) with studies reported in multiple reviews excluded
 
Full sample
One endpoint per review
Studies reported in multiple reviews excluded
 
Proportion
%
Proportion
%
Proportion
%
Ratio > 1*a
31/74
41.9
12/29
41.4
24/65
36.9
Ratio < 1*b
42/74
56.8
17/29
58.6
40/65
61.5
Ratio = 1*
1/74
1.4
0/29
0.0
1/65
1.5
Extreme difference (ratio > 1.43)
12/74
16.2
5/29
17.2
8/65
12.3
Extreme difference (ratio < 0.7)
20/74
27.0
8/29
27.6
19/65
29.2
Absence of an extreme difference (0.7 ≤ ratio ≤ 1.43)
42/74
56.8
16/29
55.2
38/65
58.5
*Does not account for direction of effect
aRelative effect larger in observational studies
bRelative effect larger in RCTs

Consistency of relative effect measures from observational studies and RCTs

In 30 of the 74 pairs (40.5%), effect estimates from observational studies and RCTs pointed in opposite directions of effect. The RCT point estimate was outside the 95% CI of the observational study in 35 of the 74 pairs (47.3%) and the observational study point estimate was outside the 95% CI of the RCT in 27 of the 74 pairs (36.5%). There was a statistically significant difference between relative effect estimates from observational studies and RCTs in 15 of the 74 pairs (20.3%). In 13 of the 74 pairs (17.6%), there was a statistically significant difference and the effect estimates of observational studies and RCTs pointed in opposite directions (Table 3). The results remained fairly consistent when the sensitivity analyses were conducted (Table 3).
Table 3
Consistency of relative effect measures from observational studies and relative effect measures from RCTs: (a) among 74 pairs of pooled estimates, (b) with only one endpoint per review included, and (c) with studies reported in multiple reviews excluded
 
Full sample
One endpoint per review
Studies reported in multiple reviews excluded
 
Proportion
%
Proportion
%
Proportion
%
Effect estimates of observational studies and RCTs in opposite directions
30/74
40.5
11/29
37.9
26/65
40.0
RCT effect estimate outside the observational study 95% CI
35/74
47.3
17/29
58.6
29/65
44.6
Observational effect estimate outside the RCT 95% CI
27/74
36.5
11/29
37.9
25/65
38.5
Statistically significant difference
15/74
20.3
7/29
24.1
14/65
21.5
Statistically significant difference and effect estimates of observational studies and RCTs in opposite directions
13/74
17.6
6/29
20.7
12/65
18.5

Discussion

Our analysis of 29 reviews comparing results of RCTs and observational studies of pharmaceuticals showed, on average, no significant differences in their relative risk ratios across all studies, but also considerable study-by-study variability. The median ratio of the relative effect measure from observational studies to RCTs was 0.92, indicating just slightly lower effectiveness/safety estimates in observational studies than corresponding RCTs. This is in fact somewhat higher than the 0.80 ratio recently found in meta-research comparing effect estimates of randomized clinical trials that use routinely collected data (i.e., from traditional observational study sources such as registries, electronic health records, or administrative claims) for outcome ascertainment with traditional trials not using routinely collected data [47]. However, whether judging by the frequency of “extreme” differences (43.2%) or statistically significant differences in opposite directions (17.6%), one could not claim that observational study results consistently replicated RCT results on a study-by-study basis in our sample.
There are a number of reasons that any given observational study result may not replicate an RCT comparing the same treatments. First, it may not have been the intent of the observational study researchers to match a specific clinical trial—they may have intentionally studied a different treatment population, setting, or protocol in order to complement or test the RCT findings. In such cases, there would be variation in effect estimates due to estimating a different causal effect. Even if the researcher does attempt to match a specific RCT, the data may not have been available to closely match it, since patient histories, test results, etc., used for RCT inclusion criteria may not be observed, or outcomes may not be captured the same way. Even given similar data, non-randomized studies have the potential for selection/channeling bias into treatment determined by factors unobservable in either type of study, and analytic attempts to correct for such confounding may have limited success. In some cases, treatment conditions may differ enough between the RCT and real-world practice that replication of results should not be expected, e.g., due to careful safety monitoring that affects subsequent treatment in RCTs. Finally, it is possible that other pharmacoepidemiologic principles, beyond the study design considerations we already mentioned, were violated in the individual RWD studies, which could have caused disagreement between their results and the RCTs. While variation in treatment effect estimates due to estimating a different causal effect in a different study population is expected and valid, biased estimates arising from issues with study design or analytical methods may be problematic.
Details in these reviews were typically insufficient to distinguish among these possible explanations, without detailed review of the individual studies, which we did not attempt here. However, some reviews did attempt to explain the differences they found. For example, in the review by Gandhi et al. (2015) [24], which compared dual-antiplatelet therapy (DAPT) to mono-antiplatelet therapy (MAPT) following transcatheter aortic valve implantation, there was a statistically significant difference in pooled relative treatment effect estimates from observational studies and RCTs. The primary outcome was more likely to occur in the DAPT group than in the MAPT group in the observational studies (OR 3.02; 95% CI 1.91–4.76); however, no statistically significant difference was found between DAPT and MAPT in the RCTs (OR 0.98; 95% CI 0.46–2.11). The authors explained that the RCTs (n = 2) and observational studies (n = 2) included in this review had variable patient inclusion/exclusion criteria and there were differences in the type of prosthetic aortic valve used, which may have introduced selection bias [24].
To allow for better use of individual observational studies to inform decision-making, their ability to replicate RCT results needs to become more reliable, and the “target trial” approach seems to be a path forward. Several systematic efforts using sophisticated observational data research designs to emulate multiple RCTs are underway [48, 49]. These efforts are intended to provide regulatory bodies and other decision makers with empirical evidence to support the development of a framework for assessing when and under what circumstances observational RWE can be used to support a wider range of regulatory decisions. RCT DUPLICATE is a collaboration between the Food and Drug Administration (FDA), Brigham and Women’s Hospital and Harvard Medical School Division of Pharmacoepidemiology, to replicate 30 completed Phase III or IV trials and to predict the results of seven ongoing Phase IV trials using Medicare and commercial claims data [50]. The RCT DUPLICATE team has recently reported results for its first 10 trials [51]. They report hazard ratio estimates within the 95% CI of the corresponding trial for 8 of 10 emulations.
The Multi-Regional Clinical Trials Center and OptumLabs are leading another effort called Observational Patient Evidence for Regulatory Approval and Understanding Disease (OPERAND) which extends the trial emulation activity and relaxes the inclusion/exclusion criteria of the trials to examine treatment effects in the broader patient population treated in routine care [52]. The FDA has also funded the Yale University-Mayo Clinic Center of Excellence in Regulatory Science and Innovation to predict the results of three to four ongoing safety trials using OptumLabs claims data [53].
It is important to understand that clinical trials emulation efforts are being conducted solely to improve understanding of when observational studies may be expected to produce robust results. Bartlett and colleagues [14] found that in a review of 220 clinical trials published in high impact medical journals in 2017, 15% could potentially be emulated using data available from medical claims or EHRs. For example, the inclusion/exclusion criteria for many oncology trials require data on genetic markers and progression free survival unavailable in EHRs. The estimate by Bartlett and colleagues may prove to be an underestimate as the ability to link different types of observational data continues to improve. Nevertheless, it is reasonable to assume that it is not possible to emulate most trials with existing observational datasets.
These efforts are critical to advance our understanding of the strengths and limitations of observational RWE, identifying issues with study design, endpoint definition, data quality, and analytical methodology that may impact the consistency of findings between RWE and RCTs. While much attention has focused on differences in study populations between observational studies and RCTs as the reason for the inconsistency in effect estimates, emerging evidence suggests that issues with study design (e.g., establishing time zero of exposure) may be equally if not more important [7]. Therefore, the results of these efforts will not provide definitive guidance to decision makers but they emphasize how even subtle differences in study design and endpoint definition can impact absolute estimates of treatment effect. Moreover, RWE studies are answering a different question than RCTs, i.e., “Does it work?” verses “Can it Work?” The former is important to a variety of stakeholders beyond regulators. Hence, they should not be expected to provide results identical to RCTs.

Conclusions

In conclusion, although our review shows no average significant difference in the relative risk ratios between published RCTs and observational studies, there is substantial study-to-study variation. It was impractical to review all individual observational study designs and examine their potential biases, but future work should elucidate how much of the variation is due to differences in study populations versus biased estimates arising from issues with study design or analytical methods. As more target trial replication attempts are conducted and published, more systematic evidence will emerge on the reliability of this approach and on the potential for observational studies to more routinely inform healthcare decisions.

Acknowledgements

Not applicable

Declarations

Not applicable.
Not applicable.

Competing interests

Yoon Duk Hong, John Guerino, Marc L. Berger, William Crown, Richard J. Willke, Wim G. Goettsch, and Lucinda S. Orsini have no conflicts of interest to report. Jeroen P. Jansen is a part-time employee of Precision Medicine Group (PMG) (PRECISIONheor) and has stock options from Precision Medicine Group. PMG provides contracted research services to pharmaceutical and biotech industry. C. Daniel Mullins has received consulting fees from AstraZeneca, Bayer, Incyte, Merck, Pfizer, and Takeda and has received support from Bayer and Pfizer for attending meetings and/or travel.
Open AccessThis article is licensed under a Creative Commons Attribution 4.0 International License, which permits use, sharing, adaptation, distribution and reproduction in any medium or format, as long as you give appropriate credit to the original author(s) and the source, provide a link to the Creative Commons licence, and indicate if changes were made. The images or other third party material in this article are included in the article's Creative Commons licence, unless indicated otherwise in a credit line to the material. If material is not included in the article's Creative Commons licence and your intended use is not permitted by statutory regulation or exceeds the permitted use, you will need to obtain permission directly from the copyright holder. To view a copy of this licence, visit http://​creativecommons.​org/​licenses/​by/​4.​0/​. The Creative Commons Public Domain Dedication waiver (http://​creativecommons.​org/​publicdomain/​zero/​1.​0/​) applies to the data made available in this article, unless otherwise stated in a credit line to the data.

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Literatur
52.
Zurück zum Zitat Evaluating RWE from observational studies in regulatory decision-making: lessons learned from trial replication analyses. Trial Emulation Studies and OPERAND. Duke-Margolis Center for Health Policy Virtual Meeting, February 16-17, 2021. Evaluating RWE from observational studies in regulatory decision-making: lessons learned from trial replication analyses. Trial Emulation Studies and OPERAND. Duke-Margolis Center for Health Policy Virtual Meeting, February 16-17, 2021.
Metadaten
Titel
Comparative effectiveness and safety of pharmaceuticals assessed in observational studies compared with randomized controlled trials
verfasst von
Yoon Duk Hong
Jeroen P. Jansen
John Guerino
Marc L. Berger
William Crown
Wim G. Goettsch
C. Daniel Mullins
Richard J. Willke
Lucinda S. Orsini
Publikationsdatum
01.12.2021
Verlag
BioMed Central
Erschienen in
BMC Medicine / Ausgabe 1/2021
Elektronische ISSN: 1741-7015
DOI
https://doi.org/10.1186/s12916-021-02176-1

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