Description of the condition
Raynaud's phenomenon (episodic colour changes of the fingers and toes, usually in response to cold exposure or to emotional stress) may be primary (idiopathic) or secondary to a variety of conditions including connective tissue diseases (especially systemic sclerosis-spectrum disorders), extrinsic vascular compression, use of vibratory equipment (hand-arm-vibration-syndrome), hyperviscosity states, and certain drugs or chemicals (Block 2001). The classic colour changes are white (ischaemia), blue (deoxygenation) and red (reperfusion).
This review of calcium channel blockers is concerned only with primary Raynaud's phenomenon. A key differentiating feature between primary and secondary Raynaud's phenomenon is that in primary Raynaud's phenomenon vasospasm is entirely reversible: irreversible tissue injury does not occur (LeRoy 1992). This is in contrast to the situation in systemic sclerosis, for example, when digital ischaemia can be severe and result in ulceration, scarring, and sometimes even gangrene. Although primary Raynaud's phenomenon is therefore often considered 'benign', it can cause considerable discomfort and distress in severely affected people.
The pathogenesis of primary Raynaud's phenomenon is not known but it is likely that abnormalities in the neural control elements of vascular tone play a key role (Herrick 2005). Primary Raynaud's phenomenon is common; its reported prevalence varies, in part related to geography, being less prevalent in warmer climates (Maricq 1997). A UK general practice-based study suggested that 15% to 19% of the population experience Raynaud's phenomenon (Silman 1990), and of these most will have primary Raynaud's phenomenon. Women are more affected than men.
A careful history and examination will usually indicate whether or not a person with Raynaud's phenomenon is likely to have the primary form or whether there is an underlying cause. The key investigations to perform are an erythrocyte sedimentation rate (normal in primary Raynaud's phenomenon), antinuclear antibody (ANA) (negative or only weakly positive in people with primary Raynaud's phenomenon) and nailfold capillaroscopy (normal in people with primary Raynaud's phenomenon) (LeRoy 1992; Wigley 2002). In the person with Raynaud's phenomenon, systemic sclerosis-specific autoantibodies and nailfold capillaroscopic abnormalities are independent risk factors for systemic sclerosis (Koenig 2008).
Many people with primary Raynaud's phenomenon are reassured to know that they do not have a serious underlying disease and respond well to conservative measures, including keeping warm and avoiding drugs with vasoconstrictive effects (Suter 2005). Smoking cessation should be strongly encouraged. Although smoking has not been proven to adversely affect primary Raynaud's phenomenon, it is likely that Raynaud's could be exacerbated via a variety of possible mechanisms including endothelial injury and increased plasma viscosity. Although some sufferers have reported benefit from temperature biofeedback, this was shown to be ineffective in a large randomised controlled trial of 313 participants with primary Raynaud's phenomenon (RTS 2000).
For those people with primary Raynaud's phenomenon who do not respond to conservative (i.e. 'non-drug') measures, drug treatment will often be required. The aim of drug treatment is to prevent vasospasm or to increase vasodilation, or both. Despite the high prevalence of primary Raynaud's phenomenon, there have been relatively few controlled clinical trials, probably reflecting how difficult clinical trials of Raynaud's phenomenon are to set up. These have to be run over the winter months, there is a large placebo effect, and there is a lack of reliable outcome measures which are sensitive to change. The Raynaud's Condition Score, a self-assessment score (0 to 10) which incorporates frequency, duration and severity of attacks, is increasingly used in clinical trials (Merkel 2002).
Description of the intervention
Calcium channel blockers are generally considered first-line in the drug treatment of primary Raynaud's phenomenon, and are the group of drugs which have been most researched. A meta-analysis published in 2005 concluded that calcium channel blockers had some effect in primary Raynaud's phenomenon in reducing the frequency and severity of Raynaud's attacks, but that this was small (2.8 to 5.0 fewer attacks per week with a 33% reduction in severity) (Thompson 2005). Other drugs currently used include angiotensin converting enzyme (ACE) inhibitors, angiotensin II receptor antagonists (Dziadzio 1999) and alpha-adrenergic blockers (Harding 1998). Intravenous prostanoids, used to treat severe exacerbations of digital ischaemia and ulceration in people with systemic sclerosis-spectrum disorders, are not generally indicated for primary Raynaud's phenomenon. Similarly, surgical intervention, for example digital (palmar) sympathectomy, may be indicated for severe secondary Raynaud's phenomenon but not in people with primary Raynaud's phenomenon.
How the intervention might work
Calcium channel blockers bind to L-type voltage-gated calcium channels on cells, preventing influx of extracellular calcium. There are two main categories of calcium channel blocker: the dihydropyridines (these include nifedipine, nicardipine and amlodipine) and the non-dihydropyridines (these include diltiazem and verapamil) (Sturgill 1998). The dihydropyridines are more selective for vascular smooth muscle than the non-dihydropyridines: they are potent vasodilators with minimal effect on cardiac contractility or conduction, and can cause reflex tachycardia. In contrast, verapamil and diltiazem are negatively inotropic and can slow cardiac conduction, and are less potent vasodilators than the dihydropyridines. The calcium channel blockers which have been most studied in primary Raynaud's phenomenon, and forming the basis of this review, are dihydropyridines.
Why it is important to do this review
Calcium channel blockers are currently the most commonly prescribed drug treatment for primary Raynaud's phenomenon.
To assess the effects of different calcium channel blockers for primary Raynaud's phenomenon as determined by attack rates, severity scores, participant-preference scores and physiological measurements.
Criteria for considering studies for this review
Types of studies
We considered all randomised and blinded trials of the effects of orally-administered calcium channel blockers compared with placebo, or other drug therapy, in people with primary Raynaud's phenomenon, using any method of randomisation. We included trials not analysed on an intention-to-treat basis, provided all randomised participants were accounted for. We included cross-over trials only if they incorporated (a) a placebo or no-treatment 'run-in' of at least one week (or provided baseline data for quantification of the severity of the condition), and (b) a 'wash-out' period of at least one week between treatment arms for all participants. We identified but did not include in the meta-analyses studies in which primary and secondary cases could not be resolved. We also excluded open studies and single-dose studies. There was no restriction on language, with publications in languages other than English translated and assessed for eligibility.
Types of participants
Adults, over 18 years of age, with clinical features of primary Raynaud's phenomenon. Trials including a mixture of primary and secondary participants were included if primary participants could be identified and analysed separately.
Types of interventions
- Oral administration of a calcium channel blocker.
- Another pharmacological therapy
- Other treatments
Types of outcome measures
- Attack rates of Raynaud's phenomenon
- Duration of attacks
- Severity scores
- Participant-preference scores
- Physiological measurements (including digital temperature and blood flow response to hand cooling)
- Adverse reactions (e.g. flushing, headache, tachycardia and ankle swelling), and withdrawal of medication.
Search methods for identification of studies
There were no language or publication status restrictions.
The Cochrane Peripheral Vascular Diseases Group Trials Search Co-ordinator (TSC) searched the Specialised Register (last searched February 2013) and the Cochrane Central Register of Controlled Trials (CENTRAL) (2013, Issue 1), part of The Cochrane Library, (www.thecochranelibrary.com). See Appendix 1 for details of the search strategy used to search CENTRAL. The Specialised Register is maintained by the TSC and is constructed from weekly electronic searches of MEDLINE, EMBASE, CINAHL, AMED, and through handsearching of relevant journals. The full list of the databases, journals and conference proceedings which have been searched, as well as the search strategies used, are described in the Specialised Register section of the Cochrane Peripheral Vascular Diseases Group module in The Cochrane Library (www.thecochranelibrary.com).
The TSC searched the following trial databases (February 2013) for details of ongoing and unpublished studies using the term 'Raynaud';
World Health Organization International Clinical Trials Registry (apps.who.int/trialsearch/)
Current Controlled Trials (www.controlled-trials.com/)
Nederlands Trials Register (www.trialregister.nl/trialreg/admin/rctsearch.asp)
Searching other resources
In February 2013, all trials listed under 'channel blockers' on www.trialscentral.org were checked by the authors for reference to Raynaud's phenomenon with no relevant results found.
Reference lists of relevant studies and reviews were also checked for potentially relevant studies.
Data collection and analysis
Selection of studies
We obtained full-text articles of all the references identified and where necessary translated them. Three review authors (HE, MA and AH) independently reviewed the articles and resolved disagreements by discussion and consensus. One review author (HE) contacted six trial authors or co-authors for additional information on articles where data from subgroups could not be identified. We obtained email addresses by searching the relevant article, the author's or co-author's most recent reference in PubMed, and 'Google'. We received replies from four.
Data extraction and management
Three review authors (HE, MA and AH) independently reviewed each study identified as being eligible for inclusion, and extracted data from included studies using the Cochrane Peripheral Vascular Diseases Group 'Data Extraction Table'. This includes method of allocation, degree of blinding, power calculations, exclusions post-randomisation, losses to follow-up, source of funding, number of participants, age and sex of participants, inclusion and exclusion criteria, treatment, control group, duration of study, and outcome measures. We resolved disagreements by discussion and consensus. Where a trial was described in multiple publications, we extracted data from the most complete report.
Assessment of risk of bias in included studies
Two review authors (MA, AH) independently assessed trial quality as being at 'low', 'high' or 'unclear' risk of bias across the following areas, with reference to the criteria listed in the Cochrane Handbook for Systematic Reviews of Interventions (Higgins 2011):
1. Adequate sequence generation.
2. Allocation concealment.
3. Blinding of participants and personnel, and of outcome assessment.
4. Incomplete outcome data addressed.
5. Free of selective reporting.
6. Free of any other bias: i.e. if no other potential sources of bias were identified.
We resolved disagreement about whether or not a trial fulfilled certain quality criteria by consensus, in discussion with the other review authors (HE, JW). All quality criteria ratings and supporting information are listed in the 'Risk of bias' tables (See Characteristics of included studies).
Measures of treatment effect
The rate of attacks was the only outcome for which we deemed it appropriate to conduct meta-analysis. The results of the remaining outcomes were described separately for each study. For attack rate, the measure of treatment effect used was the difference in mean number of Raynaud's attacks in the treatment and placebo groups. Mean numbers of attacks were reported over different time periods across the studies and, in the case of RTS 2000, geometric means were reported, requiring log transformation in order to make the comparison. Accordingly, we used a standardised mean difference as the parameterisation of treatment effect, calculated by the difference in means divided by the pooled standard deviation (SD) for each study. We calculated the individual study estimates of treatment effect in R (R Development Core Team 2012).
Numbers of attacks were reported on different scales across trials. We calculated standardised mean differences for each trial, in order to provide commensurable estimates of treatment effect. Following data synthesis, we transformed the pooled estimate of treatment effect and its 95% confidence interval to an interpretable scale by multiplying by a typical SD, obtained by pooling the standard deviations of the trials in the analysis (Higgins 2011, section 12.6.4).
Unit of analysis issues
The individual participant was the unit of analysis. For cross-over trials, the calculation of standard errors accounted for the fact that observations were paired (Elbourne 2002).
Dealing with missing data
Five trials had missing data due to participant withdrawal. In three of these (Sarkozi 1986; Vayssairat 1991; Wollersheim 1991), participant-level data were not reported and no intention-to-treat analysis had been attempted. For these studies, we analysed the data as reported based on all participants retained in the study. As such, we assumed missing data to be missing completely at random. In one of the five studies (Ettinger 1984), participant-level data were reported, but the number of participants with primary Raynaud's phenomenon was too low to permit meaningful imputation of missing values. One study (RTS 2000) presented results having conducted multiple imputation to account for missing data. We used these results in the meta-analysis. We contacted two authors for missing data or if the reporting of data was unclear and, while both responded, no additional data were available.
Assessment of heterogeneity
We examined statistical heterogeneity using a Cochrane's Q test and the I² statistic (Higgins 2003).
Wherever there appeared to be at least moderate statistical heterogeneity (I² > 30%), we performed a random-effects analysis to test the robustness of the results derived from the fixed-effect approach. Assessment of clinical heterogeneity was based on consideration of the protocols of the studies included in the analysis and their individual estimates of effect size. Wherever we suspected clinical heterogeneity, we performed a sensitivity analysis by repeating the meta-analysis with the clinically heterogeneous trials excluded.
Assessment of reporting biases
We entered primary outcome data from all included studies into a funnel plot (trial effect against trial size) to investigate the possibility of publication bias, but the small number of studies (seven) meant that we could reach no definitive conclusion.
We performed statistical analysis according to the statistical guidelines provided for authors by the Cochrane Peripheral Vascular Diseases Group and from the Cochrane Handbook for Systematic Reviews of Interventions (Higgins 2011).
We combined the results of five cross-over trials and two parallel-group trials in a meta-analysis. These compared the calcium channel blockers nifedipine and nicardipine with placebo for the outcome 'frequency of Raynaud's attacks'. We calculated estimates of treatment effect and standard errors for each of the studies.
Two studies did not report SDs for treatment or placebo groups, and did not contain sufficient information to calculate them. This did not, however, prevent the calculation of a standardised mean difference for either study. For one study (Wollersheim 1991), the mean difference in the number of attacks, and therefore the standardised difference, was zero. We made a conservative assumption that there was no correlation between participants' measurements, resulting in a larger standard error and a relatively low weighting for the study within the meta-analysis. The other study (RTS 2000) reported geometric means for attack rates and adjusted P values. By using the upper limit of the P value and log-transforming the geometric means, it was possible to calculate a standard error for the difference in means. From this, it was possible to calculate a standardised mean difference and a corresponding standard error.
We then pooled the estimates using the generic inverse variance method, whereby each study is weighted according to the precision of its estimate of treatment effect. Following synthesis, we back-transformed the pooled estimate to an interpretable scale, as described above.
We used a forest plot to show individual study estimates and 95% confidence intervals (CI), together with the pooled estimate. We conducted the meta-analysis using the Review Manager 5 software (RevMan 2012) provided by The Cochrane Collaboration.
Subgroup analysis and investigation of heterogeneity
We performed separate subgroup analyses for the nifedipine and nicardipine trials. To investigate the influence of clinically heterogeneous trials, we performed sensitivity analyses with the trials in question excluded.
We conducted sensitivity analyses to examine the robustness of the meta-analysis as follows:
- If a study had a discordant estimate of treatment effect, we repeated the analysis with that study removed.
- If a study had a weighting that seemed disproportionate to its sample size, we repeated the analysis with that study removed.
If an analysis displayed at least moderate statistical heterogeneity (e.g. I² > 30%) we repeated the analysis using the random-effects method.
Description of studies
Results of the search
See Figure 1.
|Figure 1. Study flow diagram.|
Seven trials were eligible for inclusion in the review. Three trials had multiple publications (Ettinger 1984; Kahan 1987; RTS 2000). In two cases (Ettinger 1984; RTS 2000), we extracted data from the main trial report and in one case (Kahan 1987) from the English language publication. See the Characteristics of included studies table. The seven included trials had a total of 296 participants with primary Raynaud's phenomenon. The largest study (RTS 2000) involved 158 participants, three (Sarkozi 1986; Vayssairat 1991; Wollersheim 1991) had between 69 and 16 participants and the remaining three studies (Rodeheffer 1983; Ettinger 1984; Kahan 1987) had six or fewer participants. The studies were all published between 1983 and 2000.
In the 87 full-text articles assessed for eligibility, we identified a total of 11 calcium channel blockers (48 studies involved nifedipine, eight nicardipine, five nisoldipine, four diltiazem, three felodipine, three flunarizine, two isradipine, one amlodipine, one verapamil, one phendilin and one drug in development, with some examining more than one calcium channel blocker). The seven included trials represented two different drugs (four nifedipine (Rodeheffer 1983; Ettinger 1984; Sarkozi 1986; RTS 2000) and three nicardipine (Kahan 1987; Vayssairat 1991; Wollersheim 1991)).
Types of participants
Four trials had a mixture of participants with primary and secondary Raynaud's phenomenon (Rodeheffer 1983; Ettinger 1984; Kahan 1987; Wollersheim 1991). However, results were presented separately for the two groups. The remaining trials included only participants with primary Raynaud's phenomenon. The definitions of primary Raynaud's phenomenon used within the seven trials varied substantially: two referenced specific criteria (Wollersheim 1991 used Allen 1932; Vayssairat 1991 used Priollet 1987), four provided descriptions only (Rodeheffer 1983: symptomatic Raynaud's phenomenon relating to cold or stress without demonstrable systemic disease; Ettinger 1984: episodic digital pallor and cyanosis on cold exposure without associated demonstrable disease; Sarkozi 1986: episodic, well-demarcated, digital pallor or cyanosis in response to cold or emotional stimuli, episodes associated with numbness and pain and no clinical evidence of primary disease; and Kahan 1987: symptomatic, bilateral Raynaud's without secondary disease) while RTS 2000 screened participants for primary Raynaud's using colour charts (Maricq 1988) and, along with Vayssairat 1991, used nailfold capillaroscopy, antinuclear antibody (ANA) testing and physical examinations to rule out secondary Raynaud's. In two trials (Rodeheffer 1983; Sarkozi 1986), references to positive ANA titres or the development of digital gangrene amongst trial participants indicate that some participants diagnosed as primary Raynaud's phenomenon may not have fulfilled what are now the generally accepted criteria.
Major exclusion criteria were secondary Raynaud's phenomenon in three trials (Sarkozi 1986; Vayssairat 1991; RTS 2000) although, as above, there were concerns that some participants with secondary Raynaud's phenomenon may have been included in one of these studies (Sarkozi 1986). Participants were required to be free of vasoactive medications one week prior in one case (Sarkozi 1986), four weeks prior in one case (Wollersheim 1991), and at the start of the trial in four cases (Rodeheffer 1983; Ettinger 1984; Kahan 1987; Vayssairat 1991). Ettinger 1984 also required participants not to take non-steroidal anti-inflammatory drugs (NSAIDS), aspirin and drugs affecting the nervous system. One trial did not list medications as exclusion criteria (RTS 2000).
Baseline frequency of Raynaud's attacks was provided in only one trial (RTS 2000; as an average daily attack rate during the previous cold season) and five trials had a minimum attack rate as an inclusion criterion. Participants were required to have at least two attacks per day during the previous cold season in one trial (RTS 2000), one attack per day in two trials (Ettinger 1984; Wollersheim 1991), two per week in one trial (Sarkozi 1986) and three per week in one trial (Vayssairat 1991). Two trials (Rodeheffer 1983; Kahan 1987) did not stipulate a minimum attack rate.
One study (Rodeheffer 1983) had only female participants and four trials had more female than male participants (Ettinger 1984; Sarkozi 1986; Vayssairat 1991; RTS 2000), which would reflect the prevalence of primary Raynaud's phenomenon in the population. In two trials (Kahan 1987; Wollersheim 1991), no breakdown by gender was available for participants with primary Raynaud's phenomenon. The age range across the studies also reflected the age distribution of Raynaud's phenomenon in the population by including both young and older participants: three trials included participants aged 18 to 65 years, one trial 18 to 68 years, one trial 20 to 49 years and in two cases the mean age was 39 and 45 years. Ethnicity data were provided in only one trial (RTS 2000). All seven trials were based within a hospital setting: three in the USA, two in France, one in Canada and one in the Netherlands. Five trials were conducted during the cold season, one trial ran from January to June (Kahan 1987) and in one trial (Wollersheim 1991) the timing was not specified. Data on smoking habits were provided in only one trial (Sarkozi 1986).
The included trials represented two different drugs: four trials with nifedipine and three with nicardipine. Four trials compared a calcium channel blocker with placebo using a cross-over design, two trials compared a calcium channel blocker with placebo using a parallel design and one study compared a calcium channel blocker with another active drug (dazoxiben) and with placebo using a cross-over design (Ettinger 1984).
Length of studies
The duration of the treatment periods varied between studies: two weeks (Rodeheffer 1983; Ettinger 1984; Kahan 1987; Vayssairat 1991), three weeks (Wollersheim 1991), 10 weeks (Sarkozi 1986) and 12 to 13 months (RTS 2000). Three trials had a run-in period on no medication lasting either one week (Sarkozi 1986; Kahan 1987) or three weeks (Wollersheim 1991). One trial had a one-week run-in period with single-blind placebo treatment (Vayssairat 1991) and two trials had a two-week run-in period with single-blind placebo treatment (Rodeheffer 1983; Ettinger 1984). In one trial, there was a one-month baseline observation period but it was not clear if participants were required to be off medication (RTS 2000).
All the trials were published prior to 2001 and so predated the use of the Raynaud's Condition Score (Merkel 2002). Frequency of Raynaud's attacks was recorded in all seven trials using participant diaries. In one trial (RTS 2000) participants were provided with a colour chart and each verified attack had to be at least 30 minutes apart.
Duration of attacks was recorded in only one trial (Wollersheim 1991).
Severity of Raynaud's attacks was recorded in six of the seven trials (Ettinger 1984; Sarkozi 1986; Kahan 1987; Vayssairat 1991; Wollersheim 1991; RTS 2000). Severity was recorded in five trials using participant diaries whereby each attack was rated according to a scale, although the scales used differed: mild, moderate or severe (Ettinger 1984), 0 to 3 with 0 = mild and 3 = severe (Sarkozi 1986), 0 to 10 scale (Wollersheim 1991), 1 to 4 scale with 1 = mild and 4 = highly severe (Vayssairat 1991) and a four-point scale (RTS 2000). In one trial, severity of attacks was graded at the end of each treatment period using a single scale of 1 to 4 with 1 = slight to 4 = very severe (Kahan 1987). One trial (Rodeheffer 1983) did not record severity of Raynaud's attacks.
Four trials (Rodeheffer 1983; Ettinger 1984; Wollersheim 1991; RTS 2000) used some form of participant-preference scores although again scales used differed substantially across trials. Rodeheffer 1983 asked participants to rate the overall effectiveness of treatment at the end of each treatment period on a five-point scale (marked improvement, moderate improvement, minimal improvement, no change, worse). Ettinger 1984 asked participants to rate overall response to treatment and side effects at the end of each treatment period on a four-point scale (worse or no change, slight change, moderate change, marked change). Wollersheim 1991 asked participants to rate drug effectiveness relative to baseline at the end of each treatment period on a five-point scale (much worse, worse, no difference, better, much better). Finally, RTS 2000 asked participants to rate improvement compared to baseline at quarterly visits.
Physiological measurements were used in five trials (Rodeheffer 1983; Ettinger 1984; Sarkozi 1986; Vayssairat 1991; Wollersheim 1991). Ettinger 1984 examined finger systolic pressure after local cooling at the end of each treatment period but no data were available specifically for participants with primary Raynaud's phenomenon. Rodeheffer 1983 assessed digital-artery systolic pressure at 30 and 15 degrees C. Sarkozi 1986 assessed pulse amplitude of digital blood flow (assessed by photoplethysmography) and time to return to baseline pulse amplitude following cold pressor challenge at baseline and on the last day of treatment. Vayssairat 1991 assessed cold-reactive hyperaemia as measured by skin temperature using cold challenge tests at each visit with the exception of the 'wash-out' period. Wollersheim 1991 examined response to a finger-cooling test as measured by finger skin temperature, laser Doppler flux and transcutaneous oxygen tension.
Sixty-two studies were excluded. Of the sixty-two excluded trials, twenty-three were excluded because they did not meet the review inclusion criteria of a minimum one week wash-out within cross-over designs and a minimum one week run-in at baseline or baseline data on the severity of Raynaud's phenomenon (Smith 1982; Baadsgaard 1983; Kahan 1983a; Kahan 1983b; Winston 1983; Agnelli 1984; Kahan 1985a; Kahan 1985b; Corbin 1986; Finch 1986; Gjorup 1986a; Gjorup 1986b; Hawkins 1986; Kahan 1986; Redondo 1986; White 1986; Challenor 1987; Kallenberg 1987; Rupp 1987; Wigley 1987a; Wise 1987; Van Heereveld 1988; Challenor 1989). Ten trials included participants with both primary and secondary Raynaud's phenomenon and did not present data which allowed data extraction for the primary group (and no additional data could be obtained) (Müller-Bühl 1983; Rhedda 1985; Aldoori 1986; Waller 1986; Costantini 1987; Finch 1988; Ferri 1992; La Civita 1993; La Civita 1996; Martinez 1999). In seven trials the drug was administered only once (single-dose trials) (Kahan 1981; Gush 1987; Lewis 1987; Morgan 1987; Wollersheim 1987; Weber 1990; Dompeling 1992), six trials were open studies (Shcherbakov 1987; Codella 1989; Denton 1999; Dziadzio 1999a; Coleiro 2001; Choi 2009) and participants within six trials did not have primary Raynaud's phenomenon (Sauza 1984; Da Costa 1987; Schmidt 1989; Mascagni 1994; Varela-Aguilar 1997; Scorza 2001). In four trials the active drug was not a calcium channel blocker, or a calcium channel blocker was used in combination with another therapy (Brotzu 1989; Moriau 1993; Csiki 2011; Caglayan 2012). Four trials were not randomised (Creager 1984; Pisenti 1984; Stefenelli 1986; Leppert 1989). In two trials, the definition of the outcome and description of the data collection were lacking (Nilsson 1987; Wasir 1983). Many trials had more than one reason for exclusion. The principle reasons for exclusion only are listed in Characteristics of excluded studies.
Risk of bias in included studies
See: Figure 2. All included studies were described as randomised and double-blinded. No studies were classed as being at high risk of bias.
|Figure 2. Risk of bias summary: review authors' judgements about each risk of bias item for each included study.|
Six of the seven included trials (Rodeheffer 1983; Ettinger 1984; Sarkozi 1986; Kahan 1987; Vayssairat 1991; Wollersheim 1991) did not provide full details of the method of allocation and, as a result, these were classed as having an unclear risk of bias.
All included studies were described as being double-blind with both participants and personnel unaware of the treatment allocation. Data regarding Raynaud's attack frequency, severity and duration were all collected from participant diaries. Overall, from the descriptions provided by authors, the blinding appeared to be adequate.
Incomplete outcome data
Other potential sources of bias
We identified no other potential sources of bias.
Follow-up and exclusions
Only one trial (RTS 2000) discussed intention-to-treat analysis. This study had forty-three drop-outs after randomisation (27%). Two trials (Rodeheffer 1983; Kahan 1987) had no drop-outs or exclusions. Four trials reported loss to follow-up. Ettinger 1984 had three withdrawals (50%) which, in all cases, were due to symptomatic orthostatic hypotension while on nifedipine. Sarkozi 1986 had seven withdrawals (18%), of which five were on nifedipine (two for digital gangrene, two for non-compliance and one for severe flushing) and two were on placebo (one for pregnancy and one for non-compliance). Vayssairat 1991 had nine withdrawals (13%), of which two were for personal reasons and seven due to side effects (five on placebo and two on nicardipine for headache/nausea and malaise/vertigo). Wollersheim 1991 had three (18%) withdrawals due to side effects (one on placebo and two on nicardipine) although the reasons were not recorded.
Power calculations for a comparison of a calcium channel blocker with a placebo for participants with primary Raynaud's phenomenon were given by only one trial (Vayssairat 1991). Overall, the sample sizes in the included trials were small.
Cross-over, carry-over and period effect
All five cross-over studies included in the review (Rodeheffer 1983; Ettinger 1984; Kahan 1987; Vayssairat 1991; Wollersheim 1991) incorporated a wash-out period of at least one week. In three cases (Rodeheffer 1983; Ettinger 1984; Vayssairat 1991), this involved single-blind treatment with placebo and in the remaining two cases (Kahan 1987; Wollersheim 1991) it involved no vasoactive treatment. One study reported that there was no observed order effect (Kahan 1987). In the remaining studies, no testing for period and carry-over effect was reported.
Five trials (Rodeheffer 1983; Ettinger 1984; Vayssairat 1991; Wollersheim 1991; RTS 2000) had support from a pharmaceutical company (two (Vayssairat 1991; Wollersheim 1991) received financial support and three (Rodeheffer 1983; Ettinger 1984; RTS 2000) received donations of the active drug), three trials also had support from charitable bodies (Rodeheffer 1983; Wollersheim 1991; RTS 2000) and the remaining two (Sarkozi 1986; Kahan 1987) did not state their source of funding.
Effects of interventions
Attack rates of Raynaud's phenomenon
The meta-analysis suggested a statistically significant advantage of treatment with calcium channel blockers compared to placebo as measured by the number of Raynaud's attacks (standardised mean difference of 0.23; 95% confidence interval (CI) 0.08 to 0.38, P = 0.003) although the effect size was small and the 95% CI indicated that the actual treatment effect may be very small to moderate (Figure 3; Analysis 1.1). The transformed results, using a pooled SD of 7.48, indicated a decrease in the number of Raynaud's attacks per week on calcium channel blockers as compared to placebo (1.72; 95% CI 0.60 to 2.84). Calcium channel blockers, therefore, could reduce the weekly number of attacks by as few as 0.6, or as many as 2.8.
|Figure 3. Comparison 1 Calcium channel blockers versus placebo, Outcome: 1 Number of attacks.|
Within the meta-analysis, Ettinger 1984 received a high weighting inconsistent with the number of participants but when we conducted a sensitivity analysis the effect of excluding this study made little difference to the overall results ( Analysis 1.2). RTS 2000 was the only trial with a 95% CI for the standardised mean difference which unequivocally suggested a beneficial effect of treatment (the interval did not span zero). Accordingly, the meta-analysis was repeated with RTS 2000 removed. The effect was a reduction of the point estimate and confidence limits of the overall estimate of treatment effect. We no longer observed a statistically significant treatment effect (Figure 4; Analysis 1.3).
|Figure 4. Comparison: 1 Calcium channel blockers versus placebo, Outcome: 1.5 Number of attacks - without RTS 2000 (sensitivity analysis).|
The results are also presented for each drug separately.
(ATC classification code C08CA05: Selective calcium channel blocker with mainly vascular effects; dihydropyridine derivatives)
Four trials were included (Rodeheffer 1983; Ettinger 1984; Sarkozi 1986; RTS 2000): two were parallel (Sarkozi 1986; RTS 2000) and two were cross-over (Rodeheffer 1983; Ettinger 1984). One parallel trial compared 10 mg nifedipine three times a day rising to 20 mg three times a day if no improvement shown at five weeks (Sarkozi 1986) with placebo for a total of 10 weeks. The other parallel trial compared 30 mg sustained-release nifedipine once a day rising to 60 mg if well tolerated for a period of one year (RTS 2000). One cross-over trial compared 20 mg nifedipine three times a day with placebo for a period of two weeks (Ettinger 1984), while the other cross-over trial compared 10 mg nifedipine three times a day rising to 20 mg if well tolerated with placebo for a period of two weeks (Rodeheffer 1983).
The effect of excluding the three nicardipine trials was to increase the point estimate and confidence intervals for the size of the treatment effect (Figure 5; Analysis 1.4). A significant treatment effect was found at the 5% level (P = 0.004). A pooled SD of 7.51 was calculated for the nifedipine trials to give a difference in means of 2.41; 95% CI 0.75 to 4.06.
|Figure 5. Comparison 1: Calcium channel blockers versus placebo, Outcome: 1.2 Number of attacks - Nifedipine trials only.|
Nifedipine therefore could reduce the weekly number of attacks by as few as 0.8, or by as many as 4.1. There was a considerable amount of statistical heterogeneity in the meta-analysis of nifedipine trials alone (I² = 48%). We performed a random-effects analysis to investigate whether the results were robust to heterogeneity and, while this had a negligible effect on the point estimate of treatment effect, the confidence interval was wider and the estimate of treatment effect was no longer statistically significant at the 5% level ( Analysis 1.5). The difference in means was 2.33 (95% CI -0.38 to 5.11).
(ATC classification code C08CA04: Selective calcium channel blocker with mainly vascular effects; dihydropyridine derivatives)
Three cross-over trials were included (Kahan 1987; Vayssairat 1991; Wollersheim 1991): one compared 30 mg nicardipine three times a day with placebo for three weeks (Wollersheim 1991), one compared 50 mg long-acting nicardipine twice a day with placebo for two weeks (Vayssairat 1991) and one compared 20 mg nicardipine three times a day with placebo for two weeks (Kahan 1987). Two trials included participants with both primary and secondary Raynaud's phenomenon, and only one was restricted to primary Raynaud's only.
We found no overall effect of nicardipine (Figure 6; Analysis 1.6). Multiplying the estimate and CI by the baseline SD reported in Vayssairat 1991 gave a result of 1.11 (95% CI -0.44 to 2.59). Nicardipine therefore could increase the weekly number of attacks by as much as 0.4, could decrease the weekly number of attacks by as much as 2.6 or have no effect whatsoever.
|Figure 6. Comparison 1: Calcium channel blockers versus placebo, Outcome: 1.3 Number of attacks - Nicardipine trials only.|
Duration of attacks
Duration of attacks was recorded in only one trial (Wollersheim 1991): the average duration of attacks was 11 minutes in participants in the placebo group and 13 minutes in the nicardipine group but this difference was not statistically significant. Insufficient information was available to reproduce this analysis, and no P values or confidence intervals were provided.
Severity of Raynaud's attacks was recorded in six of the seven trials (Ettinger 1984; Sarkozi 1986; Kahan 1987; Vayssairat 1991; Wollersheim 1991; RTS 2000). In one trial (Ettinger 1984), severity scores were not reported specifically for participants with primary Raynaud's and no results were provided in another trial (RTS 2000). Wollersheim 1991 reported no significant difference between placebo and treatment although insufficient information was available to reproduce the analysis, and no P values or confidence intervals were reported. Vayssairat 1991 reported no significant difference between 60 participants receiving placebo and nicardipine in a cross-over design (mean score ± SD: 1.55 ± 0.7 in placebo versus 1.36 ± 0.8 in nicardipine; difference 0.2 (95% CI of difference 0 to 0.4)). No P value was reported and it was not possible to reproduce the Wilcoxon signed rank test due to a lack of participant-level data. One trial (Sarkozi 1986) reported average percentage change in number of attacks, broken down by severity. A reduction in the average number of mild, moderate and severe attacks was reported for participants on nifedipine, although no statistical comparison was made with the placebo group. In one trial (Kahan 1987) a comparison of severity scores in primary Raynaud's participants was not made, but the authors of this review were able to make a comparison using reported participant-level data with a median score of 2 for three participants while on placebo versus 2 while on nicardipine (Wilcoxon signed rank test, P > 0.9999). We did not pool severity data for meta-analysis for the following reasons: scales were generally described and unvalidated, the timing of assessment differed in one trial (Kahan 1987), and no data on severity were reported in one trial (RTS 2000).
Four trials (Rodeheffer 1983; Ettinger 1984; Wollersheim 1991; RTS 2000) used some form of participant-preference scores although scales used differed substantially across trials. Rodeheffer 1983 asked participants to rate the overall effectiveness of treatment at the end of each treatment period on a five-point scale (marked improvement, moderate improvement, minimal improvement, no change, worse) with no significant difference between placebo and nifedipine (mean score of 0.6 in placebo group versus 2.2 in nifedipine group, P = 0.25 by Wilcoxon signed rank test). Ettinger 1984 asked participants to rate overall response to treatment and side effects at the end of each treatment period on a four-point scale (worse or no change, slight change, moderate change, marked change) and found no significant difference between nicardipine and placebo although no breakdown was available for participants with primary Raynaud's phenomenon. Wollersheim 1991 asked participants to rate drug effectiveness relative to baseline at the end of each treatment period on a five-point scale (much worse, worse, no difference, better, much better) and found no preference for nicardipine over placebo, although again no breakdown was available for participants with primary Raynaud's phenomenon. Finally, RTS 2000 asked participants to rate improvement compared to baseline at quarterly visits and found participants rated symptoms as better on nifedipine in comparison to placebo (33% in placebo group versus 73% in nifedipine group, P < 0.001). We did not pool participant-preference data for meta-analysis for the following reasons: scales were generally described and unvalidated, definitions of participant-preference differed slightly across all four trials, and one trial (RTS 2000) collected data at a different time point.
Physiological measurements (including digital temperature and blood flow response to hand cooling)
Physiological measurements were used in five trials (Rodeheffer 1983; Ettinger 1984; Sarkozi 1986; Vayssairat 1991; Wollersheim 1991). Ettinger 1984 examined finger systolic pressure after local cooling at the end of each treatment period but no data were available specifically for participants with primary Raynaud's phenomenon. Rodeheffer 1983 assessed digital-artery systolic pressure at 30 and 15 degrees C and found no significant change in the mean cold-induced fall in digital artery systolic blood pressure between nifedipine and placebo (P = 0.625 and P > 0.999 respectively, from Wilcoxon signed rank test conducted by the review authors). Sarkozi 1986 assessed pulse amplitude of digital blood flow (assessed by photoplethysmography) and time to return to baseline pulse amplitude following cold pressor challenge at baseline and on the last day of treatment. No significant differences were found between baseline and at the end of treatment in either the nifedipine or placebo groups. Between-group comparisons of physiological measurements after 10 weeks of treatment conducted by the authors of this review found no significant differences for pulse amplitudes (P = 0.896 and P = 0.815 by unpaired t-tests, right and left hands respectively) or for responses to the cold pressor test (P = 0.542 and P = 0.066) for right index and middle fingers respectively). Vayssairat 1991 assessed cold-reactive hyperaemia as measured by skin temperature using cold challenge tests at each visit with the exception of the 'wash-out' period but found no significant increase in reactive hyperaemia between the nicardipine and placebo groups (mean of 0.49 ± -2.4 in placebo versus 0.77 ± -1.6 in treatment group, difference of -0.3 ± -3, 95% CI -1.0 to 0.5, but no P value given and insufficient information to reproduce the analysis). Wollersheim 1991 examined response to a finger cooling test as measured by finger skin temperature, laser Doppler flux and transcutaneous oxygen tension. No effects of nicardipine were found for any of these three parameters, nor were there any differences between participants with primary and secondary Raynaud's phenomenon. P values were not reported but were calculated by the review authors using unpaired t-tests, as insufficient data were available to calculate the standard error of the mean difference required for paired t-tests (finger skin temperature P = 0.830, laser Doppler flux P = 0.570, transcutaneous oxygen tension P = 0.812). Accordingly, we acknowledge that these tests will have reduced power. We did not pool physiological data for meta-analysis because the methodology used varied substantially and the results could not be synthesised.
Adverse reactions for participants with primary Raynaud's phenomenon were reported differently across the seven trials. Rodeheffer 1983 reported a significantly higher incidence of mild transient headaches in the nifedipine group versus the placebo group but no breakdown was available for participants with primary Raynaud's phenomenon. Ettinger 1984 provided no adverse event breakdown but (as above) referred to three withdrawals in the nifedipine group due to hypotension. Sarkozi 1986 indicated that all participants on nifedipine reported side effects and, in addition, there was a significantly higher incidence of palpitations in the nifedipine group versus the placebo group (1/18 in placebo and 7/18 in treatment group, P = 0.020) as well as flushing (0/18 in placebo and 6/18 in treatment group, P = 0.010). Kahan 1987 did not provide a breakdown by treatment group. Wollersheim 1991 did not provide a breakdown but flushing, palpitations and headaches were referred to as common adverse events – 39 adverse events/16 participants in placebo group and 50 adverse events/16 participants in treatment group plus 1 discontinued therapy/16 participants in placebo group and 2 discontinued therapy/16 participants in treatment group. Vayssairat 1991 reported that the number of participants with at least one side effect was significantly higher in the nifedipine group versus placebo (19 versus 7, P < 0.05) with headaches (12 versus 1, P < 0.05) and oedema (9 versus 2, P < 0.05) listed as occurring significantly more often when taking nifedipine compared to placebo. They reported no significant differences in occurrence of flushing and/or erythema, vertigo, nausea, cutaneous rash, 'various' and withdrawal from the trial (no P values reported from McNemar's test, with insufficient detail to reproduce the analysis). RTS 2000 reported a higher incidence of oedema, flushing and headaches (8/77 headaches in placebo group versus 13/77 in treatment group, P < 0.001) in the nifedipine group compared to the placebo group, and two participants in the nifedipine group reported tachycardia.
Summary of main results
This review summarises the evidence for the use of calcium channel blockers in the treatment of primary Raynaud's phenomenon. Of the 69 studies identified by the search (which examined 11 different calcium channel blockers), seven trials with a total of 296 participants were included in the review. The remaining 62 studies did not match the inclusion criteria of this review. Four of the seven included trials investigated nifedipine, and three investigated nicardipine.
Only two trials (Kahan 1987; Vayssairat 1991) provided any detail of statistical comparisons of (unvalidated) severity scores between treatment groups: one of these trials (60 participants) reported a mean severity score of 1.55 on placebo and 1.36 on nicardipine, difference 0.2 (95% CI of difference 0 to 0.4, no P value reported) and the other trial (three participants only with primary Raynaud's phenomenon) reported a median severity score of 2 on both nicardipine and placebo treatment (P > 0.999). The contrasting results may be partially explained by differences in how severity was quantified across the trials, using unvalidated scales. Participant-preference scores were included in four trials, but in only two were results specific to participants with primary Raynaud's phenomenon, and scoring systems differed between trials: scores differed between treatments in only one trial, in which 33% of participants on placebo and 73% on nifedipine reported improvement in symptoms (P < 0.001). Physiological measurements were included as outcome measures in five trials (different methodologies were used in each): in none of these trials were any between-treatment group differences found. Treatment with calcium channel blockers was associated with a number of adverse reactions including headaches, flushing and oedema.
The meta-analysis of the seven trials indicated a statistically significant advantage of treatment with calcium channel blockers compared to placebo as measured by the number of Raynaud's attacks per week. However, this significant treatment effect was no longer observed when the analysis was repeated without RTS 2000. This trial included the largest number of participants (158), was conducted over a longer period and was the one which suggested most benefit from a calcium channel blocker. It is possible that the larger effect size was attributable to the longer follow-up period. The use of multiple imputation and the number of participants within the treatment arm who received 0 mg of nifedipine may contribute to the different findings in RTS 2000, although these factors would be expected to decrease rather than increase the estimate of effect.
When the meta-analysis was repeated for each of the two calcium channel blockers individually, the estimate of treatment effect was statistically significant for the four nifedipine trials but this result was not robust to a secondary analysis using a random-effects method. No overall effect of nicardipine was found.
All the studies reported adverse events although it was not always possible within studies including participants with both primary and secondary Raynaud's phenomenon to determine the adverse reactions for the primary Raynaud's group alone.
Overall completeness and applicability of evidence
This meta-analysis highlighted several difficulties inherent in the existing clinical trials of calcium channel blockers in primary Raynaud's phenomenon:
a) The overall quality of the data was very variable with the result that the only outcome measure that could be compared using a meta-analysis across all seven included trials was the frequency of Raynaud's attacks. This meta-analysis therefore emphasises the need for validated outcome measures for Raynaud's phenomenon.
b) A large number of trials were excluded because it was not possible to separate out participants with primary and secondary Raynaud's phenomenon, or because of inadequate wash-out periods within cross-over trial designs.
c) Sample size was small, resulting from the small number of trials which could be included, and the small numbers of participants included in most of these seven trials.
d) The dose of calcium channel blockers used may not reflect maximum doses used in current clinical practice: for example, many clinicians increase the nifedipine dose to 80 mg daily. Also, in only one study was treatment duration longer than 10 weeks, with most studies having much shorter treatment periods.
e) The categorisation and assessment of some of the participants classed as having primary Raynaud's phenomenon was sometimes unsatisfactory with reports of digital gangrene and positive ANA titres suggesting that some participants had secondary rather than primary Raynaud's phenomenon. However, this has to be set against the 'real-world' background in which the distinction between primary and secondary Raynaud's phenomenon can be difficult; around 1% to 2% per year of those who present with what initially seems to be primary Raynaud's phenomenon go on to develop a systemic sclerosis spectrum disorder or some other underlying disease (Hirschl 2006).
Quality of the evidence
We conducted an assessment of overall quality of the evidence according to the GRADE approach (GRADE Working Group 2004; Schunemann 2006; Guyatt 2008a; Guyatt 2008b; Higgins 2011). All seven trials included in the review were described as randomised, double-blind, placebo-controlled trials and were judged to be sound in terms of design. Allocation concealment was not fully described in six out of the seven trials, and although described as 'double-blinded', details of how blinding was ensured were not provided. Although detail was lacking in this regard, the review authors did not feel that there was a high likelihood of bias as a result.
Our population of interest was people with primary Raynaud's phenomenon and as such we excluded trials where results for primary Raynaud's participants were not identifiable. Although the demographic information reported across the seven included trials was relatively limited, the sample was broadly representative in terms of gender balance and age range. It should be noted, however, that this sample was predominantly hospital-based and all seven trials recruited from hospital clinics, with the exception of RTS 2000 where participants were also recruited via advertisement. A significant proportion of people with primary Raynaud's phenomenon are not referred to hospital for evaluation and treatment: it is likely that those who are may be more severely affected or may not have responded to conservative measures, or both, and these are the participants mainly presented in this meta-analysis. Another potential concern (highlighted above) is that the doses of calcium channel blockers used within the seven trials are lower than the maximum doses administered in clinical practice. The estimates of treatment effect observed in these seven trials may therefore underestimate the effect of calcium channel blockers.
With the exception of RTS 2000, which had a higher 95% lower confidence limit for the standardised mean difference in the number of Raynaud's attacks, the 95% CIs were generally consistent across the trials. A sensitivity analysis with RTS 2000 excluded failed to reproduce the significant treatment difference observed in the meta-analysis of all seven trials. In this review, the longer follow-up time of RTS 2000 has been proposed as a possible explanation for this potentially discordant interval (the difference is not so severe that the intervals do not overlap). Small numbers of participants with primary Raynaud's phenomenon were included in several trials (Rodeheffer 1983; Ettinger 1984; Kahan 1987). The 95% confidence intervals for the standardised mean difference in the number of Raynaud's attacks in both Rodeheffer 1983 and Kahan 1987 were wide, corresponding to a lack of precision in the estimates of treatment effect. The standard error for the estimate of treatment effect in Ettinger 1984, however, was small due to the high within-participant correlation (ρ = 0.953) and the 95% confidence interval calculated from this study was not too wide. Although Rodeheffer 1983 and Kahan 1987 received a very low weighting in the meta-analysis on account of their large standard errors, the presence of studies with small samples constitutes a limiting feature of these trials as a body of evidence. On account of this, the review authors revised their assessment of the quality of evidence provided by these trials to 'moderate'.
Due to the small number of trials meeting the inclusion criteria of this review, a definitive assessment of publication bias using a funnel plot was not possible. However, the search strategy employed in this review was robust, including conference abstracts in addition to published papers. Furthermore, several of the studies included here contained results best described as null findings. We did not deem the risk of publication bias to be substantial enough to downgrade our assessment of the quality of this body of evidence.
Potential biases in the review process
None of the review authors was involved in any of the included or excluded studies, and none of us has any relevant commercial or other conflicts of interest. The Trials Search Co-ordinator conducted extensive searches for relevant articles and for details of ongoing studies, to reduce the risk of publication bias. Three review authors independently assessed all studies. As described in the review, we also attempted to deal with missing information and data by contacting authors. No further information was available, however, and missing data could introduce bias into the results of four of the seven studies where multiple imputation had not been attempted and could not be performed by the review authors. This, in turn, could introduce bias into the results of the meta-analyses presented here. Two trials (Ettinger 1984; Vayssairat 1991) are of concern in this respect, as both received high weighting in the meta-analysis of all trials.
Agreements and disagreements with other studies or reviews
A previous meta-analysis (Thompson 2005) indicated a statistically significant advantage of treatment with calcium channel blockers over placebo in primary Raynaud's phenomenon, with a decrease of between 2.8 to 5 attacks per week. The estimate of treatment effect was statistically significant for nifedipine but not significant for nicardipine or nisoldipine. Inclusion criteria, however, differed from those for this Cochrane review. Trials were included if the number of participants with primary Raynaud's phenomenon exceeded 75%, and trials with a cross-over design but no wash-out period were also included. As a result, the following trials were included by Thompson 2005 but excluded from this review: Kahan 1983a; Kahan 1985a; Kahan 1985b; Aldoori 1986; Corbin 1986; Gjorup 1986a; Gjorup 1986b; Redondo 1986; Waller 1986; Challenor 1987; Nilsson 1987; Challenor 1989. In one case, a trial was excluded by Thompson 2005 on the grounds of insufficient data for primary Raynaud's participants but was included within this review (Wollersheim 1991). Despite these different inclusion criteria, the results presented here can be seen as consistent with the findings of Thompson 2005 .
Thompson 2005 also included data on the severity of Raynaud's attacks within the meta-analysis. However, we decided not to pool these data because of variation across the seven included trials in terms of scale, description and usage. Instead, we have reported the severity results separately for each trial in which they were reported.
Implications for practice
There is moderate-quality evidence that oral calcium channel blockers are minimally effective in the treatment of primary Raynaud's phenomenon as measured by the frequency of attacks. Participants experienced 1.72 (95% CI 0.60 to 2.84) fewer attacks per week on calcium channel blockers compared to placebo. This effect size was small, although may be greater with longer duration of treatment. The results of this review were limited by small sample size and variable overall data quality.
Implications for research
Given that calcium channel blockers are considered by most clinicians as a first-line treatment for primary Raynaud's phenomenon (and are likely to be the comparator drug in clinical trials of new therapies), there is a need to better define their efficacy in adequately powered, well-designed clinical trials. One concern is the current lack of validated, objective outcome measures for the assessment of Raynaud's attacks. The Raynaud's Condition Score (Merkel 2002) has been validated as an outcome measure for Raynaud's phenomenon and future clinical trials should at least include this.
We would like to thank the editorial team for the Peripheral Vascular Disease Group for their advice and comments on this review.
The first published version of the protocol for this review was conceived and written by Dr Ed Housley. A revised version of the protocol was published in Issue 1, 2005 with a new team of authors led by Mr Ian Quirk.
Data and analyses
- Top of page
- Authors' conclusions
- Data and analyses
- Contributions of authors
- Declarations of interest
- Sources of support
- Differences between protocol and review
- Index terms
Appendix 1. CENTRAL search strategy
Protocol first published: Issue 2, 2000
Review first published: Issue 1, 2014
Contributions of authors
HE, MA and AH identified trials, confirmed eligibility, assessed the quality of trials, extracted data and co-authored the review. JW assessed the quality of trials, performed the statistical analysis and co-authored the review.
Declarations of interest
MA has received honoraria and financial support to attend advisory board and educational meetings from Actelion Pharmaceuticals (Coronary artery hypertension meeting and Consensus best practice for digital ulcer management meeting presentations) and Bristol-Meyers Squibb (national presentations). MA's institution receives National Digital Ulcer Registry participation payments (supported by Actelion Pharmaceuticals).
AH has undertaken consultancy work for Actelion Pharmaceuticals and Pfizer, and has spoken at meetings sponsored by Actelion Pharmaceuticals. AH was a principal investigator (PI) on a study sponsored by Orion (study on systemic sclerosis-related Raynaud's phenomenon) and is a PI on studies sponsored by Actelion Pharmaceuticals (studies on systemic sclerosis-related Raynaud's phenomenon and digital ulceration).
HE and JW report no known conflicts of interest.
Sources of support
- No sources of support supplied
- Chief Scientist Office, Scottish Government Health Directorates, The Scottish Government, UK.The PVD Group editorial base is supported by the Chief Scientist office.
Differences between protocol and review
Subgroup analyses for smoking status, pre-interventional severity, gender and age were proposed in the protocol but this was not deemed possible due to the data quality of the included trials.
The outcomes 'physiological measurements for example, finger systolic pressure response to digital cooling' and 'digital temperature and blood flow response to hand cooling' have been combined into the outcome 'Physiological measurements (including digital temperature and blood flow response to hand cooling)'.
Medical Subject Headings (MeSH)
MeSH check words
* Indicates the major publication for the study