Use of statins and risk of haematological malignancies:
a meta-analysis of six randomized clinical trials and
eight observational studies
Stefanos Bonovas,1,2Kalitsa Filioussi,1Argirios Tsantes3& Nikolaos M. Sitaras1
1Department of Pharmacology, School of Medicine, University of Athens,2Centre for Disease Control and Prevention and3Laboratory of
Haematology, ‘Attikon’General Hospital, School of Medicine, University of Athens, Athens, Greece
Stefanos Bonovas, MD, MSc,
Department of Pharmacology, School
of Medicine, University of Athens,
Tel: +30 210 6040115
Fax: +30 210 8235657
malignancy, meta-analysis, statins
20 January 2007
29 March 2007
19 June 2007
Statins have been suggested to prevent haematological malignancies. Several epide-
miological studies have evaluated this association, while randomized controlled trials
(RCTs) on cardiovascular outcomes have provided relevant data as secondary end-
points. Our aim was to examine the strength of this association through a detailed
meta-analysis of the studies published in peer-reviewed literature.
A comprehensive search for articles published up to December 2006 was performed,
reviews of each study were conducted and data abstracted. Prior to meta-analysis, the
studies were evaluated for publication bias and heterogeneity. Pooled relative risk (RR)
estimates and 95% confidence intervals (CIs) were calculated using the random
Fourteen studies (six RCTs, seven case–control and one cohort study) contributed to
the analysis. Studies were grouped on the basis of study design, and two separate
meta-analyses were conducted. There was no evidence of an association between
statin use and haematological malignancies among either RCTs (RR = 0.92, 95% CI
0.72, 1.16) or the observational studies (RR = 0.83, 95% CI 0.53, 1.29). Similarly, we
found no evidence of publication bias. However, high heterogeneity was detected
among the observational studies.
Our meta-analysis findings do not support a potential role of statins in the prevention
of haematological malignancies.
Statins are some of the most widely prescribed drugs
worldwide . They have been developed as treatment
for lowering cholesterol by inhibiting hepatic 3-
hydroxy-3-methylglutaryl coenzyme A reductase 
and have been shown to prevent cardiovascular events,
not only in the patients with hypercholesterolaemia, but
also in patients with a wide range of cholesterol levels in
a number of large clinical trials.
Although early studies hinted that these agents might
promote several malignant neoplasms at doses similar
to those used in humans , a significantly elevated risk
of cancer has not been reported in the setting of cardio-
vascular disease prevention. On the contrary, recent
British Journal of Clinical Pharmacology
© 2007 The Authors
Journal compilation © 2007 Blackwell Publishing Ltd
Br J Clin Pharmacol
experimental studies have suggested that statins may
have chemopreventive potential against haematopoietic
and lymphatic malignancies [4–6].
A number of meta-analyses have been reported in the
literature, demonstrating no association between statins
and cancer risk [7–11]. However, it is very unlikely that
exposure such as to statins affects the incidence of all
types of cancer, and increases or decreases in a specific
type of cancer are likely to be masked by random varia-
tion in the effects of statins on all other types. The
end-point of all cancers is therefore not very sensitive.
Thus, the effect of statins on the risk of haematological
malignancies remains to be determined. To address this
issue, we carried out a detailed meta-analysis of
the studies published on the subject in peer-reviewed
Materials and methods
Studies were identified by a systematic literature search
of Medline (1966 to December 2006) and Web of
Science (1970 to December 2006) databases. Search
terms included: ‘HMG-CoA reductase inhibitor(s)’ or
‘statin(s)’ combined with ‘leuk(a)emia’ or ‘lymphoma’
or ‘multiple myeloma’or ‘cancer(s)’or ‘neoplasm(s)’or
‘malignancy(ies)’. The titles and abstracts of studies
identified were scanned to exclude any that were clearly
irrelevant. The full text of the remaining articles was
read to determine whether it contained information on
the topic of interest. The reference lists of articles with
information on the topic were also reviewed for addi-
tional pertinent studies.
The studies considered in this meta-analysis were either
randomized controlled trials (RCTs) or observational
studies (case–control or cohort) that evaluated exposure
to statins and risk of haematological malignancies.
Articles were excluded if there were insufficient pub-
lished data for estimating relative risk or a confidence
interval. RCTs were considered eligible if they (i) evalu-
ated a statin therapy compared with placebo or no treat-
ment, (ii) had no other intervention difference between
the experimental and the control group, (iii) had a
minimum duration of 3 years, (iv) enrolled at least 2000
participants, and (v) reported incidence of haematologi-
cal malignancies during the trial.
We did not assess the methodological quality of the
primary studies, since quality scoring in meta-analyses
of observational studies is controversial, as it is for RCTs
[12, 13], because scores constructed in an ad hoc fashion
may lack demonstrated validity and results may not be
associated with quality . Instead, several subgroup
analyses were performed.
Two reviewers abstracted the data independently. The
following data were collected from each study: (i) pub-
lication data, first author’s last name, year of publica-
tion, and country of the population studied; (ii) study
design; (iii) number of subjects; (iv) relative risks (RR)
and 95% confidence intervals (CIs); (v) types of haema-
tological malignancies studied; (vi) definition of statin
exposure; and (vii) control for confounding factors by
matching or adjustments, if applicable.
Risk ratios and 95% CIs were calculated for each
RCT by reconstructing contingency tables based on the
number of subjects randomly assigned and the number
of subjects with incident haematological malignancies
(intention-to-treat analysis). In observational studies, the
risk estimates were extracted that reflected the greatest
degree of control for potential confounders. Differences
in data extraction were resolved by consensus, referring
back to the original article.
Included in this meta-analysis are studies reporting dif-
ferent measures of RR: RCTs (risk ratio), case–control
studies (odds ratio) and cohort studies (rate ratio). In
practice, these measures of effect yield very similar esti-
mates of RR, since the absolute risk of haematological
malignancies is very low .
Studies were grouped on the basis of study design, and
two separate meta-analyses were conducted: one meta-
analysis of RCTs and a second one of observational
studies. This was done to examine consistency of results
across varying study designs with different potential
Summary RR estimates with their corresponding 95%
CIs were derived with the method of DerSimonian and
Laird  by the use of the assumptions of a random
effects model, which incorporates both within- and
between-study variation. Publication bias was assessed
using the Begg and Mazumdar adjusted rank correlation
test  and the Egger regression asymmetry test .
To evaluate whether the results of the studies were
homogeneous, the Cochran’s Q-test was used .
We also calculated the quantity I2that describes the
percentage variation across studies that is due to hetero-
geneity rather than chance . Negative values of
I2were put equal to zero, so that I2lies between 0% (i.e.
no observed heterogeneity) and 100%. High values
would show increasing heterogeneity. Furthermore, the
S. Bonovas et al.
Br J Clin Pharmacol
summary RR estimates derived from the two separate
meta-analyses (meta-analysis of RCTs vs. meta-analysis
of observational studies) were compared with a test of
All P-values are two-tailed. For all tests, a probability
level <0.05 was considered to be statistically significant.
This work was performed according to the guidelines
proposed by the Meta-analysis of Observational Studies
in Epidemiology (MOOSE) group , and the Quality
of Reporting of Meta-analyses (QUOROM) recommen-
dations for improving the quality of meta-analyses of
RCTs . Stata software was used for the statistical
analyses (Stata Corp., College Station, TX, USA).
The initial search strategy yielded 1368 records. After
screening the titles and abstracts, we retrieved 46 poten-
independent studies met the predefined inclusion criteria
[24–37]. Six out of 14 were randomized, double blind,
comes [24–29], seven were case–control studies [30–32,
34–37] and one was a cohort study . The number of
116 in the RCTs, from 24 to 2362 in the case–control
studies and was 1626 in the cohort study.
All RCTs reported site-specific cancer outcomes
including haematological malignancies. It was therefore
possible to conduct a post hoc analysis of these trials and
calculate risk ratios for haematological malignancies in
an intention-to-treat analysis. All observational studies
evaluated exposure to statins and risk of haematological
malignancies, and were controlled for potential con-
founding factors (at least for age) by matching or adjust-
ments. The publication dates of the studies included in
the meta-analysis were between 1996 and 2006. Study
designs, along with the estimated RRs and 95% CIs, are
shown in Table 1 for the RCTs and Table 2 for the obser-
Meta-analysis of RCTs
Six large RCTs contributed to the analysis [24–29]. A
total of 46 852 individuals (females ~21%) participated
in these trials: 23 432 in treatment groups and 23 420 in
placebo groups. The participants had a mean age of
61 years at enrolment and a mean follow-up of approxi-
mately 6.1 years. A total experience of 286 000 person-
years was reached.
Randomized, double-blind, placebo-controlled trials included in the meta-analysis
Statin group Placebo group
outcome RR (95% CI)
4S * Simvastatin4444 Median: 10.417 of 222119 of 2223 0.90 (0.47, 1.72)Incident
ALERT  Fluvastatin2094Mean: 5.1 11 of 1045 18 of 1049 0.61 (0.29, 1.29)
HPS  Simvastatin20536 Mean: 5.0 64 of 10 26952 of 10 267 1.23 (0.85, 1.77)
LIPID Pravastatin9014 Mean: 8.037 of 451252 of 4502 0.71 (0.47, 1.08)
AFCAPS  Lovastatin6605 Mean: 5.212 of 330411 of 33011.09 (0.48, 2.47)
CARE  Pravastatin4159 Mean: 4.88 of 208110 of 20780.80 (0.32, 2.02)
RR, Relative risk (risk ratio); CI, confidence interval. *Numbers in parentheses, reference citation.
Statin use and risk of haematological malignancies
Br J Clin Pharmacol
Four trials [24, 25, 27, 29] reported a lower risk of
haematological malignancies in the treatment group,
whereas the other two trials [26, 28] reported a higher
risk (Table 1). None was statistically significant. Meta-
analysis of all six reports showed no evidence for an
association between statin treatment and risk of haema-
tological malignancies. The overall rate of haematologi-
cal malignancies was 0.64% in the statin group (149
incident cases) and 0.69% in the placebo group (162
incident cases). The association of statin use with hae-
matological malignancies was not statistically signifi-
cant (RR = 0.92, 95% CI 0.72, 1.16). The Cochran’s
Q-test had a P-value of 0.38 [Q = 5.3 on 5 degrees of
freedom (d.f.)] and the corresponding quantity I2was
6%, both indicating very little variability between
studies that cannot be explained by chance.The P-values
for the Begg’s and the Egger’s tests were P = 0.99 and
P = 0.48, respectively, both suggesting that an assump-
tion of no publication bias is reasonable. After stratify-
ing the data in two subgroups (lipophilic [24–26, 28] vs.
lipophobic statins [27, 29]), no statistically significant
association was found between lipophilic or lipophobic
statins andrisk ofhaematological
(Table 3). Figure 1 shows the RRs and 95% CIs from the
individual trials and the pooled results.
Meta-analysis of observational studies
Seven case–control studies [30–32, 34–37] and one
cohort study  evaluated exposure to statins and risk
of haematological malignancies. The meta-analysis
encompassed these eight studies in a total of 365 201
individuals, of whom 5308 had a haematological malig-
nancy. This time, the Cochran’s Q-test had a P-value of
<0.001 (Q = 38.1 on 7 d.f.) and the quantity I2was 82%,
both indicating high heterogeneity between the studies
(Table 3). In contrast, the P-values for the Begg’s and
Observational studies included in the meta-analysis
cases RR (95% CI)
confounders* Type of HM studied
Fortuny et al.
Ireland, Italy and
C-C4568 23620.61 (0.45, 0.84)1–3 Incident lymphoma
Iwata et al.
Landgren et al.
Friis et al.
Graaf et al.
Zhang et al.
Blais et al.
Traversa et al.
C-C1100 221 2.24 (1.37, 3.66)1, 2, 4–6 Incident lymphoma and
Incident myeloma USA C-C 870 179 0.4 (0.2, 0.8)1, 7–9
Denmark Cohort334 754 16260.88 (0.60, 1.29) 1, 2, 10–13 Incident haematological
Incident lymphoma The NetherlandsC-C 2010593 0.28 (0.06, 1.30)1, 2, 12–22
USA C-C1318 6010.5 (0.4, 0.8)1, 9, 23, 24Incident non-Hodgkin
Incident lymphomaCanada C-C264 24 2.17 (0.38, 12.36) 1, 2, 4, 18, 25, 26
ItalyC-C 2222202 1.5 (0.8, 2.6)1, 2Incident leukaemia
HM, Haematological malignancy; RR, relative risk; CI, confidence interval. *1, age; 2, gender; 3, country; 4, year of visit; 5,
serological status for antihepatitis B surface antigens; 6, serological status for antihepatitis C virus antibodies; 7, race; 8,
education; 9, body mass index; 10, calendar period; 11, use of cardiovascular drugs; 12, use of nonsteroidal anti-inflammatory
drugs; 13, use of hormone replacement therapy; 14, geographical region; 15, duration of follow-up; 16, diabetes mellitus; 17,
prior hospitalizations; 18, chronic disease score; 19, chronic use of diuretics; 20, chronic use of angiotensin-converting enzyme
inhibitors; 21, chronic use of calcium channel blockers; 22, use of other lipid-lowering therapy; 23, menopausal status; 24, family
history of non-Hodgkin lymphoma in first-degree relatives; 25, previous neoplasm; 26, use of fibric acids. †Numbers in
parentheses, reference citation.
S. Bonovas et al.
Br J Clin Pharmacol
Results of meta-analysis
RR (95% CI)
Tests of homogeneity
Tests of publication bias
RCTs of lipophilic statins
RCTs of lipophobic statins
0.85 (0.64, 1.12)
0.92 (0.72, 1.16)
1.04 (0.79, 1.37)
0.72 (0.49, 1.06)
0.83 (0.53, 1.29)
0.82 (0.47, 1.41)
0.88 (0.60, 1.29)
RR, Relative risk; CI, confidence interval; d.f., degrees of freedom, RCT, randomized controlled trial.
Randomized trials: pooled estimate
Fortuny et al. (30)
Iwata et al. (31)
Landgren et al. (32)
Friis et al. (33)
Graaf et al. (34)
Zhang et al. (35)
Blais et al. (36)
Traversa et al. (37)
Observational studies: pooled estimate
All studies (n = 14): pooled estimate
Relative Risk (logarithmic scale)
Results from individual studies and meta-analyses. The relative risk and 95% confidence interval for each study are displayed on a logarithmic scale.
Pooled estimates are from a random effects model
Statin use and risk of haematological malignancies
Br J Clin Pharmacol
the Egger’s tests were P = 0.54 and P = 0.66, respec-
tively, both suggesting a low probability of publication
bias. Statin use did not significantly affect the risk of
haematological malignancies (RR = 0.83, 95% CI 0.53,
1.29) (Table 3). Figure 1 shows the RRs and 95% CIs
from the individual studies and the pooled results.
When the analysis was restricted to the seven case–
control studies (exclusion of the cohort study ), the
results did not substantially change (RR = 0.82, 95% CI
0.47, 1.41; Cochran’s P < 0.001 and I2= 84%; Begg’s
P = 0.76 and Egger’s P = 0.66).
We compared the summary RR estimates derived from
the two separate meta-analyses with a test of interaction
.The difference between estimates of statin effect on
the risk of haematological malignancies, in RCTs and
observational studies, was not statistically significant
(Z = 0.3, P = 0.7).
A combined analysis of RCTs and observational
studies was also performed. The Cochran’s Q-test had a
P-value of <0.001 (Q = 40.0 on 13 d.f.) and the quantity
I2was 71%, both indicating high heterogeneity between
the 14 studies (Table 3). In contrast, the P-values for
the Begg’s and the Egger’s tests were P = 0.91 and
P = 0.70, respectively, both suggesting a very low prob-
ability of publication bias. Statin use did not signifi-
cantly affect the risk of haematological malignancies
(RR = 0.85, 95% CI 0.64, 1.12).
There is a long-standing debate concerning the associa-
tion between use of statins and cancer. During preclini-
cal and clinical development of statin-class drugs,
animal studies have shown an increased risk of malig-
nant neoplasms in rodents exposed to statins at doses
similar to those administered to humans . Interest-
ingly, the commonly used statin pravastatin caused
malignant lymphomas in mice at doses that ranged from
0.5 to fivefold the maximum recommended dose for
humans . In contrast, several recent mechanistic, in
vivo and observational studies have suggested that
statins may actually have chemopreventive potential
against cancer at various sites , including haemato-
poietic and lymphatic tissue [4–6, 30].
Meta-analysis serves as a valuable tool for studying
rare and unintended effects of a treatment. It extends
prior randomized and nonrandomized studies by permit-
ting synthesis of data and providing more stable esti-
mates of effect. This meta-analysis of published studies
does not provide evidence that statin use is associated
with a substantially decreased or increased risk of neo-
plastic diseases of the haematopoietic and lymphatic
tissue. Furthermore, our findings are in line with recent
meta-analyses on the association between use of statins
and other site-specific cancers. Likewise, they indicate
that statin use does not substantially affect respiratory
, colorectal  or breast cancer risk . However,
since our meta-analysis pulled all haematological malig-
nancies, the possibility of a beneficial (or harmful) effect
of statins on a specific haematological malignancy
cannot be totally excluded.
When meta-analysis of published literature is per-
formed, consideration of study bias is critical. Existence
of a bias in favour of publication of statistically signifi-
cant results is well documented [41–43]. However, the
likelihood of important selection or publication bias in
our results is small. During the identification and selec-
tion process, no article was excluded because of meth-
odological characteristics or any subjective quality
criteria, and the Begg’s as well as the Egger’s test
revealed no relation between the estimate of RR and
study size. We are therefore confident that important
publication bias due to preferential publication of large
studies with significant findings is unlikely to have
On the contrary, the tests of heterogeneity indicated
very high variability between the observational studies
included in our analysis. A likely explanation is that
these observational studies, whether they showed a ben-
eficial or a harmful effect of statin use on the risk of
haematological malignancies, all appear to suffer from
limitations resulting from selection bias, information
bias and unaccounted confounding. As is usually the
case in any epidemiological study, there are limits to
what can be done to reduce unmeasured or unknown
sources of bias.
Limitations of this meta-analysis stem from the limi-
tations of the primary studies included in the analysis.
The first meta-analysis included a group of trials of
statins for cardiovascular outcomes, which reported the
incidence of haematological malignancies. The exam-
ined populations varied and the risk of haematological
malignancies was about 1.1 in 1000 per year, which
could make it difficult to detect any effects, beneficial or
harmful.Treatment and follow-up times were an average
of only 6.1 years, which might be considered a short
period on which to base definite conclusions compared
with the latency time between the initiation and the
clinical detection of a malignancy.
Detection or surveillance of adverse events, such as
malignancy, may have varied considerably between
trials, for many reasons, including differences in the
threshold of patients and physicians to report; and the
S. Bonovas et al.
Br J Clin Pharmacol
mode of data collection (in particular, active vs. passive
surveillance for harm). Furthermore, the fact that occur-
rence of haematological malignancies was not the
primary objective of these trials might have affected the
detection rate, but this factor would probably have
affected both arms of the trials equally, which would
bias the results towards the null.
On the other hand, the second meta-analysis included
observational studies that lacked the experimental
random allocation of the intervention necessary to test
exposure–outcome hypotheses optimally.They may also
suffer from the ‘healthy volunteer’ bias, which is an
important limitation of observational studies. These
studies were also different in terms of study design and
definitions of drug exposure.
Although randomized and observational designs
sometimes produce equivalent results [44–46], system-
atic reviews have found that they often give different
results, and that the difference is in all directions . In
our case, it is noteworthy that the findings were similar
in both meta-analyses of RCTs and observational
studies, although the primary studies had varying study
designs with different potential biases. This fact
strengthened our confidence in the validity of our
In conclusion, meta-analysis of existing data does not
support a potential role of statins in prevention of hae-
matological malignancies. However, given the high and
growing prevalence of statin use, it is important to
monitor the experience with statins for extended
follow-up periods, to identify potential effects in the
longer term. Until then, physicians need to be vigilant in
ensuring that use of statins remains restricted to the
Competing interests: None declared.
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