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VI. Meta-Analysis of Calcitonin for the Treatment of Postmenopausal Osteoporosis

	A. Abstract

Top A. Abstract B. Background C. Methods D. Results E. Discussion References

 
Objective: To review the effect of calcitonin on bone density and fractures in postmenopausal women.

Data Source: We searched MEDLINE and EMBASE from 1966 to 2000 and examined citations of relevant articles and the proceedings of international osteoporosis meetings. We contacted osteoporosis investigators to identify additional studies and primary authors for unpublished data.

Study Selection: We included 30 studies that randomized women to calcitonin or an alternative (placebo or calcium and/or vitamin D) and measured bone density or fracture incidence for at least 1 yr.

Data Extraction: For each trial, three independent reviewers assessed the methodological quality and abstracted data.

Data Synthesis: Calcitonin reduced the incidence of vertebral fractures, with a pooled relative risk (RR) of 0.46 [95% confidence interval (CI) 0.25–0.87, P = 0.02, n = 1404, 4 trials]. However, the RR from the one relatively large randomized controlled trial (RCT) was 0.79 (95% CI 0.62–1.00, P = 0.05, n = 1108). For nonvertebral fractures, the pooled RR was 0.52 (95% CI 0.22–1.23, P = 0.14, n = 1481, 3 trials). Once again, the single large trial showed a less impressive effect than the smaller trials (RR 0.80, 95% CI 0.59–1.09, P = 0.16, n = 1245). For bone density of the lumbar spine, the pooled weekly dose of 250 to 2800 IU per week resulted in significant increase in the weighted mean difference (WMD) of 3.74 (2.04–5.43, P < 0.01, n = 2260, 24 trials). The combined forearm showed a similar effect, with a WMD of 3.02 (95% CI 0.98–5.07, P < 0.01, n = 468, 9 trials). At the femoral neck, the pooled weighted mean difference showed a nonsignificant trend toward benefit, WMD 3.80 (95% CI -0.32–7.91, P = 0.07, 9 trials, n = 513). Methodologically weaker studies tended to show greater effects on bone density, and the lumbar spine results suggested the possibility of publication bias.

Conclusions: Calcitonin likely increases bone density in postmenopausal women predominantly at the lumbar spine and forearm for weekly doses of greater than 250 IU, although the true effect may be smaller than the pooled estimate would suggest. Calcitonin likely reduces the risk of vertebral fracture; its effect on nonvertebral fracture remains uncertain.

	B. Background

Top A. Abstract B. Background C. Methods D. Results E. Discussion References

 
CALCITONIN IS AN endogenous polypeptide hormone that inhibits bone resorption by osteoclasts (1). A number of randomized trials have suggested that sc or intranasal calcitonin is effective in prevention of trabecular bone loss in late menopause (2, 3, 4, 5). An observational study by Kanis et al. (6) demonstrated a 30% reduction in hip fractures in patients treated with injectable calcitonin. In addition, calcitonin may have an analgesic effect in women with acute vertebral fractures, which appears to be independent of its effect on osteolastic resorption (4, 7).

Salmon calcitonin is approximately 40–50 times more potent than human calcitonin. and the majority of the randomized trials have used salmon calcitonin(7). Calcitonin was initially administered by injection but is now available in an intranasal formulation, which provides 25–50% of the biological activity of the parenterally administered dose (200 IU of nasal calcitonin would be equivalent to 50 IU of injectable). The nasal formulation is widely available in the United States, Canada, and Europe. Although the development of antibodies to calcitonin presents a potential problem, the biological or clinical significance of antibody development remains speculative (5, 8). There is also concern that prolonged exposure to calcitonin may down-regulate the calcitonin receptors on osteoclasts, which could allow the osteoclasts to recover from the suppressive action of calcitonin (9). Intermittent administration of calcitonin has been recommended as a strategy to avoid clinical resistance.

National Osteoporosis Foundation-published guidelines have recommended calcitonin as an alternative to hormone replacement therapy or alendronate for patients who have found other treatments unsuccessful or difficult to tolerate (10).

Two previous meta-analyses of calcitonin demonstrated the efficacy of calcitonin in increasing bone density and decreasing vertebral fractures, but suffered from a number of limitations (11, 12). The authors of one meta-analysis (11) used the number of fractures as opposed to number of individuals with fractures. Neither group of investigators contacted authors or industry sources to confirm data accuracy and obtain important information omitted in the original reports. The authors noted the heterogeneity in treatment effect, but failed to fully explore sources of heterogeneity. The Kanis and McCloskey (12) meta-analyses included pre- and postmenopausal as well as steroid-induced postmenopausal women. Finally, and most important, these reviews failed to include all trials, and in particular were completed prior the recent fracture trial, the Prevent Recurrence of Osteoporotic Fractures trial (PROOF) (13, 14).

We therefore conducted a systematic review and meta-analysis of the efficacy of calcitonin on bone mineral density (BMD) and fractures. Our goals included considering all published and unpublished RCTs, estimating effects on vertebral and nonvertebral fractures, and determining the consistency of calcitonin effect on postmenopausal women across different treatment groups.

	C. Methods

Top A. Abstract B. Background C. Methods D. Results E. Discussion References

 
We followed the procedure defined by the Cochrane Collaboration (15) for conducting systematic reviews as outlined in Section I.

1. Inclusion criteria.
Trials satisfied the following inclusion criteria: 1) RCTs of at least 1-yr duration, comparing calcitonin therapy vs. placebo or calcium and/or vitamin D; 2) outcomes included BMD (sites included outlined in the Section I) or fracture incidence; and 3) participants were postmenopausal women.

2. Study search and selection.
Using the strategy presented in Section I, we searched for relevant studies published from 1966 to October 2000 and hand-searched conference abstract books from international meetings and the proceedings of the Food and Drug Administration. Our search included the following key and text terms: calcitonin, nasal calcitonin, Miacalcin, postmenopausal, fractures, and bone mineral density.

Two reviewers (A.C., V.R.) examined all potentially relevant trials. For abstracts consistent with study eligibility, we obtained the full text. Reviewers resolved disagreements in study selection by consensus.

3. Methodological quality.
Three reviewers (A.C., V.R., N.Z.) independently extracted all data, which included evaluation of each trial for four characteristics related to methodological quality: concealment, intention-to-treat analysis, blinding, and the completeness of follow-up.

4. Data collection.
The reviewers also abstracted data related to study population, treatment duration, dosage, and patient status with respect to outcomes at baseline and end of study. When the article presented inadequate or unclear data, we contacted the authors for additional information.

5. A priori hypotheses regarding heterogeneity.
We developed a priori hypotheses that might explain the heterogeneity of study results. Specifically, we compared groups according to: 1) population—prevention vs. treatment; 2) duration of treatment; 3) route of administration—sc, ip, and rectal vs. nasal; 4) dose of calcitonin; 5) concurrent treatments including total calcium intake or vitamin D; 6) administration—daily vs. intermittent; 7) SD provided or estimated; and 8) individual components of the quality assessment—including concealed randomization, blinding, and loss to follow-up (less than 20% vs. greater).

If the T-score was available, we divided studies into those that restricted their population to women whose bone density was at least two SD values below peak bone mass (treatment) and those that included women in which the bone density was within 2 SD values of the mean (prevention) mass. The precision of quantitative computed tomography (QCT) is not as good as dual x-ray absorptiometry of the posterior/anterior spine (16). QCT provides a measure of volumetric density, and higher rates of bone change have been reported with QCT. Thus, we decided to pool all bone density measures but QCT.

Given that there are different types of osteoporotic fractures and that treatments may impact more on certain types of fractures, we conducted analyses on the following endpoints: all vertebral fractures and nonvertebral fractures.

6. Statistical analysis.
We used a random-effects model, which includes differences between studies in calculating the variance estimate, for all final analyses of all treatment effects (17).

As outlined in detail in Section I, we conducted separate analyses for each bone density site (lumbar spine, femoral neck, total body, and combined forearm; one third distal radius and then one third distal radius and ulna). We constructed regression models, which included parameters for each year, and for nasal vs. other routes of administration. Thus, independent variables were year and route of administration, and the dependent variable the size of the treatment effect. We sought the most parsimonious model for our final analysis.

In considering the impact of dose, we dealt with the great variability in individual calcitonin dose and in frequency of administration by calculating a weekly dose of calcitonin based on the daily dose and the number of times the patients received that dose in the course of a week.

After using the regression results to decide on the extent of pooling across groups with different routes of administration or length of follow-up, we calculated the impact of treatment on bone density for each stratum that remained separate. We decided that in addition to using the regression analyses to inform our pooling decisions, we would pool any two doses if the random-effects CI for one was completely contained within the CI for the other.

We calculated heterogeneity between studies using the ² distribution with N-1 degrees of freedom, where N is the number of studies (17). Irrespective of the results of the tests of heterogeneity, for each of the a priori hypotheses, we divided the studies into two groups based upon the heterogeneity analyses. For each a priori hypothesis, we then tested whether we could reject chance as the explanation of the variability in WMDs between the two groups (18).

For fractures, a weighted average of the relative risk was calculated. We constructed two-by-two tables for vertebral and nonvertebral fractures in each study for which the data were available and calculated the associated RR.

	D. Results

Top A. Abstract B. Background C. Methods D. Results E. Discussion References

 
1. Trial characteristics.
We identified 770 articles from our electronic search strategy and 6 from hand searching the reference lists, and we retrieved 75 for closer examination. Eleven of the trials had no placebo or control group to compare (19, 20, 21, 22, 23, 24, 25, 26, 27, 28, 29). Thirteen trials included control groups, but treatment allocation was determined by a mechanism other than randomization (30, 31, 32, 33, 34, 35, 36, 37, 38, 39, 40, 41, 42). Six of the trials were duplicate populations described in other trials (43, 44, 45, 46, 47, 48). One trial examined only premenopausal women (49). Seven trials presented BMD for less than 1 yr, or treatment duration was less than 1 yr (50, 51, 52, 53, 54, 55, 56). One trial (57) reported only metacarpal index. Three trials measured only the ultra-distal radius site, which was excluded from analyses a priori (58, 59, 60), and three trials measured BMD using QCT measurements (61, 62, 63). For the PROOF trial (13), originally identified in abstract form, we used the most recent data from the 2000 PROOF publication by Chesnut et al. (14).

In total, 30 trials (14, 64, 65, 66, 67, 68, 69, 70, 71, 72, 73, 74, 75, 76, 77, 78, 79, 80, 81, 82, 83, 84, 85, 86, 87, 88, 89, 90, 91, 92) were RCTs (n = 3993, treatment n = 2569) (Fig. 1 and Table 1). All of the trials except two (79, 81) used salmon calcitonin. Sixteen trials were classified as treatment trials, thirteen were classified as prevention trials, and one trial was a combination prevention and treatment trial (68). Fifteen trials used some blinding, and 16 concealed allocation to treatment. Twenty-eight trials provided a description of dropouts and withdrawals; 4 trials had a loss to follow-up of less than 1%, 2 trials from 2–4%, 13 trials from 5 to 20%, and of the remaining 9 trials over 20% were lost to follow-up (Table 1). Large loss to follow-up is a threat to trial validity and can result in biased estimates of the treatment effect, spuriously increasing or decreasing the magnitude of treatment effect. For example, if individuals at greater risk of fracture are lost preferentially from the control arm, the results may bias against the treatment.

View larger version (34K): [in this window] [in a new window]   Figure 1. Search results for the calcitonin review.

View larger version (14K): [in this window] [in a new window]   Figure 2. RR for vertebral fracture after treatment with calcitonin.

View larger version (13K): [in this window] [in a new window]   Figure 3. RR for nonvertebral fracture after treatment with calcitonin.

View larger version (39K): [in this window] [in a new window]   Figure 4. WMD for lumbar spine after treatment with calcitonin (final year, doses >80 IU/wk).

4. Publication bias.
The funnel plots suggested the possibility of publication bias for the lumbar spine BMD outcome (Fig. 5). Although the single large RCT showed a very small effect on lumbar spine bone density, a number of small trials showed a much larger effect. The femoral neck funnel plot did not suggest publication bias. Although the small number of trials available for fracture incidence prevents a strong inference, the observation that the single large trial yielded a much smaller effect than the smaller trials raises the concern of publication bias (Figs. 2 and 3).

View larger version (11K): [in this window] [in a new window]   Figure 5. Heterogeneity of difference in BMD after treatment with calcitonin (final year, doses >80 IU/wk).

	E. Discussion

Top A. Abstract B. Background C. Methods D. Results E. Discussion References

 
In this meta-analysis, we conducted a comprehensive literature search, specified inclusion and exclusion criteria, and conducted a rigorous data analysis. We made a systematic effort to obtain complete data from all published trials.

Our pooled analysis suggests that calcitonin reduces the incidence of vertebral fractures by over 50% (Table 2). In this case, however, there are a number of reasons for skepticism regarding this pooled estimate. The large effect size is driven by the results of 3 small trials with a sample size of 45–164 (Fig. 3). This raises two concerns. Most serious is the possibility of publication bias. Large studies are more likely to provide precise estimates of treatment effect, and one would anticipate the results of smaller studies to be more or less uniformly distributed around the results of larger studies. In this case, the larger PROOF trial demonstrated a point estimate in fracture RR reduction, 21%, appreciably lower than did the small studies. It is possible that other small studies that failed to show a benefit remain unreported.

A second concern relates to the random-effects model we have chosen for our analyses. In general, the random-effects model estimates yield wider CIs than fixed-effect models, and are thus more conservative. However, random-effects models give relatively larger weight to small studies in comparison to fixed-effect models. In this case, in which smaller studies have yielded larger effects, this drives the point estimate of RR downward, potentially inflating the treatment effect.

A final concern relates to the large loss to follow-up, particularly in the PROOF trial. Particularly given the possibility of publication bias we raised above, one would like to look to the PROOF trial to provide the most robust estimate of the treatment effect. The PROOF trial’s loss to follow-up of over 50% (Table 1) makes relying on this study’s results appreciably less secure. Although we failed to find a systematic effect of loss to follow-up in this meta-analysis, or any other of our osteoporosis meta-analyses, loss to follow-up of this magnitude must reduce the strength of any inferences.

All these considerations suggest that the magnitude of the impact of calcitonin on vertebral fracture remains uncertain. One approach to interpretation would suggest that the magnitude of the relative risk reduction is closer to the 21% suggested by the PROOF trial than the 54% suggested by the pooled estimate. Alternatively, it is possible that the large loss to follow-up has biased the PROOF trial against the active treatment. This would occur if those at greater risk of fracture were preferentially lost to follow-up from the control arm. Unfortunately, we find it difficult to escape the conclusion that inferences regarding the magnitude of calcitonin’s effect on vertebral fracture remain weak.

Exactly the same issues apply to our estimate of the effect of calcitonin on nonvertebral fractures. Two of the small trials show large effects (RR reductions of 75% and 40%), whereas the PROOF trial provides a point estimate of the RR reduction of 20%. In this case, the random-effects CI overlaps no effect, giving a nonsignificant point estimate. Whether calcitonin reduces nonvertebral fractures remains unestablished.

Our meta-analysis confirms that calcitonin increases bone density of the lumbar spine and combined forearm, but again a number of issues suggest that the point estimates of the magnitude of effect may be inflated. First, the pattern of results in the lumbar spine analysis, in which the single larger trial reveals a substantially smaller point estimate than many of the smaller trials, suggests the possibility of publication bias. Indeed, Fig. 5 resembles the classic pattern of a funnel plot suggesting publication bias, with studies missing in the left lower quadrant. Several larger studies with point estimates approximating those of the PROOF trial would strengthen this hypothesis.

Second, in the lumbar spine meta-analysis, the methodologically stronger studies demonstrated smaller treatment effects than the weaker studies (Table 4). In particular, concealed trials showed an appreciably smaller effect than unconcealed trials.

Given that nasal calcitonin is the most widely used commercial preparation, further concern about the lumbar spine estimate arises from the fact that effect sizes for lumbar spine proved substantially smaller in studies using nasal calcitonin than in studies of parenteral calcitonin (Table 4). This observation is consistent with the fact that nasal calcitonin has a variable bioavailability and may therefore be less effective. On the other hand, nasal and parenteral calcitonin demonstrated similar effects on femoral bone density (Table 4).

The pooled estimates of calcitonin’s effect on bone density are somewhat lower than those of the bisphosphonates. Reflections on the relation between bone density and fracture reduction in calcitonin vs. other agents would, in our view, be completely speculative. The reasons include the considerable remaining uncertainty about calcitonin’s effect on both bone density and fracture reduction, and the many mechanisms that may impact on antiresorptive drugs’ impact on fracture reduction.

We did not detect any difference in the pooled estimates for lumbar spine BMD when comparing daily with intermittent administration, suggesting that the issue of down-regulation of calcitonin receptors may not be of clinical significance. However, we were not able to adequately determine whether an extended break/holiday from calcitonin, which would theoretically allow for the up-regulation of the receptors, and would result in a larger treatment effect.

It was difficult to confidently estimate pooled RRs for adverse effects due to inadequate reporting in all the trials.

Our results will be useful to clinicians and policy-makers involved in the development and implementation of guidelines for the treatment of osteoporosis. Our results highlight the considerably uncertainty that remains concerning the effects of calcitonin on bone density, and in particular on vertebral and nonvertebral fractures.

	Footnotes

 
Abbreviations: BMD, Bone mineral density; CI, confidence interval; PROOF, Prevent Recurrence of Osteoporotic Fractures; RCT, randomized controlled trial; RR, relative risk; QCT, quantitative computed tomography; WMD, weighted mean difference.

	References

Top A. Abstract B. Background C. Methods D. Results E. Discussion References

 

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