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Creighton University
Osteoporosis Research Center
601 North 30th Street
Suite 4841
Omaha, NE 68131
E-mail: jmlappe{at}creighton.edu
Dear Sir:
Ojha et al voice concern that we did not provide enough key information about baseline data and compliance in our study (1) and that a more subdued interpretation of the results might be required. We remind Ojha et al that our random assignment of subjects to treatment groups was designed to eliminate systematic bias between groups with respect to attributes that might have affected the dependent variable, ie, the incidence of cancer. Randomization is a trustworthy and accepted method of allocating unrecognizable responders to the various contrast groups.
Nonetheless, we are pleased to provide additional information. There were no significant differences between the randomized groups in age, height, weight, body mass index, baseline calcium intake, use of hormones, smoking, any of the bone scan variables, prevalence of spine fracture, or any of the laboratory tests (parathyroid hormone, albumin, bone-specific alkaline phosphatase, calcium, serum 25-hydroxyvitamin D, or creatinine concentrations).
Although there was a statistically significant difference in compliance with calcium supplements between those in the calcium-only group and those in the placebo group (mean adherence: 76% compared with 82%, respectively; P = 0.004), there were no significant differences in compliance with vitamin D supplementation between the treatment groups. This should allay the concerns of Ojha et al about potential differential misclassification of the exposure (intervention). Finally, there were no statistically significant differences between groups in the percentage of subjects completing the study (calcium-only group: 86.5%; calcium and vitamin D group: 85.4%, and placebo group: 83.3%).
Ojha et al raise questions about our statistical modeling. We remind them, and assure the Journal's readers, that the purpose of Cox and logistic models is to increase sensitivity and that our results were sufficiently clear cut as to be evident by simple chi-square analysis, as we stated in our article. We used logistic regression mainly to refine our estimate of the effect size.
Ojha et al also states that an insufficient duration of follow-up may have contributed uncertainty to our findings. They support this concern by citing a recent observational study (2) that reported a reduced breast cancer risk associated with self-selected vitamin D intake >800 IU/d and that found that the risk estimate converged to null after 10 y of follow-up. An important congruence between our randomized study and the study by Robien et al is that a higher intake of vitamin D in both instances was associated with a dramatic decrease in the incidence of cancer within a relatively short period of time (4–5 y). Even if the long-term effect of vitamin D were simply to delay the onset of cancer, that outcome would clearly be beneficial and one that individuals at risk of cancer would accept with alacrity. However, it is unlikely to be just a delay. The reduction in cancer incidence that we reported was precisely of the magnitude found in a large number of observational vitamin D studies and meta-analyses of such studies (eg, 2–5). Moreover, the serum 25-hydroxyvitamin D concentration predicted risk as well as did the fact of treatment.
The concerns of Ojha et al about the validity of our study have no sound basis. The design of our study (population-based, random assignment, double-blind, and placebo-controlled), the low dropout rate, and the excellent compliance with treatment provide confidence that the findings are valid. Furthermore, our findings are supported by a large body of epidemiologic, observational, and case-control evidence that vitamin D decreases cancer risk. Finally, vitamin D supplementation is safe and inexpensive. We argue that it is in the public interest to strongly support optimal vitamin D nutritional status.
ACKNOWLEDGMENTS
Neither author had a conflict of interest.
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