J Orthop Sci (2014) 19:199–203 DOI 10.1007/s00776-013-0500-4
LETTER TO THE EDITOR
Influential factors in bisphosphonates for periprosthetic bone loss after total joint arthroplasty Tiao Lin • Xi Zuo • Shi-Gui Yan
Received: 15 August 2013 / Accepted: 29 October 2013 / Published online: 3 December 2013 Ó The Japanese Orthopaedic Association 2013
To the Editor: We read with interest the article, ‘‘A meta-analysis of bisphosphonates for periprosthetic bone loss after total joint’’, which was based on 17 randomized controlled trials (RCTs) involving 781 patients to evaluate the effect of bisphosphonates (BPs) [1]. Respecting the highly influential status of meta-analysis in the hierarchy of evidence, we wish to bring attention to three important issues. First, the authors adopted the change from baseline (MD) in bone mineral density (BMD, g/cm2) values in order to reduce the bias of different baselines. However, time-point values or percentage changes of BMD are more appropriate as the standard deviations (S) of absolute change values (SD, g/cm2), which were not directly provided in all of the studies included. Furthermore, it is obviously unacceptable to impute the missing SD using S of baseline data (S0) since they are totally different statistical concepts [2]. Statistically, it will widen the 95 % confidential interval (CI) of effect sizes in each study and
This comment refers to the article available at doi:10.1007/s00776-013-0411-4.
Electronic supplementary material The online version of this article (doi:10.1007/s00776-013-0500-4) contains supplementary material, which is available to authorized users. T. Lin S.-G. Yan (&) Department of Orthopaedic Surgery, School of Medicine, Second Affiliated Hospital, Zhejiang University, 88 Jiefang Road, Hangzhou 310009, Zhejiang, People’s Republic of China e-mail: [email protected] X. Zuo Department of Epidemiology and Biostatistics, School of Public Health, Drexel University, Philadelphia, PA, USA
underestimate the power of overall effects, leading to some unexpected insignificant results. With great concern over this, we are encouraged to impute the missing SD values following the instruction of Cochrane handbook 5.0.2 [2]. The SD in four articles were imputed by the formula, SD = MD/t, as t value for MD were achievable (data not shown). The correlation coefficients (Corr) were also calculated to assess the reliability of MD. Due to the same measurement scale (g/cm2), the same time-points, and the consistent MD (Corr [ 0.5), which were very reliable as changes from baseline, the imputed SD in BPs and control groups from Soininvaara’s study [3] was tentatively adopted for other studies. The reason why we picked Soininvaara’s study was its median imputed SD and its similar Corr value between the BPs and control groups (Table 1). Second, compared to previous meta-analysis [4, 5], the most distinct perspective of this article is its subgroup analysis. The authors concluded from their subgroup analysis that the efficacy of BPs in the gender-balanced, shorter duration, and non-nitrogenous BPs groups was not different from that for controls. However, in this metaanalysis, the results of their subgroup analysis should be treated with greater caution, mainly due to imbalanced number of studies between subgroups. Furthermore, these results were confounded by many potential influential factors in the meantime, including varied surgeries, prosthesis, BPs, durations of treatment, and calcium supplementation across studies, which needed further stratification and identification. For example, there were only three studies [6–8] in the ‘‘gender-balanced group’’ subgroup of 6 months (cut-off value: M:F = 1), of which two studies [6, 7] were also presented in the subgroup ‘‘shorter duration’’ (cut-off value: 6 months, 3 studies [6, 7, 9] in total); a similar situation existed in the 12-month group as
123
200
T. Lin et al.
Table 1 Imputation of SD from Soininvaara’s study [3] Group
M0
S0
M6
S6
M12
S12
N
M6D
S6D
t6
Corr6
M12D
S12D
t12
Corr12
BPs
1.45
0.17
1.4
0.22
1.38
0.19
8
0.05
0.0982898
0.5087
0.9042664
0.07
0.0901365
0.7766
0.8804243
Con
1.39
0.22
1.18
0.17
1.11
0.16
11
0.21
0.083829
2.5051
0.9394746
0.28
0.08202
3.4138
0.9555783
M0,6,12: mean of BMD at baseline or 6 or 12 months; S0,6,12: standard deviations of M0,6,12; M6,12D: mean of BMD change from baseline at 6 or 12 months; S6,12D: standard deviations of M6,12D; Corr6, 12: correlation coefficient of M6,12D; t6,12: t value of M6,12D; SD = MD/t; Corr6, 12 = (S20 ? S26,12 - S26,12D)/(2 9 S0 9 S6,12) BPs bisphosphonates, Con control, Corr correlation coefficient
Fig. 1 Weighted mean difference in bone mineral density at 6 months
well. When the authors concluded the insufficient effectiveness of BPs in the gender-balanced and shorter duration groups at the same time, the above two confounders should have been extensively clarified first. Third, we would like to offer some observations regarding their included studies. Three groups of authors provide multiple reports [6, 10–14], two of which (Wang et al. and Yamaguchi et al.) probably reported the same study with different time points. As indicated in the Supplementary Appendix, two reports of each group shared great similarities in the characteristics of the study, including exactly the same intervention in the same
123
institute, a similar baseline data of included patients, and an overlapped period of the recruited patients. Duplicate publication can introduce substantial biases if studies are inadvertently included more than once in a meta-analysis [15]. To detect duplicate publication is tricky; uncertainties could remain even after carefully checking and consideration. In this case, to correspond with the authors of the reports for identification is necessary. If this attempt turns out to be unsuccessful, the data from both studies might be conservatively considered as one trial for reporting purposes based on the great similarities in the two articles. At the least, a sensitivity analysis conducted by omitting the
Influential factors in bisphosphonates
201
Fig. 2 Weighted mean difference in bone mineral density at 12 months
Table 2 Subgroup analysis by other factors in THA studies Factors
BPs
6 months
12 months b
Subgroups
WMD, 95 %CI, Ran
P
N (8 trials) NN (4 trials)
-0.06 (-0.09, -0.04) -0.03 (-0.06, -0.01)
\0.001 0.02
Subgroups
WMD, 95 % CI, Ran
Pb
N (8 trials) NN (4 trials)
-0.07 (-0.10, -0.04) -0.06 (-0.10, -0.01)
\0.001 0.01
Pa = 0.06 M:F (0.5)
Pa = 0.59
IB (7 trials)
-0.06 (-0.09, -0.04)
\0.001
IB (6 trials)
-0.09 (-0.13, -0.05)
\0.001
B (5 trials)
-0.04 (-0.07, -0.02)
\0.001
B (6 trials)
-0.04 (-0.07, -0.02)
\0.001
Pa = 0.21 M:F (1)
Pa = 0.05
IB (9 trials)
-0.06 (-0.08, -0.04)
\0.001
IB (8 trials)
-0.08 (-0.11, -0.05)
\0.001
B (3 trials)
-0.03 (-0.05, -0.00)
0.05
B (4 trials)
-0.03 (-0.07, -0.00)
0.02
Pa = 0.03 Duration (6 months)
Pa = 0.02
L (9 trials)
-0.05 (-0.07, -0.03)
\0.001
L (9 trials)
-0.08 (-0.11, -0.05)
S (3 trials)
-0.05 (-0.09, -0.01)
0.02
S (3 trials)
-0.03 (-0.08, 0.01)
a
0.15
a
P = 0.82 Duration (12 months)
\0.001
P = 0.01
NA
NA
L (7 trials)
-0.08 (-0.12, -0.05)
\0.001
NA
NA
S (5 trials)
-0.04 (-0.07, -0.02)
0.003
NA
Pa = 0.12
BPs (N nitrogenous, NN non-nitrogenous); M:F (the ratio of male to female); IB imbalance, B balance; L longer duration, S shorter duration; NA not applicable; WMD weighted mean differences, CI confidence interval, Ran random-effect model a
Comparison between subgroups
b
Comparison between BPs and control
123
202
T. Lin et al.
Table 3 Subgroup analysis by other factors in THA studies, without Shetty’s study [7] 6 months Factors BPs
Subgroups
12 months b
WMD, 95 % CI, Ran
P
N (7 trials)
-0.09 (-0.11, -0.06)
\0.001
NN (4 trials)
-0.03 (-0.06, -0.01)
0.02
WMD, 95 % CI, Ran
Pb
N (7 trials)
-0.08 (-0.11, -0.05)
\0.001
NN (4 trials)
-0.06 (-0.10, -0.01)
0.01
Subgroups
Pa = 0.008 M:F (0.5)
Pa = 0.34
IB (7 trials)
-0.06 (-0.09, -0.04)
\0.001
IB (6 trials)
-0.09 (-0.13, -0.05)
\0.001
B (4 trials)
-0.05 (-0.07, 0.02)
\0.001
B (5 trials)
-0.05 (-0.07, -0.03)
\0.001
Pa = 0.38 M:F (1)
IB (9 trials) B (2 trials)
-0.06 (-0.08, -0.04) -0.03 (-0.07, 0.00)
Pa = 0.10 \0.001 0.04
IB (8 trials) B (3 trials)
Pa = 0.12 Duration (6 months)
L (9 trials)
-0.05 (-0.07, -0.03)
\0.001
L (9 trials)
-0.08 (-0.11, -0.05)
S (2 trials)
-0.07 (-0.10, 0.03)
\0.001
S (2 trials)
-0.05 (-0.10, 0.00)
NA NA
\0.001 0.003
Pa = 0.07
Pa = 0.56 Duration (12 months)
-0.08 (-0.11, -0.05) -0.05 (-0.07, -0.02)
\0.001 0.07
Pa = 0.33
NA
L (7 trials)
-0.08 (-0.12, -0.05)
\0.001
NA
S (4 trials)
-0.06 (-0.08, -0.03)
\0.001
Pa = 0.26
NA
The differences between subgroups turned out to be insignificant after omitting Shetty’s study [7], and the associated P values were italicized BPs (N nitrogenous, NN non-nitrogenous); M:F (the ratio of male to female); IB imbalance, B balance; L longer duration, S shorter duration; NA not applicable; WMD weighted mean differences, CI confidence interval, Ran random-effect model a
Comparison between subgroups
b
Comparison between BPs and control
potential duplicated studies to check the stability of the overall effect size should be explicitly stated, especially because Wang’s study [12] in 2003 contained the largest sample size. Unfortunately, in this meta-analysis, we did not find that any concerns were addressed on this issue. With these considerations, we performed a second metaanalysis with updated SD using Review Manager 5.1. As expected, the CIs of overall effects were shorter, but moderate heterogeneities across included studies were observed at both 6 months (I2 = 41 %) and 12 months (I2 = 66 %), which could be explained by the involvement of total knee arthroplasty (TKA) studies (Figs. 1, 2). To sort out the bias of TKA studies, subgroup analysis was repeated, restricted to only total hip arthroplasty (THA) studies. As illustrated by Table 2, generally nitrogenous BPs, longer duration and gender-imbalanced subgroups have stronger BPs effects, however, which turned out to be very unstable in the following sensitivity analysis. We noticed that in Shetty’s study [7], only minimal BPs efficacy was found due to only a single intravenous pamidronate administration, which is the most influential confounding factor. After omitting this study, strikingly, the differences between subgroups became insignificant (Table 3). On the other hand, the significant efficacies of BPs versus control were achieved in non-nitrogenous BPs, shorter duration and balanced gender subgroups at both time points with or without Shetty’s study (Tables 2, 3).
123
Therefore, the results on which the authors’ conclusions are based are substantially unsound. Nevertheless, the data we showed is still not optimal, and its accuracy of should be kept under careful consideration. To sum up, the negative conclusions regarding non-nitrogenous BPs, shorter duration and gender balanced group should be re-considered due to three major methodological flaws: the inappropriate imputation of SD, the limited number of trials in subgroups, and the failure in ruling out confounding factors. Conflict of interest of interest.
The authors declare that they have no conflict
References: 1. Zhu L, Zheng W, Zhao FC, Guo Y, Meng BY, Liu HT, Guo KJ. A meta-analysis of bisphosphonates for periprosthetic bone loss after total joint arthroplasty. J Orthop Sci. 2013;18:762–73. 2. Higgins JPT, Green S. Cochrane collaboration. Cochrane handbook for systematic reviews of interventions. Chichester/Hoboken: Wiley-Blackwell; 2008. 3. Soininvaara TA, Jurvelin JS, Miettinen HJ, Suomalainen OT, Alhava EM, Kro¨ger PJ. Effect of alendronate on periprosthetic bone loss after total knee arthroplasty: a one-year, randomized, controlled trial of 19 patients. Calcif Tissue Int. 2002;71:472–7. 4. Bhandari M, Bajammal S, Guyatt GH, Griffith L, Busse JW, Schu¨nemann H, Einhorn TA. Effect of bisphosphonates on
Influential factors in bisphosphonates
5.
6.
7.
8.
9.
10.
periprosthetic bone mineral density after total joint arthroplasty. A meta-analysis. J Bone Joint Surg Am. 2005;87:293–301. Lin T, Yan SG, Cai XZ, Ying ZM. Bisphosphonates for periprosthetic bone loss after joint arthroplasty: a meta-analysis of 14 randomized controlled trials. Osteoporos Int. 2012;23:1823–34. Arabmotlagh M, Pilz M, Warzecha J, Rauschmann M. Changes of femoral periprosthetic bone mineral density 6 years after treatment with alendronate following total hip arthroplasty. J Orthop Res. 2009;27:183–8. Shetty N, Hamer AJ, Stockley I, Eastell R, Willkinson JM. Clinical and radiological outcome of total hip replacement five years after pamidronate therapy. A trial extension. J Bone Joint Surg Br. 2006;88:1309–15. Trevisan C, Ortolani S, Romano P, Isaia G, Agnese L, Dallari D, Grappiolo G, Cherubini R, Massari L, Bianchi G. Decreased periprosthetic bone loss in patients treated with clodronate: a 1-year randomized controlled study. Calcif Tissue Int. 2010;86:436–46. Yamasaki S, Masuhara K, Yamaguchi K, Nakai T, Fuji T, Seino Y. Risedronate reduces postoperative bone resorption after cementless total hip arthroplasty. Osteoporos Int. 2007;18:1009–15. Arabmotlagh M, Rittmeister M, Hennigs T. Alendronate prevents femoral periprosthetic bone loss following total hip arthroplasty:
203
11.
12.
13.
14.
15.
prospective randomized double-blind study. J Orthop Res. 2006;24:1336–41. Wang CJ, Wang JW, Ko JY, Weng LH, Huang CC. Three-year changes in bone mineral density around the knee after a sixmonth course of oral alendronate following total knee arthroplasty. A prospective, randomized study. J Bone Joint Surg Am. 2006;88:267–72. Wang CJ, Wang JW, Weng LH, Hsu CC, Huang CC, Chen HS. The effect of alendronate on bone mineral density in the distal part of the femur and proximal part of the tibia after total knee arthroplasty. J Bone Joint Surg Am. 2003;85-A:2121–6. Yamaguchi K, Masuhara K, Yamasaki S, Fuji T. Efficacy of different dosing schedules of etidronate for stress shielding after cementless total hip arthroplasty. J Orthop Sci. 2005;10:32–6. Yamaguchi K, Masuhara K, Yamasaki S, Fuji T, Seino Y. Effects of discontinuation as well as intervention of cyclic therapy with etidronate on bone remodeling after cementless total hip arthroplasty. Bone. 2004;35:217–23. Tramer MR, Reynolds DJ, Moore RA, McQuay HJ. Impact of covert duplicate publication on meta-analysis: a case study. BMJ. 1997;315:635–40.
123
LETTER TO THE EDITOR
Influential factors in bisphosphonates for periprosthetic bone loss after total joint arthroplasty Tiao Lin • Xi Zuo • Shi-Gui Yan
Received: 15 August 2013 / Accepted: 29 October 2013 / Published online: 3 December 2013 Ó The Japanese Orthopaedic Association 2013
To the Editor: We read with interest the article, ‘‘A meta-analysis of bisphosphonates for periprosthetic bone loss after total joint’’, which was based on 17 randomized controlled trials (RCTs) involving 781 patients to evaluate the effect of bisphosphonates (BPs) [1]. Respecting the highly influential status of meta-analysis in the hierarchy of evidence, we wish to bring attention to three important issues. First, the authors adopted the change from baseline (MD) in bone mineral density (BMD, g/cm2) values in order to reduce the bias of different baselines. However, time-point values or percentage changes of BMD are more appropriate as the standard deviations (S) of absolute change values (SD, g/cm2), which were not directly provided in all of the studies included. Furthermore, it is obviously unacceptable to impute the missing SD using S of baseline data (S0) since they are totally different statistical concepts [2]. Statistically, it will widen the 95 % confidential interval (CI) of effect sizes in each study and
This comment refers to the article available at doi:10.1007/s00776-013-0411-4.
Electronic supplementary material The online version of this article (doi:10.1007/s00776-013-0500-4) contains supplementary material, which is available to authorized users. T. Lin S.-G. Yan (&) Department of Orthopaedic Surgery, School of Medicine, Second Affiliated Hospital, Zhejiang University, 88 Jiefang Road, Hangzhou 310009, Zhejiang, People’s Republic of China e-mail: [email protected] X. Zuo Department of Epidemiology and Biostatistics, School of Public Health, Drexel University, Philadelphia, PA, USA
underestimate the power of overall effects, leading to some unexpected insignificant results. With great concern over this, we are encouraged to impute the missing SD values following the instruction of Cochrane handbook 5.0.2 [2]. The SD in four articles were imputed by the formula, SD = MD/t, as t value for MD were achievable (data not shown). The correlation coefficients (Corr) were also calculated to assess the reliability of MD. Due to the same measurement scale (g/cm2), the same time-points, and the consistent MD (Corr [ 0.5), which were very reliable as changes from baseline, the imputed SD in BPs and control groups from Soininvaara’s study [3] was tentatively adopted for other studies. The reason why we picked Soininvaara’s study was its median imputed SD and its similar Corr value between the BPs and control groups (Table 1). Second, compared to previous meta-analysis [4, 5], the most distinct perspective of this article is its subgroup analysis. The authors concluded from their subgroup analysis that the efficacy of BPs in the gender-balanced, shorter duration, and non-nitrogenous BPs groups was not different from that for controls. However, in this metaanalysis, the results of their subgroup analysis should be treated with greater caution, mainly due to imbalanced number of studies between subgroups. Furthermore, these results were confounded by many potential influential factors in the meantime, including varied surgeries, prosthesis, BPs, durations of treatment, and calcium supplementation across studies, which needed further stratification and identification. For example, there were only three studies [6–8] in the ‘‘gender-balanced group’’ subgroup of 6 months (cut-off value: M:F = 1), of which two studies [6, 7] were also presented in the subgroup ‘‘shorter duration’’ (cut-off value: 6 months, 3 studies [6, 7, 9] in total); a similar situation existed in the 12-month group as
123
200
T. Lin et al.
Table 1 Imputation of SD from Soininvaara’s study [3] Group
M0
S0
M6
S6
M12
S12
N
M6D
S6D
t6
Corr6
M12D
S12D
t12
Corr12
BPs
1.45
0.17
1.4
0.22
1.38
0.19
8
0.05
0.0982898
0.5087
0.9042664
0.07
0.0901365
0.7766
0.8804243
Con
1.39
0.22
1.18
0.17
1.11
0.16
11
0.21
0.083829
2.5051
0.9394746
0.28
0.08202
3.4138
0.9555783
M0,6,12: mean of BMD at baseline or 6 or 12 months; S0,6,12: standard deviations of M0,6,12; M6,12D: mean of BMD change from baseline at 6 or 12 months; S6,12D: standard deviations of M6,12D; Corr6, 12: correlation coefficient of M6,12D; t6,12: t value of M6,12D; SD = MD/t; Corr6, 12 = (S20 ? S26,12 - S26,12D)/(2 9 S0 9 S6,12) BPs bisphosphonates, Con control, Corr correlation coefficient
Fig. 1 Weighted mean difference in bone mineral density at 6 months
well. When the authors concluded the insufficient effectiveness of BPs in the gender-balanced and shorter duration groups at the same time, the above two confounders should have been extensively clarified first. Third, we would like to offer some observations regarding their included studies. Three groups of authors provide multiple reports [6, 10–14], two of which (Wang et al. and Yamaguchi et al.) probably reported the same study with different time points. As indicated in the Supplementary Appendix, two reports of each group shared great similarities in the characteristics of the study, including exactly the same intervention in the same
123
institute, a similar baseline data of included patients, and an overlapped period of the recruited patients. Duplicate publication can introduce substantial biases if studies are inadvertently included more than once in a meta-analysis [15]. To detect duplicate publication is tricky; uncertainties could remain even after carefully checking and consideration. In this case, to correspond with the authors of the reports for identification is necessary. If this attempt turns out to be unsuccessful, the data from both studies might be conservatively considered as one trial for reporting purposes based on the great similarities in the two articles. At the least, a sensitivity analysis conducted by omitting the
Influential factors in bisphosphonates
201
Fig. 2 Weighted mean difference in bone mineral density at 12 months
Table 2 Subgroup analysis by other factors in THA studies Factors
BPs
6 months
12 months b
Subgroups
WMD, 95 %CI, Ran
P
N (8 trials) NN (4 trials)
-0.06 (-0.09, -0.04) -0.03 (-0.06, -0.01)
\0.001 0.02
Subgroups
WMD, 95 % CI, Ran
Pb
N (8 trials) NN (4 trials)
-0.07 (-0.10, -0.04) -0.06 (-0.10, -0.01)
\0.001 0.01
Pa = 0.06 M:F (0.5)
Pa = 0.59
IB (7 trials)
-0.06 (-0.09, -0.04)
\0.001
IB (6 trials)
-0.09 (-0.13, -0.05)
\0.001
B (5 trials)
-0.04 (-0.07, -0.02)
\0.001
B (6 trials)
-0.04 (-0.07, -0.02)
\0.001
Pa = 0.21 M:F (1)
Pa = 0.05
IB (9 trials)
-0.06 (-0.08, -0.04)
\0.001
IB (8 trials)
-0.08 (-0.11, -0.05)
\0.001
B (3 trials)
-0.03 (-0.05, -0.00)
0.05
B (4 trials)
-0.03 (-0.07, -0.00)
0.02
Pa = 0.03 Duration (6 months)
Pa = 0.02
L (9 trials)
-0.05 (-0.07, -0.03)
\0.001
L (9 trials)
-0.08 (-0.11, -0.05)
S (3 trials)
-0.05 (-0.09, -0.01)
0.02
S (3 trials)
-0.03 (-0.08, 0.01)
a
0.15
a
P = 0.82 Duration (12 months)
\0.001
P = 0.01
NA
NA
L (7 trials)
-0.08 (-0.12, -0.05)
\0.001
NA
NA
S (5 trials)
-0.04 (-0.07, -0.02)
0.003
NA
Pa = 0.12
BPs (N nitrogenous, NN non-nitrogenous); M:F (the ratio of male to female); IB imbalance, B balance; L longer duration, S shorter duration; NA not applicable; WMD weighted mean differences, CI confidence interval, Ran random-effect model a
Comparison between subgroups
b
Comparison between BPs and control
123
202
T. Lin et al.
Table 3 Subgroup analysis by other factors in THA studies, without Shetty’s study [7] 6 months Factors BPs
Subgroups
12 months b
WMD, 95 % CI, Ran
P
N (7 trials)
-0.09 (-0.11, -0.06)
\0.001
NN (4 trials)
-0.03 (-0.06, -0.01)
0.02
WMD, 95 % CI, Ran
Pb
N (7 trials)
-0.08 (-0.11, -0.05)
\0.001
NN (4 trials)
-0.06 (-0.10, -0.01)
0.01
Subgroups
Pa = 0.008 M:F (0.5)
Pa = 0.34
IB (7 trials)
-0.06 (-0.09, -0.04)
\0.001
IB (6 trials)
-0.09 (-0.13, -0.05)
\0.001
B (4 trials)
-0.05 (-0.07, 0.02)
\0.001
B (5 trials)
-0.05 (-0.07, -0.03)
\0.001
Pa = 0.38 M:F (1)
IB (9 trials) B (2 trials)
-0.06 (-0.08, -0.04) -0.03 (-0.07, 0.00)
Pa = 0.10 \0.001 0.04
IB (8 trials) B (3 trials)
Pa = 0.12 Duration (6 months)
L (9 trials)
-0.05 (-0.07, -0.03)
\0.001
L (9 trials)
-0.08 (-0.11, -0.05)
S (2 trials)
-0.07 (-0.10, 0.03)
\0.001
S (2 trials)
-0.05 (-0.10, 0.00)
NA NA
\0.001 0.003
Pa = 0.07
Pa = 0.56 Duration (12 months)
-0.08 (-0.11, -0.05) -0.05 (-0.07, -0.02)
\0.001 0.07
Pa = 0.33
NA
L (7 trials)
-0.08 (-0.12, -0.05)
\0.001
NA
S (4 trials)
-0.06 (-0.08, -0.03)
\0.001
Pa = 0.26
NA
The differences between subgroups turned out to be insignificant after omitting Shetty’s study [7], and the associated P values were italicized BPs (N nitrogenous, NN non-nitrogenous); M:F (the ratio of male to female); IB imbalance, B balance; L longer duration, S shorter duration; NA not applicable; WMD weighted mean differences, CI confidence interval, Ran random-effect model a
Comparison between subgroups
b
Comparison between BPs and control
potential duplicated studies to check the stability of the overall effect size should be explicitly stated, especially because Wang’s study [12] in 2003 contained the largest sample size. Unfortunately, in this meta-analysis, we did not find that any concerns were addressed on this issue. With these considerations, we performed a second metaanalysis with updated SD using Review Manager 5.1. As expected, the CIs of overall effects were shorter, but moderate heterogeneities across included studies were observed at both 6 months (I2 = 41 %) and 12 months (I2 = 66 %), which could be explained by the involvement of total knee arthroplasty (TKA) studies (Figs. 1, 2). To sort out the bias of TKA studies, subgroup analysis was repeated, restricted to only total hip arthroplasty (THA) studies. As illustrated by Table 2, generally nitrogenous BPs, longer duration and gender-imbalanced subgroups have stronger BPs effects, however, which turned out to be very unstable in the following sensitivity analysis. We noticed that in Shetty’s study [7], only minimal BPs efficacy was found due to only a single intravenous pamidronate administration, which is the most influential confounding factor. After omitting this study, strikingly, the differences between subgroups became insignificant (Table 3). On the other hand, the significant efficacies of BPs versus control were achieved in non-nitrogenous BPs, shorter duration and balanced gender subgroups at both time points with or without Shetty’s study (Tables 2, 3).
123
Therefore, the results on which the authors’ conclusions are based are substantially unsound. Nevertheless, the data we showed is still not optimal, and its accuracy of should be kept under careful consideration. To sum up, the negative conclusions regarding non-nitrogenous BPs, shorter duration and gender balanced group should be re-considered due to three major methodological flaws: the inappropriate imputation of SD, the limited number of trials in subgroups, and the failure in ruling out confounding factors. Conflict of interest of interest.
The authors declare that they have no conflict
References: 1. Zhu L, Zheng W, Zhao FC, Guo Y, Meng BY, Liu HT, Guo KJ. A meta-analysis of bisphosphonates for periprosthetic bone loss after total joint arthroplasty. J Orthop Sci. 2013;18:762–73. 2. Higgins JPT, Green S. Cochrane collaboration. Cochrane handbook for systematic reviews of interventions. Chichester/Hoboken: Wiley-Blackwell; 2008. 3. Soininvaara TA, Jurvelin JS, Miettinen HJ, Suomalainen OT, Alhava EM, Kro¨ger PJ. Effect of alendronate on periprosthetic bone loss after total knee arthroplasty: a one-year, randomized, controlled trial of 19 patients. Calcif Tissue Int. 2002;71:472–7. 4. Bhandari M, Bajammal S, Guyatt GH, Griffith L, Busse JW, Schu¨nemann H, Einhorn TA. Effect of bisphosphonates on
Influential factors in bisphosphonates
5.
6.
7.
8.
9.
10.
periprosthetic bone mineral density after total joint arthroplasty. A meta-analysis. J Bone Joint Surg Am. 2005;87:293–301. Lin T, Yan SG, Cai XZ, Ying ZM. Bisphosphonates for periprosthetic bone loss after joint arthroplasty: a meta-analysis of 14 randomized controlled trials. Osteoporos Int. 2012;23:1823–34. Arabmotlagh M, Pilz M, Warzecha J, Rauschmann M. Changes of femoral periprosthetic bone mineral density 6 years after treatment with alendronate following total hip arthroplasty. J Orthop Res. 2009;27:183–8. Shetty N, Hamer AJ, Stockley I, Eastell R, Willkinson JM. Clinical and radiological outcome of total hip replacement five years after pamidronate therapy. A trial extension. J Bone Joint Surg Br. 2006;88:1309–15. Trevisan C, Ortolani S, Romano P, Isaia G, Agnese L, Dallari D, Grappiolo G, Cherubini R, Massari L, Bianchi G. Decreased periprosthetic bone loss in patients treated with clodronate: a 1-year randomized controlled study. Calcif Tissue Int. 2010;86:436–46. Yamasaki S, Masuhara K, Yamaguchi K, Nakai T, Fuji T, Seino Y. Risedronate reduces postoperative bone resorption after cementless total hip arthroplasty. Osteoporos Int. 2007;18:1009–15. Arabmotlagh M, Rittmeister M, Hennigs T. Alendronate prevents femoral periprosthetic bone loss following total hip arthroplasty:
203
11.
12.
13.
14.
15.
prospective randomized double-blind study. J Orthop Res. 2006;24:1336–41. Wang CJ, Wang JW, Ko JY, Weng LH, Huang CC. Three-year changes in bone mineral density around the knee after a sixmonth course of oral alendronate following total knee arthroplasty. A prospective, randomized study. J Bone Joint Surg Am. 2006;88:267–72. Wang CJ, Wang JW, Weng LH, Hsu CC, Huang CC, Chen HS. The effect of alendronate on bone mineral density in the distal part of the femur and proximal part of the tibia after total knee arthroplasty. J Bone Joint Surg Am. 2003;85-A:2121–6. Yamaguchi K, Masuhara K, Yamasaki S, Fuji T. Efficacy of different dosing schedules of etidronate for stress shielding after cementless total hip arthroplasty. J Orthop Sci. 2005;10:32–6. Yamaguchi K, Masuhara K, Yamasaki S, Fuji T, Seino Y. Effects of discontinuation as well as intervention of cyclic therapy with etidronate on bone remodeling after cementless total hip arthroplasty. Bone. 2004;35:217–23. Tramer MR, Reynolds DJ, Moore RA, McQuay HJ. Impact of covert duplicate publication on meta-analysis: a case study. BMJ. 1997;315:635–40.
123
